When the Consequences Precede Rulemaking: The Case of CEO
Stock Option Grants
Gabriel Bonilla
January, 2017
[Most Recent Version Here]
Abstract
Studies show that firms modified their CEO compensation practices following the adoption in 2005 of SFAS
123R, which changed the accounting treatment of stock options. However, using data from 2000 to 2006, I
document that the most significant changes in firms’ CEO compensation practices, particularly in the use of
stock option grants, took place in 2002 following the accounting scandals that rocked corporate America. I also
document that the magnitude of these changes was proportional to the perceived accounting benefits of using
stock options stemming from the nonneutral accounting treatment to which option grants were subjected under
the accounting rules that prevailed when the wave of scandals occurred. Consistent with the notion that the
perceived accounting benefits were a function of the firm’s perceived ability to manage investors’ perception of
firm profitability, I find that the decline in the use of CEO option grants varied cross-sectionally according to the
composition of the firm’s investor base. I also document that the decrease in the use of stock option grants was
more pronounced for highly visible firms with boards that have greater reputational concerns, consistent with
the view that negative media exposure constitutes a threat that affects corporate boards’ behavior. My findings
suggest that the changes in investors’ focus and public scrutiny experienced during 2002 largely negated the
perceived accounting benefits of using stock option grants before the adoption of SFAS 123R did so formally in
2005.
UPDATE (03/28/2017): Additional results on how mispricing arising from investors’ inattention shaped firms’
financing and investment decisions during the sample examined are available upon request. I find that following the accounting scandals firms with higher values in the proposed accounting benefits metric stopped earning
higher returns than otherwise similar firms and reduced equity issuances and investment. This evidence is consistent with a view that firms exploit investors’ biases to boost their stock price, and consequently secure equity
funding at better conditions.
1
1
Introduction
Chief executive officer (CEO) compensation in publicly traded companies is under constant political scrutiny.
Evidence of how the political climate and political actors—including Congress, the Securities and Exchange Commission (SEC), and organizations responsible for setting accounting and other standards—influence CEO compensation through legislation, tax policies, and accounting rules can be traced as far back as the Securities Act of
1934, which required publicly traded firms to disclose the names and total compensation of their three highestpaid executives, and continue today with the ongoing implementation of the Dodd-Frank Act of 2010. Political
factors have affected the patterns of CEO compensation for many years, as in the surge of stock-option-based
compensation starting in the early 1980s (Murphy, 2013). This political influence has been mainly understood as
operating on the socially desirable implications of protecting investors from wealth transfers that benefit powerful
corporate executives. This influence has come about via requiring shareholder advisory voting and compensation disclosure requirements, as well as on narrowing wage disparities, which some argue can adversely affect
firm value (see Thomas, 2003, and references therein).1 Although any attempt to explain CEO compensation
is incomplete without a consideration of the political climate and its actors, with a few notable exceptions, the
literature has largely overlooked political factors in favor of more traditional explanations, such as market-based
mechanisms and rent extraction.2
This paper studies the influence of political factors on firms’ CEO compensation policy. Using the accounting
scandals that rocked corporate America in late 2001 through 2002 as a quasi-experiment, I document how a
change in political climate—expressed as changes in investors’ focus, public scrutiny, and the regulatory environment—experienced during 2002 largely canceled the effect that a standard-setter—via a nonneutral accounting
rule—had on the use of stock option grants as a form CEO compensation.3 My approach differs from studies
that attribute the cancelation of this effect to the implementation of SFAS No. 123-R in 2005 (see Hayes et al.,
2012; Skantz, 2012; and Ferri and Li, 2016).4 In general, my findings underscore the need to consider with care
the events that lead to rulemaking when analyzing firms’ presumed response to such a rule, instead of simply
assuming that the implementation of a new rule (here an accounting standard) is the turning point.
I analyze data from approximately 1,000 nonfinancial firms operating during a seven-year period, 2000–2006,
1
As part of the Dodd-Frank Act, the SEC has enhanced its protective role mandating that all firms subject to the federal proxy rules provide
shareholders with an advisory vote on executive compensation. The rule adopted in January 15, 2011, specifies that these say-on-pay votes
are required at least once every three years, beginning with the first annual shareholders’ meeting taking place on or after January 21, 2011.
2
See, for example, Baker (1999), who examines the influence of political costs in how aggresively firms report the estimates of the value
of CEO option awards. For surveys of studies that attempt to explain CEO compensation trends and levels using market-based mechanisms
and managerial power rationales, seeBertrand (2009), Edmans and Gabaix (2009), and Frydman and Jenter (2010).
3
The resulting damage caused by this wave of accounting scandals included “a decline in the worldwide reputation of a wide variety of
U.S. institutions, including U.S. generally accepted accounting principles (GAAP), auditors, security analysts, regulators, and financial markets
generally” (Ball, 2009).
4
Skantz (2012) considers the option expensing decision date as the pivotal point in the elimination of accounting benefits. However, as
he describes, most of his sample (80%) involuntarily adopts option expensing with the implementation of SFAS 123-R.
2
surrounding the accounting scandals. In these years, stock options for the typical CEO went from representing
the predominant component of compensation, accounting for roughly half of compensation before 2002, to less
than 20% of compensation by 2006.5 This intriguing and drastic shift away from stock options affords us a
lens through which to identify the role of accounting—specifically, of a nonneutral accounting rule that gave
firms using a particular form of compensation the ability to choose between disclosure and recognition of a
compensation expense—on the design of CEO compensation. Although both tax and accounting considerations
played an important role in the surge in stock option grants, my focus here is on accounting rules, since only these
were modified between 2000 and 2006.6
Under rule SFAS No. 123 (FAS123), implemented in 1995 by the Financial Accounting Standards Boards
(FASB), firms issuing fixed stock options (in which the exercise price and the number of shares are fixed at grant
date) could choose between expensing the fair-market value of the option grants or reporting them following the
“intrinsic value” methodology, with the requirement that they disclose the fair value in a footnote on their financial
statement. The intrinsic value methodology did not require the use of fixed options awarded with an exercise
price (greater or) equal to the grant-date market price to be recorded as an expense. Before 2002, presumably
due to the lack of a charge of option grants’ value against reported earnings, it was standard practice to award
fixed options with strike price equal to the market price and report them under the intrinsic value methodology.7
Not surprisingly, an accounting standard that permitted firms to avoid expensing a compensation cost was highly
controversial.
The original rationale for this accounting standard, which dates back to 1972, when the Accounting Principles
Board (APB)—the predecessor of the FASB—issued Opinion No. 25—the predecessor of FAS123—was that there
were no mechanisms for firms to value stock options accurately.8 In the 1980s, despite the surge of firms’ issuance
of fixed stock option grants and the availability of widely accepted option valuation methodologies, the FASB’s
attempt to require options expensing faced extraordinary opposition from the business world, Congress, and even
President Bill Clinton (see Dechow et al., 1996; and Murphy, 2013).9 Not until scandals rocked the financial world
in the early 2000s was action finally taken to revise the accounting standard. With the implementation of SFAS
5
Frydman and Jenter (2010) and Murphy (2013) document similar patterns.
Under Section 162(m) of the tax code, introduced as part of the Omnibus Budget Reconciliation Act in 1993, CEO compensation (and that
of a firm’s four other highest-paid executives) above $1 million became nondeductible for tax purposes, unless compensation was considered
performance based. Fixed stock option grants generally qualify as performance-based compensation, thus incentivizing the use of options for
motives other than as stimuli to increase firm performance.
7
In the sample studied, over 90% of firms awarded stock options with strike price equal to the market price at the date of grant before
2002. I exclude the handful of firms that expensed the fair-value of option grants before the summer of 2002; see Mcconnell et al. (2004).
8
This argument shortly became obsolete with the publication in 1973 of the Black-Scholes formula, which triggered a huge boom in markets
for publicly traded options; see Bodie et al. (2003).
9
The unprecedented participation of the U.S. President and Congress in the standard-setting process (SFAS No. 123, para. 376) was the
result of the intense lobbying efforts of corporations; see Cowan (1993). Firms’ decision to spend resources to affect accounting standards
substantiates my conjecture that firms valued the difference between disclosure and recognition of an expense for stock option grants. As
Zmijewski and Hagerman (1981) state: “A firm would not incur lobbying costs if it could counteract the effect of a decision made by the FASB
by changing its set of accounting principles if the cost of this change would be less than the cost of lobbying.”
6
3
No. 123-R (FAS123R), beginning on June 15, 2005, firms were required to recognize the use of options awards
as an expense.
The heterogeneous accounting treatment of stock option grants under FAS123 enabled firms using option
grants instead of other forms of compensation to lawfully boost reported earnings by avoiding the compensation
expense.10 This form of compensation thereby gave firms benefits in addition to the economic advantages of
using option grants (for instance, greater incentives to increase firm performance) to the extent that reported
earnings figures became the key metrics considered by outsiders.11 I refer to these added benefits hereafter as
the accounting benefits of using stock option grants. Using the accounting benefits notion, I classify firms into
“benefiting” firms, or firms with positive accounting benefits, and “nonbenefiting” firms, or firms with neutral or
negative accounting benefits. A firm could perceive positive accounting benefits by using option grants under
several scenarios.12 In this paper, I focus on two such scenarios. The first is when a firm perceives the ability to
manage financial reporting perceptions—to obtain higher valuations—via reported earnings figures because its
investors do not unscramble the information contained in the firm’s financial statement.13 The second is when
a firm is encouraged to take actions to inflate reported earnings because its investors are focused too heavily
on short-term firm performance.14 I construct a firm-specific proxy for the accounting benefits of using options
to capture firms’ differing ability and need to manage financial reporting perceptions before the onset of the
accounting scandals of the early 2000s.
The shocking collapse of the Enron Corporation in December 2001—and the subsequent unearthing of an unprecedented number of accounting malpractices throughout 2002—profoundly shook the confidence of investors
and prompted an intense scrutiny of accounting standards, compensation practices, and the quality of corporate
governance and financial reporting.15,16 This climate of distrust led the nation’s top executives to press for enhanced regulation of corporate governance and increased oversight of accounting practices to restore investors’
10
Additionally, Core et al. (2002a) argues that the treasury stock method –used to incorporate the effect of “potentially dilutive” securities
–such as warrants, convertible debt, and employee stock options (ESOs)– systematically understates the dilutive effect of outstanding stock
options, upwardly biasing reported fully-diluted EPS.
11
Graham et al. (2005) present survey evidence that chief financial officers (CFOs) believe that earnings, not cash flows, are the key metric
considered by external constituents.
12
See Healy and Wahlen (1999); Dechow and Skinner (2000); Fields et al. (2001); and Graham et al. (2005), who discuss several motivations that firms have to influence accounting numbers.
13
Because this mechanism implies that investors price recognized and disclosed information differently, explanations of this sort have faced
considerable resistance by academics who, from a market efficiency perspective, argue that investors should fully internalize and price the use
of stock option grants as long as their value is fully disclosed; see Dechow and Skinner (2000). That said, Hirshleifer and Teoh (2003) model
how investors’ inattention and limited processing ability can result in the mispricing of disclosed, unrecognized information; Cai et al. (2008)
provide empirical support their hypotheses. Moreover, Sloan (1996) and Xie (2001) document evidence suggesting that investors “fixate” on
earnings, thereby questioning market efficiency.
Admittedly, the jury is still out on whether recognition and disclosure are substitutes. For evidence in favor of this assertion, see, e.g.,
Espahbodi et al. (2002); Bell et al. (2002); Cheng and Smith (2013); Yu (2013); Israeli (2015); for evidence against it, see, e.g., Aboody
et al. (2004a); Laux and N’Dir (2007); Bratten et al. (2013).
14
Jacobs (1991); Porter (1992); and Laverty (1996; 2004) discuss how the focus in short-term performance leads firms to take decisions
that boost reported earnings.
15
The Corporate Scandal Sheet (Patsuris, 2002) lists the most prominent cases exposed from 2001 to 2002.
16
For examples of press coverage on investors’ lack of confidence, see Browning and Weil (2002); Frank (2002); Sandberg et al. (2002);
and Yergin (2002).
4
plummeting confidence.17 Other salient voices of influence, including former SEC chairman Arthur Levitt and
Federal Reserve chairman Alan Greenspan, focused on what they called “poorly structured” CEO compensation
packages—with an emphasis on the use of stock option grants—to explain the tsunami of accounting scandals
and call for reform.18 The business press duly echoed the public commentary about the use of stock option grants
and reinforced a growing stigma toward this form of compensation.19 But the media went beyond echoing comments, covering the accounting scandals with an intensity that some considered exaggerated (Ball, 2009) and
others deemed of high social value and worthy of recognition.20 The negative public sentiment was such that
the media seemed to have license to criticize and instruct corporate America, with corporate boards as the most
criticized party (Joe et al., 2009). In response, lawmakers and regulators rushed to design, debate, and pass
distinct bodies of reforms with the objective of restoring investors’ confidence, among them the Sarbanes-Oxley
Act (SOX) and the NYSE’s and Nasdaq’s increased listing requirements.
Although the possibility that these multiple events—which both directly and indirectly led to the passing of
FAS123R—effectively eliminated the accounting benefits of using stock options well before FAS123R was implemented is certainly not negligible, the literature has overlooked the analysis of these events in this context.
Ultimately, if FAS123 had a sizable effect on the design of CEO compensation, then observed changes in compensation must be a function of the timing of the effective elimination of the perceived accounting benefits stemming
from the rule rather than a function of the date of its modification, FAS123R, which presumably only formally
eliminated these benefits. A variety of mechanisms could have prompted the elimination of accounting benefits following the events set off by the scandals. First, investors’ increased focus on the quality of earnings and
compensation practices may have lessened firms’ ability to manage investors’ perceptions about firm profitability.
Second, pay-setters may have decided to forgo the accounting benefits of using stock option grants and reduce
their use for fear of being portrayed negatively in the press for using a form of compensation seen as one of the
chief culprits of the corporate scandals. Third, the change in the regulatory environment that increased the pressure to provide more transparent reported earnings may have curbed the firms’ perceived accounting benefits of
using stock option grants as compensation. It is difficult and beyond the scope of this study to rank these mech17
For two notable examples, Goldman chairman and CEO Henry M. Paulson Jr., speech at the National Press Club, “Restoring Investor Confidence: An Agenda for Change,” http://www.content.gs.com/media-relations/press- releases/archived/2002/paulson-restore-confidencespeech.pdf; and James Turley, CEO of Ernst & Young, “How Accounting Can Get Back Its Good Name,” Wall Street Journal, February 4,
2002.)
18
Testimony of Alan Greenspan, Chairperson of the Federal Reserve Board, before the Committee on Banking, Housing, and Urban Affairs,
U.S. Senate, July 16, 2002 (Greenspan, 2002): “The highly desirable spread of shareholding and options among business managers perversely
created incentives to artificially inflate reported earnings in order to keep stock prices high and rising. This outcome suggests that the options
were poorly structured, and, consequently, they failed to properly align the long-term interests of shareholders and managers, the paradigm
so essential for effective corporate governance. The incentives they created overcame the good judgment of too many corporate managers. It
is not that humans have become any more greedy than in generations past. It is that the avenues to express greed had grown so enormously.”
19
For one of many examples, see David Wessel, “Why Boardroom Bad Guys Have Now Emerged en Masse,” June 20, 2002, in which he
describes CEO stock option awards as one of the culprits of the corporate transgressions.
20
The Wall Street Journal received the 2003 Pulitzer Prize in explanatory reporting “for its clear, concise and comprehensive stories that
illuminated the roots, significance and impact of corporate scandals in America”; http://www.pulitzer.org/winners/staff-57.
5
anisms by their ability to eliminate accounting benefits. Yet approaching the analysis from varying perspectives
sheds light on the source of the accounting benefits of using options to firms as well as firms’ incentives to modify
compensation practices in response to changes in the political and economic environment.
To identify the effect of the nonneutral accounting of options in the design of CEO compensation, the empirical
strategy exploits that firms differed in their level of perceived accounting benefits before the unexpected arrival of
the accounting scandals. Changes in firms’ CEO compensation practices should therefore be proportionate to their
pre-2002 accounting benefits if the scandals and events that followed catalyzed a market-wide revision of these
benefits. I quantify this effect using a Difference-in-Differences regression model (DiD) that allows firms’ CEO
compensation practices to be treated with differing intensity, where I define treatment intensity using the proxy of
firm-specific perceived accounting benefits of using stock options.21 This approach identifies the role of accounting
for options by comparing changes in CEO compensation practices in firms that differed in their perceived level
of accounting benefits under the assumption that, in the absence of the accounting scandals, firms’ pay practices
would have followed common trends. In other words, the setup identifies this effect by comparing benefiting
(treated) firms both against nonbenefiting (control) firms, as in the standard DiD, and against benefiting firms
that differed in their perceived level of accounting benefits, treating them as controls as well.22
My analyses reveal that firms’ use of stock option grants between 2000 and 2006 experienced a significant
decline following the accounting scandals in 2002 and continued to decrease over the subsequent four years,
although at a slower rate. Moreover, I confirm that these temporal patterns were a function of the accounting
benefits of options. In terms of their perceived accounting benefits before the scandals (treatment intensity), the
typical firm decreased its use of stock option grants by approximately 44% by 2002 and further altered its use to
reach a 65% decrease by 2006. When I compare compensation practices of two firms that differ only in terms of
their perceived accounting benefits before the scandals by one standard deviation, I find that the firm treated with
more intensity (high accounting benefits) reduced its use of option awards by 27% more relative to the other firm.
These results are robust to alternative definitions of accounting benefits, alternative definitions of industry used
to control for industry-wide effects, and the inclusion of firm characteristics to control for known determinants of
compensation policy, and are not driven by nonparallel trends in compensation practices before 2002.23
If the estimated decrease in the use of stock option grants reflects a heightened investors’ focus on the quality of
earnings and CEO compensation because of the lack of confidence in firms’ financial reporting practices following
the corporate scandals, then these changes should be more pronounced in firms that perceived a greater ability
21
It is important to point out that estimations using the accounting benefits construct as a regressor are subject to the usual measurement
error caveats since accounting benefits is a generated regressor measured with error.
22
I use the term “standard DiD” to refer to a model that does not accommodate differing treatment intensity and classifies firms into two
groups only: treated and control firms (e.g., Angrist and Krueger, 1999).
23
I include firm characteristics in a way that avoids biasing the DiD estimates of interest. See the discussion of “bad controls” in Angrist and
Pischke (2008).
6
and need to manage market participants’ financial reporting perceptions since they are presumed to have used
options for this purpose before the scandals of 2002. For instance, firms with an investor base largely composed of
unsophisticated or sophisticated short-term investors should implement more significant revisions than firms with
a base dominated by sophisticated long-term investors. Consistent with this conjecture, in one of my specifications,
I find that firms with an investor base dominated by either unsophisticated investors or sophisticated short-term
investors reduce their use of options by 50%; whereas, firms with an investor base dominated by sophisticated
long-term investors reduce their use of options by 23%, significantly less than the other two groups of firms. These
results are consistent with the hypothesis that the composition of the investor base was an important determinant
of perceived accounting benefits before the scandals.
Business media could have also played an important role in disincentivizing the use of option awards by
changing the norm of “socially acceptable” CEO compensation practices following the accounting scandals. Because readers in the United States rely on the press to form their opinions (Zingales, 2000), business media’s
stigmatization of the use of stock option grants might have constituted a threat to the reputation of pay-setters in
firms that used option-based compensation before the scandals.24 Certainly, this threat should be more relevant
for pay-setters of firms that are more likely to attract media attention. Consistent with this logic, I find that the
compensation practices of firms with high analyst following (highly visible), which are more likely to be featured
in media headlines (Brown and Caylor, 2005), exhibited a differential response depending on the importance of
their boards’ reputational concerns. Firms governed by boards with stronger reputational concerns reduced the
use of stock option grants 30% more than firms governed by boards with weaker reputational concerns. This difference is statistically significant. In contrast, firms with low analyst following (less visible) show no differentiated
response as a function of their boards’ reputational concerns. These results suggest that pay-setters’ reputational
concerns affected firms’ response to the changes in the political climate and that under certain conditions media
can act as a substitute for regulatory action in making financial reporting practices more transparent. Moreover,
from a policy perspective, the results suggest the potential to affect corporate governance and its practices via
shaming mechanisms.25
I also conduct a battery of tests to rule out alternative explanations that could be confounding my analyses.
For instance, I corroborate that my results are robust to the exclusion of firms that decided to expense options
voluntarily (Aboody et al., 2004a); firms that were targeted with shareholder proposals calling to expense options
24
Dyck and Zingales (2002) discuss how media pressures CEOs to behave according to societal norms.
The change in regulatory environment introduced by the implementation of SOX in 2002 may have also played a significant role in
disincentivizing the use of stock option grants to boost reported earnings. Determining its effect is challenging, however, since most of
its provisions affected all public firms equally, with the exception of Section 404, in which compliance was exempted to smaller reporting
companies—firms with a public float below $75 million—on the presumption that compliance could hinder their viability. Section 404 requires
firms to elaborate and periodically assess procedures to ensure the adequacy of their financial reporting practices (SEC press release 33-8238).
I use this exemption to examine whether section 404 led to differential modifications of the use of stock option awards. I find no significant
evidence. However, this analysis presents several limitations that I discuss in more detail in section 5.4.
25
7
(Ferri and Sandino, 2009); and “new economy” (technology and communications) firms, the most prominent
users of stock option grants before 2002. The first two sets of firms have been documented to reduce the usage
of option-based compensation during the period I examine; their exclusion rules out the possibility that signaling
purposes or shareholder activism are behind these findings. The exclusion of the third group of firms alleviates
concerns that the dot-com crash drives my results and that the reported effects are concentrated among these
firms. The enactment of SOX and the improved board-related rules imposed by the NYSE and Nasdaq could
have also affected CEO compensation practices through mechanisms other than the accounting benefits of using
options. SOX, for example, effectively eliminated the ability to backdate option grants by requiring firms to report
them within two business days (see Heron and Lie, 2007). To ameliorate the concern that this new requirement
drives my results, I show that firms that use only scheduled option grants—which offer less room for backdating
(Aboody and Kasznik, 2000)—also exhibit a sharp reduction in the use of option awards after the scandals. In
regards to the new board requirements imposed by the two major U.S. stock exchanges, I show that my results still
hold after controlling for the changes in the structure of boards induced by these listing requirements. Finally,
I conduct placebo tests with a twofold objective: first, to provide assurance that the results are not a function
of the assumptions made in the construction of the accounting benefits metric, and second, to corroborate the
notion that the perception about the accounting benefits of using options was revised shortly after the arrival of
the accounting scandals, thereby leading to a revision of compensation practices in 2002.
To complement the analysis discussed to this point, I present an analysis that does not depend on the accounting benefits construct and is instead based on the discontinuities in earnings distributions at key earnings
benchmarks. As in the accounting literature, I interpret these discontinuities as evidence of firms’ intent to boost
earnings in order to avoid missing these benchmarks.26 In particular, I focus on the structuring of CEO compensation contracts as one of the tactics employed to achieve this financial reporting objective. Although this strategy
is limited in the sense that it does not capture the magnitude of the presumed widespread practice of boosting
earnings through stock option grants, it allows me to test whether there is a revision of the perceived accounting
benefits of using options in the postscandal period through the lens of a group of firms in this particular financial
reporting scenario. Specifically, I test whether during the prescandal period firms close to missing the prior performance benchmark had the ability to meet or beat this benchmark using CEO compensation and whether they
lose this ability after the scandals. I find that, on average, firms suspected of meeting or beating the benchmark
using stock option grants nearly doubled the use of this form of compensation relative to a scenario in which they
were considered nonsuspect before the scandals, whereas after the scandals, this difference became statistically
26
This interpretation, however, is not without controversy. Evidence in Durtschi and Easton (2005; 2009) suggests alternative explanations
for the discontinuities. That said, Gilliam et al. (2014) and Burgstahler and Chuk (2015) have both provided complementary arguments
that challenge the results of Durtschi and Easton (2005; 2009) and further support the view that interprets the empirical discontinuities as
evidence of firms’ intent to manage reported earnings with a financial reporting objective.
8
indistinguishable from zero. These results suggest that firms lose the ability to boost reported earnings through
stock option grants after the scandals, consistent with the evidence documented using the accounting benefits construct. In addition, I document that the earnings discontinuity at the prior performance benchmark disappears
after 2002. Gilliam et al. (2014) document a similar finding with respect to the loss avoidance (zero-earnings)
discontinuity.
Overall, I interpret the results as follows. Following the events set off by the accounting scandals of 2001–2002,
firms changed their CEO compensation practices in a manner consistent with a market-wide revision of the perceived accounting benefits of using options. Whereas prior research has associated similar effects to the implementation of FAS123R (see Hayes et al., 2012; Skantz, 2012; and Ferri and Li, 2016), I show that changes
begin to take place well before 2005, induced by events that most likely motivated and shaped the design of
this accounting rule. The most affected component of CEO compensation was stock option grants, which was
substantially and progressively reduced between 2002 and 2006. I complement existing studies by associating
the accounting benefits of using options to firms’ investor base composition and by documenting that changes are
more pronounced in firms with boards (pay-setters) with stronger reputational concerns.
This paper builds on the idea that favorable accounting for options affected boards’ perception of the cost of
options, presented in Murphy (2002) and Hall and Murphy (2003a). Unlike these authors, who view boards as
agents incapable of aptly valuing option grants, I conceive of boards of directors as the better-informed agents
who exploited any possibility to manage investors’ perceptions about firm profitability in order to maintain or
increase stock prices.27 Earlier studies on the accounting treatment for stock options and its effect on their use
find mixed evidence.28 More recently, using the implementation of SFAS123R as an exogenous source of variation,
the literature seems to accept the view that favorable accounting played an important role in the design of CEO
compensation (Hayes et al., 2012; Skantz, 2012; Ferri and Li, 2016). I advance this literature in two ways.
First, I establish a different timing for the elimination of the accounting benefits—stemming from the ability to
disclose and not recognize a compensation expense (accounting benefits)—of using stock option grants. This
result, in turn, is pertinent to studies that focus on the relation between the structure of CEO compensation and
managerial actions, particularly those that use SFAS123R as their exogenous shock to compensation practices
(e.g., Chava and Purnanandam, 2010; Hayes et al., 2012; Ferri and Li, 2016), since an anticipated elimination
27
My results suggest that managing financial reporting perceptions was one of the key drivers of the use of stock option grants in the
prescandal period. Graham et al. (2005) offer a pragmatic view of the practice of actively managing outsiders’ perceptions through reported
earnings: “The world is a complicated place, though corporate decision rules often are not. Executives often employ simple decision rules or
heuristics in response to a handful of widely held beliefs about how outsiders and stakeholders will react. These anticipated reactions are the
‘rules of the game’ that dictate the playing field for many earnings management and disclosure decisions.”
28
Among earlier work, Matsunaga (1995), Core and Guay (1999a), Carter et al. (2007) document evidence consistent with this view. By
contrast, Dechow et al. (1996), Yermack (1997), Bryan et al. (2000), Aboody et al. (2004a,b, 2006) ) are unable to document a relation
between compensation practices and financial reporting considerations.
For a comprehensive review of studies examining the relation between CEO compensation and financial reporting, see Aboody and Kasznik
(2009); for studies examining the use of stock options in CEO compensation in general, see Murphy (1999) and Core et al. (2003).
9
of the accounting benefits of using options raises doubts about the validity of their identification assumption if
they consider observations between the years 2002 and 2004 as part of their preshock period.29 (In light of this
evidence, in a subsequent paper I will reexamine the relation between CEO compensation and firms’ financing
and investment policies.) Second, I associate the composition of firms’ investor base to their perceived accounting
benefits of using stock option grants, which suggests that the value relevance of disclosed (but not recognized)
information varies in the cross section of firms.30
In addition, this paper contributes with empirical evidence that firms structure transactions (here CEO optionbased compensation) to achieve financial reporting objectives, consistent with survey evidence in Graham et al.
(2005).31 In this paper, I present evidence suggesting that firms rely heavily on the use of CEO stock option grants
to avoid a compensation expense against reported earnings with the ultimate objective of manipulating outsiders’
perceptions about firm profitability.
This paper also contributes to a small literature that studies the influence of media and public opinion on
corporate governance practices by analyzing the effect that the increased probability of being exposed in media
headlines has on the firm’s compensation practices.32 Finally, the evidence in this paper is related to a literature
that documents changes in reporting behavior in the period following the accounting scandals of 2001–2002.33
The remainder of the paper is organized as follows. Section 2 reviews background information and develop
the hypotheses tested. Section 3 discusses the identification strategy and describes the construction of the variable
used to proxy for the accounting benefits of using CEO stock option grants. Section 4 describes the data. Section 5
presents and discusses evidence of the decrease in the use of CEO option awards following the events set off by the
accounting scandals and documents its heterogeneous effect across firms. Section 6 presents the complementary
analysis that employs earnings discontinuities. Section 7 concludes.
2
Hypotheses Development
The economic cost incurred by a firm when awarding a stock option grant is equal to the value that outside welldiversified investors would be willing to pay provided that they could freely sell, swap, pledge as collateral, or
29
For instance, (Chava and Purnanandam, 2010) overcome this problem by running a model of changes over the period 2001 to 2005
–i.e.,(Y2005 − Y2001 ) = (X 2005 − X 2001 ) β + (" 2005 − "2001 ).
30
Relatedly, Yu (2013) documents that the value relevance of disclosed versus recognized information varies with the level of institutional
ownership in the context of pension liabilities.
31
See, e.g., Marquardt and Wiedman (2005) for evidence on firms structuring their convertible bond transactions to manage diluted earnings
per share and Choudhary et al. (2009), who document that firms accelerate the vesting of ESOs to avoid a charge against earnings for unvested
ESO grants after the implementation of FAS123R.
32
See, e.g., Dyck and Zingales (2002); Core et al. (2008); Joe et al. (2009); Kuhnen and Niessen-Ruenzi (2012).
33
See, e.g., Entwistle et al. (2006) ) for evidence that fewer firms do pro forma reporting, and when they do, they do it in a manner
that is less likely to mislead investors; and Lobo and Zhou (2006; 2010), who document an increase in conservatism following SOX, mainly
evidenced by lower discretionary accruals. Cohen et al. (2008) also document both a reduction in discretionary accruals after the passage
of SOX and an increase of real earnings management activities. Also related, Gilliam et al. (2014) documents the disapearance of the zero
earnings discontinuity in 2002.
10
hedge these options. For the risk-averse CEO receiving the grant, however, a stock option grant has a lower
valuation mainly for two reasons. First, the CEO is disproportionately invested in the firm he or she controls.
Second, existing restrictions limit the CEO’s ability to trade the options or short sell the firm’s stock.34 Therefore,
it should be expected that the firm’s board decides on a level of option-based compensation such that the economic
incentive benefits generated, that is, the increased firm performance, exceed the value-cost differential. Now, if
in addition to economic incentives benefits, the firm reports other benefits from using stock options as a means
of compensation, we should expect pay-setters to rely more heavily on this form of compensation, other things
being equal.
Under the accounting rules that prevailed in the early 2000s, the use of stock option grants received favorable
accounting treatment that differentiated them from other forms of compensation. Specifically, under FAS123,
compensating employees with stock option grants gave firms the ability to avoid taking a charge against earnings when a fixed number of stock options with an exercise price equal to (or exceeding) the grant-date market
price was employed, which was the most common practice during this period (see note 7). The perceived cost
hypothesis, proposed in Murphy (2002) and Hall and Murphy (2003b), suggests that this favorable accounting
treatment led directors’ perception of the costs of option awards to be much lower than their economic cost. The
authors argue that this view helps explain not only the features of compensation practices but also the escalation
of the use of stock option grants during this period.
Several papers have explored whether favorable option accounting treatment played a role in the design of
CEO compensation. Carter et al. (2007) find evidence that financial reporting concerns are positively related
to the use of CEO stock option grants. Yet the literature also includes studies that fail to document a reliable
relation between accounting consideration and the use of option grants. For instance, Yermack (1995) finds no
evidence that firms rely more on stock option grants whenever they face the high implicit costs of reporting low
profitability. Relatedly, Dechow et al. (1996) find no evidence consistent with the option accounting treatment
driving the decision to submit a letter opposing FASB’s 1993 Exposure Draft, which (unsuccessfully) proposed
the recognition of option expense. More, although Aboody et al. (2004a) find that firms voluntarily recognizing
an option expense under FAS123 have a smaller expense to report, they find no significant relation between the
likelihood of voluntary recognition and the magnitude of option expense (under fair-value methodology) after
controlling for other factors.
This inconclusive evidence has been recently reexamined following the implementation of FAS123R, which
requires all publicly traded firms to expense all types of option awards under the fair-value methodology, consequently leveling the playing field between (fixed) stock options and alternative equity-based forms of compensa34
See Hall and Murphy (2002), who model the divergence in the valuation of stock option grants from the perspectives of the firm and CEO
to rationalize several of the standard features of these contracts.
11
tion vis-à-vis financial reporting considerations. Recent work seems to settle this debate by providing convincing
evidence of a substantial modification of compensation practices as a response to the implementation of the modifying rule. In particular, Hayes et al. (2012) and Skantz (2012) both report that firms substantially reduce their
use of CEO stock option grants in favor of (full) stock awards, and they show that these changes were mainly
driven by the implementation of FAS123R, which, they argue, represented an exogenous shock to the noneconomic incentives to use stock options grants, which I refer to as accounting benefits.35 In this paper, however, I
contest the notion that the implementation of FAS123R in 2005 represented an exogenous shock to the accounting
benefits of using options. Instead, I argue that the accounting scandals of 2001–2002 (and the events following
the scandals) significantly and unexpectedly shocked the perceived accounting benefits of using options in 2002.
This reasoning implies that any study that obviates the effect that the accounting scandals had on the use of stock
option grants underestimates the role that accounting had on compensation policy.
Let us begin by considering the motivations that firms might have had to boost reported earnings using option
awards. The accounting literature compiles several motivations for managers to boost earnings even when there
exist no cash flow effects as in this case (see Healy and Wahlen, 1999; Dechow and Skinner, 2000; Fields et al.,
2001; and Graham et al., 2005). Based on this literature, I identify the following motivations for pay-setters, or
boards of directors, to exercise discretion over the structuring of CEO compensation to achieve a financial reporting objective such as boosting the reported earnings figures. First, pay-setters might consider that some investors
do not unscramble the cash-flow implications of the information contained in financial statements and therefore
perceive the ability to manage these investors’ perceptions about firm profitability (see note 13). Most certainly,
if this is the case, the ability to mislead investors must be a function of the composition of the firm’s investor
base (Hand, 1990). Second, from an alternative viewpoint, investors’ excessive focus on short-term performance
might force firms to take actions to inflate reported earnings (see note 14). For instance, it has been documented
that firms with a large proportion of transient (or short-term) investors experience large-scale selling by these
investors in anticipation of negative earnings-related news.36 Third, because most pay-setters (corporate directors) are managers or important decision agents in other organizations and are aware of market overreaction to
negative earnings-related news, they might have a preference for stock option awards over alternative forms of
compensation that entail a charge against earnings to avoid sending negative signals to managerial labor markets.37 Finally, pay-setters might have incentives to avoid a compensation expense to reduce the likeliness of
35
The switch towards stock awards seems to support the view that managerial incentives are more efficiently provided through restricted
stock than through options if companies are free to reduce existing forms of pay when making equity grants, in the absence of accounting
and tax concerns, see Hall and Murphy (2002). Dittmann and Maug (2007) also show that under the efficient contracting paradigm, and
ignoring accounting considerations, most CEOs should not hold any stock options. Instead, CEOs should have lower base salaries and receive
additional shares in their companies.
36
See Bushee (1998; 2001), who develops an institutional investors classification and identifies as “transient” investors those who trade
actively to maximize short-term profits. Using this classification, Ke and Petroni (2004) show that transient investors begin to dump the stock
when a break in the string of earnings increases is imminent.
37
Fama and Jensen (1983) suggest board members use their directorships to signal their competence as decision experts. Relatedly, Graham
12
violating earnings-based contractual covenants, thus reducing the expected cost of debt (see Watts and Zimmerman, 1986); or to manage the perceptions of other stakeholders (Bowen et al., 1995, Burgstahler and DIchev,
1997).
During 2002, several events with the potential to affect the accounting benefits of using option awards took
place—including a crisis in investor confidence, intense media coverage of corporations’ accounting and compensation practices, and the introduction of regulatory reforms. (For convenience, hereafter I refer to both the wave
of accounting scandals and the multitude of events that followed as the “scandals.”) For instance, the investor
confidence crisis—which resulted from the numerous accounting scandals that shook U.S. markets—increased
market participants’ focus on the quality of earnings and compensation practices, thus hindering firms’ abilities
to manage investors’ perceptions about firm profitability using reported earnings. Intense media scrutiny of the
scandals and corporate practices could also have played a role in correcting investors’ fixation on reported earnings. Related to this is a literature documenting a change in financial reporting behavior during this period and
associating it with the introduction of SOX. Entwistle et al. (2006), for example, find a sharp shift in firms’ pro
forma reporting behavior between 2001 and 2003; far fewer firms use pro forma reporting after the enactment of
SOX, and they do so in a manner that is less likely to mislead investors.38 In a similar vein, Lobo and Zhou (2006)
document more conservatism in financial reporting after the enactment of SOX, and Cohen et al. (2008) document
that accrual-based earnings management increased steadily until 2001 and decreased in 2002, coinciding with
the passage of SOX. Although the scandals likely affected boards’ perceptions of the accounting benefits of using
option grants, studies that analyze the design of CEO compensation during this period have largely overlooked
this point.39
Figure 1 offers a narrative consistent with the ideas put forth above. That is, changes in CEO compensation
ought to be a function of the timing of the effective modification of accounting benefits of using option awards.
Figure 1 shows that the most dramatic change in the use of stock option grants takes place between 2001 and
2002. Stock options usage further decreases after 2002, yet at a slower pace. Besides, the use of stock awards
increases gradually after 2002, whereas the use of cash payments seems to experience only a temporal incremental
change.
H1 Firms modified their compensation practices following accounting scandals in 2001–2002.
As reasoned above, I presume that firms modified their compensation practices proportionally to their perceived
et al. (2005) present survey evidence that CFOs believe that the inability to achieve certain reporting objectives might be interpreted by
managerial labor markets as a “managerial failure”; I conjecture that corporate directors hold a similar belief.
38
Under FAS123, firms that elected to not expense options were required to disclose the pro forma value of stock option grants in footnotes
of the financial statements.
39
One aspect that seems unlikely to have been affected by the scandals was the ability to manage the likelihood of violating earnings-related
contractual covenants. Most likely, firms enjoyed this motivation to use options until the adoption of FAS123R since debt contracts are usually
written using “floating GAAP” (Fields et al., 2001).
13
Figure 1: Composition of CEO Compensation
The figure displays the annual averages of the components of CEO compensation between 2000-2006. The sample consists of firms included in the ExecuComp
database in which there was no turnover with respect to the CEO in office during fiscal year 2001. Sample size varies from 938 firms in 2000 to 533 in 2006. The
components include cash pay (salary and bonuses), stock and option awards (valued using Black-Scholes-Merton approach), and other compensation (long-term
incentive plans, pension plans, and other compensation). Monetary amounts are converted to 2002-constant US dollars (USD) using the Consumer Price Index.
(firm-specific) accounting benefits.
H2 Firms that perceived higher accounting benefits of using options exhibited the largest modifications in their
compensation practices following the elimination of such benefits.
To see how the scandals affected CEO compensation practices, I explore the impact of the scandals on different
subsamples that are defined in order to determine the relevance of the explanations proposed above. The following hypotheses therefore help me assess whether the evidence is consistent with a revision of the perceived
accounting benefits of using options. The literature provides pervasive evidence that the probability of accurately
unscrambling the true cash-flow implications of accounting data is positively related to the sophistication of the
marginal investor (Hand, 1990). It has also been documented that institutional investors differ significantly in
terms of their investment styles and investment horizons (Fung, 1997) and that a large presence of institutional
investors with a short-term investment horizon—or transient investors, according to the classification proposed
by Bushee (1998; 2001)—exerts pressure to boost reported earnings. Based on this work, I conjecture that the
elimination of accounting benefits affected firms differently based on the composition of the investor base.40
Specifically, I hypothesize that firms with a large presence of unsophisticated (proxied by low institutional ownership) and transient investors were more likely to rely heavily on options before the scandals and therefore more
likely to have to correct their compensation practices afterward.
H3 Firms with a large presence of either unsophisticated or transient investors experienced larger decreases in
the use of options following the elimination of accounting benefits of using options than firms with a base
40
Closely related to this argument, Yu (2013) provides evidence that institutional ownership affects the valuation difference between
disclosed and recognized information in the context of pension liabilities.
14
of investors mainly composed of sophisticated investors with a long-term investment horizon.
Media provided intense coverage of the scandals that rocked U.S. markets in 2001–2002. Strong criticism of
corporate practices populated business media outlets throughout 2002, and blame was quickly assigned (see note
19). Corporate boards and their managerial compensation practices, particularly their use of stock option grants,
came in for heavy criticism. Other actors, chiefly academics and government officials, also participated in the
stigmatizing debate.41 These public concerns added to the perennial interest of regulators to improve governance
quality, placing corporate governance and its practices in the limelight.
Figure 2: Media Coverage - CEO Compensation
The figure displays the annual number of articles published in 5 major financial media outlets—The Wall Street Journal, The New York Times, The Washington
Post, Barron’s, and Investor’s Business Daily—relating to CEO pay. The bars (left-hand axis) represent the annual total number of articles that included the phrase
“CEO pay”, or variations of it. The broken line (right-hand axis) depicts the number of articles that included combinations of the following words and phrases:
“Enron” & “stock-option grants” & “ CEO pay”. The dotted line (right-hand axis) depicts the number of articles that included combinations of the following
words and phrases: “Expense/Expensing” & “Stock-Option grants” & “ CEO pay”. The solid line (right-hand axis) depicts the number of articles that included
combinations of the following words and phrases: “Board of Directors/Directors” “Excessive/egregious/lavish/scandalous” & “stock-option grants” & “ CEO pay”.
Source: Factiva.
Although the media’s focus on CEO compensation is a springtime tradition, in the spring of 2002 there was
a threefold increase in articles on this topic, as shown in Figure 2.42 In addition, negatively toned articles that
referred directly to the collapse of Enron, called for option expensing, and blamed corporate boards for what
pundits deemed lavish compensation packages represented a substantial portion of coverage.
These patterns beg the question of whether public opinion affects corporate governance. Several studies have
investigated these issues with mixed results. Core et al. (2008) find no consistent evidence that total compensation
declines after a CEO receives negative press coverage. In contrast, Dyck et al. (2008), Joe et al. (2009), and
Kuhnen and Niessen-Ruenzi (2012) report evidence that media do influences governance choices.43 Here, I
41
See, for example, Greenspan (2002).
This timing is due to the fact that the majority of public firms have fiscal years that coincide with the calendar year and typically release
their proxy statements around March following the end of the fiscal period.
43
Negative public opinion might have also played a role in attracting the attention of entities that may drive reforms; see Miller (2006) and
Dyck et al. (2010).
42
15
explore whether boards modified CEO compensation practices in a manner consistent with hypothesis that the
intense negative media coverage of 2002 played a relevant role. My objectives are twofold: first, to add to this
small literature, and second, to support the hypothesis that boards revised their perceived accounting benefits of
using options in 2002 well in advance of the implementation of FAS123R.
My strategy, like that of others in this area, follows the logic of Zingales (2000) and Dyck and Zingales (2002).
These authors build on the work of Fama (1980) and Fama and Jensen (1983), who argue that reputational concerns are the chief factor motivating corporate directors and who also note that directors’ reputation is a function
not only of their past performance but also of the system that transforms this behavior into public information,
which in this context is represented by the business media. Building on their logic, I conjecture that corporate
boards with strong reputational concerns should react more emphatically to the media’s stigmatization of the
use of option grants as a form of compensation due to the threat of being negatively exposed for not behaving
according to societal norms (in this case abiding to “socially acceptable” forms of compensation). Accordingly,
if the threat of being exposed in the media is insignificant, then we should observe no difference on pay-setters’
response of firms with differing strength of reputational concerns.
My strategy relies on two variables to identify firms with larger reputational concerns: firm visibility and
the board’s average age. I divide the sample into two groups according to the number of financial analysts
covering the firm, used to proxy for the firm’s probability of making it into business media “headlines” (see Brown
and Caylor, 2005). Taking as given the use of CEO option awards, a high level of analyst coverage for a firm
increases the expected cost of the “threat” of being exposed in public media as an incompetent director for using
a highly controversial form of compensation at the time. Importantly, I refrain from drawing any conclusion from
comparing firms with high and low levels of analyst coverage since the relation between analyst coverage and a
firm’s intent to boost reported earnings using options is arguably confounded by several factors.44
Next, I subclassify firms using the board’s average age. To the extent that the threat of being negatively exposed
in the media is relevant, “younger” boards should perceive this threat to be more expensive than “more senior”
boards because they risk losing more expected future income. On the other hand, if the threat is not relevant,
then there should exist no differential response between firms with young and senior boards.
H4 Younger corporate boards reduced more their use of option grants as a response to the scandals than more
senior corporate boards for firms with low analyst following.
44
On the one hand, a higher number of analysts following a firm might make it harder for firm officials to mislead its investor base. For
instance, Yu (2013) finds evidence suggesting that off-balance-sheet (disclosed) information is more value relevant for firms with high analyst
following; and Yu (2008) finds that firms with high analyst following manage their earnings less. On the other hand, other studies find
evidence suggesting some degree of alignment between firms’ management and analysts (see Michaely and Womack, 1999 and Chan et al.,
2007) and that high analyst coverage enhances the incentives to meet and beat earnings benchmarks (see, for example, Degeorge et al.,
1999). Furthermore, it is also well known that firms with a high number of analysts covering them tend to be much larger firms (see ?), which
also tend to exhibit a larger use of equity-based awards.
16
H5 Younger corporate boards reduced more their use of option grants as a response to the scandals than more
senior corporate boards for firms with high analyst following.
3
Research Methodology
3.1
Statistical Model: Identification Strategy
The wave of Accounting Scandals
I investigate the role of accounting in the use of CEO stock option grants through the lens of the wave of the
accounting scandals (and the events it set off), which served as the catalyst for the revision of the perceived accounting benefits of using options. Focusing on this episode that is mainly characterized by a revision of investors’
confidence on corporate financial reporting practices provides an opportunity to ameliorate concerns that the relation documented in studies between accounting factors and the use of option grants is confounded by other
explanations, for example, contracting theory rationales.45
Figure 1 exhibits the drastic shift from options that characterized this episode and that constitutes my “laboratory” to determine whether there exists a causal role of accounting in the design of CEO compensation. Yet
Figure 1 conceals a great deal of cross-sectional information that motivates my identification strategy. Although
the revision brought about by the accounting scandals affects all firms simultaneously, as argued above, it affects
firms differentially and proportionally to their prescandal level of perceived accounting benefits.
To capture the cross-sectional variation on the modification of CEO compensation practices, I employ a Differencein-Difference regression framework (DiD) adapted to exploit the differing intensity with which firms perceived
accounting benefits are affected following the scandals. For ease of presentation and following the treatment
effects literature, I refer interchangeably to the market-wide revision brought about by the scandals as the “treatment” and to the differing intensity with which firms are affected—that is, firm’s accounting benefits of using
stock options—as “treatment intensity.” This approach identifies the role of accounting in the use of CEO option
awards from comparing changes in the use of this form of compensation in firms that differed in terms of treatment intensity under the classical DiD “common trend” assumption—that is, in the absence of treatment, firms’
compensation practices would have followed common trends regardless of treatment intensity.
For the sake of presentation, let us start by considering a first-differences model with two periods: PRE and
45
It has been well documented that growth firms (Smith and Watts, 1992) and opaque firms (Demsetz and Lehn, 1985) rely more on
market-based incentive plans, such as stock option grants options, to remunerate CEOs due to the high cost of monitoring managerial actions. Because there also exists evidence that transient investors—which is (presumably) related to the level of accounting benefits of using
options—have a preference for growth firms (Bushee and Noe, 2000) and opaque firms (Maffett, 2012), one might be concerned that the previously documented relation between accounting factors and option use is driven by contracting considerations. The advent of the accounting
scandals ameliorates this concern because it is unlikely that this episode affected firms’ investment opportunity set or the complexity of their
operations.
17
POST treatment. In that case, changes in compensation practices are defined by the following linear model:
4 SOG i j = γ j + β · ABi + Γ · X iPRE + "i j
(1)
where 4SOGi j is the change in CEO stock option grants around treatment in firm i, operating in industry j; ABi is
the firm-specific perceived accounting benefits measured before treatment; and X iPRE is a set of firm controls also
measured before treatment. It is evident from (1) that the effect of accounting on the use of option awards, β,
is identified from comparing changes in the use of stock option grants in firms affected with differing treatment
intensity after controlling for firm-specific controls and industry fixed effects. This set of firm and industry controls
is selected to capture all sources of variation that would lead to differential time trends, thus strengthening the
common trend assumption.46 In particular, note that I include the pretreatment levels of firm characteristics and
not their changes around treatment due to the concern that firm characteristics are also affected by the treatment
and would then bias the estimated effect.47 The set of controls includes firm characteristics identified in the
literature as determinants of CEO option-based compensation such as firm size (proxied using sales), book-tomarket, annual stock return and its variance, indicators of cash-flow shortfall and dividend constraints.48 Note
also that the inclusion of industry fixed effects ameliorates the concerns that the estimated effect is driven by
shocks to particular industries.49
Although the first-differences model in (1) identifies the role of accounting in the design of CEO compensation,
I choose to harness the availability of longitudinal data and estimate a fixed-effects model for three reasons. First
and foremost, the availability of panel data permits me to assess the credibility of the identifying assumption, as I
discuss in a later section. Second, under the validity of the common trend assumption, considering a temporally
flexible model adds efficiency. Third, the availability of panel data permits assessing the evolution of the use
of CEO stock option grants. Documenting the evolution dynamics of CEO compensation practices allows me to
compare the importance of the events that take place during the period examined, chiefly the accounting scandals
and the implementation of FAS123R. The baseline models estimated are the following:
SOGi j t = αi + γ j t +
X
2001
βk · ABi · 1 t=k + Γ · X i,t
+ "i j t
(3)
k≥2002
46
Common Trend assumption:
E[4SOG 0i j |i, j, AB i = a b, X = x] = E[4SOG 0i j |i, j, X = x] = γ j + Γ · X i,PRE ; ∀ab ∈ AB, ∀x ∈ X ,
(2)
Equation (2) that potential nontreatment outcomes (denoted by the superscript 0) need to be independent of treatment intensity. Note that
this assumption allows to identify treatment effects even when treatment status is not randomly assigned, as long as all firms share temporal
trends; see Besley and Case (2000), Imbens and Wooldridge (2009).
47
This would be the case if innovations in managerial incentives induce changes in the firm’s financial and investment policies. See Angrist
and Pischke (2008) for more on the problems of including “endogenous controls”.
48
See Core and Guay (1999b) and references therein.
49
In particular, one might be concerned that the estimated effect is driven by the dot-com crash that principally affected the technology
and communication industries since it is well known that firms in these industries rely heavily on the use of stock options grants as means of
compensation, see Ittner et al. (2003).
18
2001
SOGi j t = αi + γ j t + β POST · ABi · 1 t≥2002 + Γ · X i,t
+ "i j t
(4)
where model in (4) is just a collapsed version of (3). In addition to the variables described above, I include a
firm-manager fixed effect, αi , that accounts for firm and CEO time-invariant traits; and industry-year fixed effect,
γ j,t , that accounts for industry-wide dynamics interplaying in the determination of executive compensation.50 To
PRE
avoid biasing the DiD estimates, I modify the set of firm controls, X i,t
, to prevent them from being subsumed in
the firm-manager fixed effect since they are measured before treatment and constructed to be time invariant (as
the perceived accounting benefits metric). Because this set of controls is included to adjust only for potentially
2001
differing trends absent treatment, I construct X i,t
to take nonzero values for the postscandal period, and zero
otherwise.51 In addition, to account for potential serial correlation in the error term, I present robust standard
errors clustered at the treatment level (that is, firm-manager level) as suggested by Bertrand et al. (2004) when
estimating DiD models with more than two periods of data. Finally, to ease the interpretation of all the estimates
presented below, the perceived accounting benefits is normalized by its sample median. Therefore, βk can be
interpreted as an estimate of the treatment effect (on the treated) by year k on a firm with a typical pretreatment
level of perceived accounting benefits, whereas β POST represents an estimate of the treatment effect (on the
treated) during the postscandal period on a firm with a typical pretreatment level of perceived accounting benefits.
One final word before moving on to describe the construction of the metric used to proxy for the firm-specific
perceived accounting benefits of using options. Note that the discussion presented in section 2 does not exclude
the possibility that firm-specific accounting benefits of using options were time varying; however, assuming that
these are time invariant, or at least a slow-moving variable, permits us to identify the role of accounting in the
use of option awards using the simple and intuitive empirical setting described above.52
3.2
Quantifying the accounting benefits of using stock option grants
Accounting motivations to use options, or perceived accounting benefits, are not directly observable. Quantifying
them thus requires the construction of a proxy. Existing studies, for instance, have used measures based on the
footnote-disclosed value of option expense—usually scaled by market value equity or total assets (see Hayes et al.,
2012; Skantz, 2012; Ferri and Li, 2016)—before the implementation of FAS123R to proxy accounting motivations.
Although such measures are certainly informative about accounting motivations, they also contain information
50
αi also controls for CEO time-invariant traits because I exclude observations after CEOs exit.
The exclusion of this set of controls does not affect my estimations, yet their inclusion results in more precisely estimated coefficients, as
expected.
52
This assumption has been made by studies that evaluate the role of accounting on the use of option grants (see Hayes et al., 2012; Skantz,
2012; and Ferri and Li, 2016).
51
19
on other nonaccounting motivations.53 In order to discriminate accounting from nonaccounting motivations, I
propose an strategy based on training a model that captures a firm’s motivations to use stock option grants (in
lieu of alternative forms of compensation) at different points in time as a function of variables defining the firm’s
economic environment.
Specifically, I train the model in two samples that differ most prominently by the existence of favorable option
accounting under FAS123. I choose the first sample, AC C + N ON , to occur before the advent of the scandals
such that the model estimates capture both accounting and nonaccounting motivations to use options. And I
choose the second sample, N ON , to take place after the implementation of FAS123R, which formally eliminated
the accounting-related motivations to use stock option grants, therefore the model estimates capture only nonaccounting motives. Accounting motivations become discernible on subtracting the predictions obtained in first
sample from those obtained in the second sample.
Two assumptions underlie this strategy. First, firms decide on the optimal proportion of option-based compensation based on their economic environment, where optimal compensation refers to the CEO contract that
maximizes shareholder wealth subject to legal and regulatory constraints. Second, nonaccounting motivations to
use option awards should not vary significantly between the two samples.
Following the literature that examines CEO equity-based compensation (see Smith and Watts, 1992; Garen,
1994; Yermack, 1995; Core and Guay, 1999b; and Aggarwal and Samwick, 1999), I model the optimal use of
option-based compensation as follows:
Š
€
b · Ci j t
l n 1 + SOG/T C ?i j t = Θ
(5)
where SOG/T C ? is the optimal proportion of CEO compensation paid in stock option grants, and Ci t is a vector of
firm and manager characteristics used to capture the firm’s economic environment.54 The tobit estimates of (5),
h
i
bAC C+N ON Θ
bN ON , that define the expected optimal proportion of option-based compensation in the presence
Θ
and absence of accounting benefits are shown in Table 1, Panel A. In panel B, I present descriptive statistics of
the variables used to calculate the estimates presented in Panel A; whereas Panel C and D, present descriptive
statistics of the computed Perceived Accounting Benefits of Options (AB). The construction of ABi is described in
the header of Table 1.
Columns 1 and 2, present estimates using all available information (limited by the availability of CEO compensation data in ExecuComp) during the periods 2000-2001 (AC C + N ON ) and 2010-2014 (N ON ). The coefficients
on firm size (sales) and market-to-book are positive in both samples, consistent with the notion that larger firms
53
For instance, Ittner et al. (2003) present evidence that the use of stock option grants can be partly explained by attraction (for new hires)
and retention motives.
54
I model the proportion of option-based compensation instead of its level for the sake of comparability across firms.
20
with more investment opportunities use larger proportions of equity-based compensation to motivate managers
to act in the shareholders’ interest due to the increased difficulty of monitoring managerial actions (Smith and
Watts, 1992). Note that the coefficient on size for the AC C + N ON sample is considerably larger than for the
N ON sample, which is consistent with larger firms offering more scope to gain financial reporting flexibility from
the use of option awards; whereas the difference in the coefficients on market-to-book suggests that incentive
alignment played a less important role in the use of option awards in the AC C + N ON sample. The coefficients on
tenure are negative suggesting that firms rely more on option grants to incentivize incoming CEOs and that the
use of option grants decay as tenure increases. Moreover, these coefficients ameliorate the concern that the use
of option grants during the prescandal period (AC C + N ON ) was driven by entrenched CEOs’ intention to extract
compensation rents, as suggested by advocates of the managerial power view (see, for example, Bebchuk et al.,
2002, and Bebchuk and Fried, 2004). The estimated coefficients on cash-flow shortfalls and dividend constrained
firms are positive during the prescandal period, consistent with the notion that firms facing high costs of reporting
low profitability shift the mix of CEO compensation towards options. As expected, this effect inverts after firms
are required to expense options. The coefficients of the variance of stock returns show a similar pattern, suggesting that riskier firms perceived a greater ability to manage outsiders’ perceptions about firm profitability by using
option grants to boost reported earnings. These results are consistent with the ideas proposed in section 2.55
In columns 3 and 4, I present estimates of model (5) using the same periods of time as in columns 1 and
2, respectively, but limiting my attention to those firms that I can observe in both samples, AC C + N ON and
N ON . By restricting my attention to these firms, I ameliorate the concern that the estimates in each sample differ
because of sample composition and not because of the absence of perceived accounting benefits of options. The
results reported in columns 3 and 4 are very similar to those found in columns 1 and 2, respectively. Moreover,
the correlation between the perceived accounting benefits of options computed using estimates from columns 1
and 2 , AB AL L , and those computed using estimates from columns 3 and 4, AB BALAN C E D , is over 94%. Because
coefficients in columns 1 and 2 are estimated more precisely than those in 3 and 4, I prefer to use the sample that
includes all available data instead of the balanced sample. In addition, in columns 5 and 6, I present a placebo
test in which I use postscandal data only. The placebo AC C + N ON sample (denoted by the superscript “PLAC”),
AC C + N ON P LAC , is defined by data from 2010 to 2012; and the placebo N ON sample, N ON P LAC , is defined
by data from 2013 and 2014. Although this does not show that the proposed proxy captures firms’ accounting
benefits of using options, it is comforting to find that estimates in column 1 and 5 are substantially different; and,
more importantly, that the correlation between AB AL L , and the perceived accounting benefits of options computed
using estimates from columns 5 and 6, AB P LAC EBO , is barely 8%.
55
I control for interest coverage and stock return (zero coefficients) to remain consistent with the literature; their exclusion does not affect
the calculations presented in Table 1 (untabulated).
21
22
ABi =
¨
1
2
0
s={2000,2001}
P
ma x
¦”
?,AC C+N ON
(SOG/T C )i js
?,N ON
− (SOG/T C )i js
—
,0
SOG
SOG
© ( /T C )i,2000 = ( /T C )i,2000 = 0
other wise
-0.001*
(-2.39)
-0.027**
(-3.36)
-0.076
(-1.38)
0.0000**
(3.71)
0.021**
(7.04)
0.174**
(27.43)
2,255
1,229
YES
YES
Observations
Number of firms
Industry FE
Year FE
Stock Return
2
F σRetur
n
CF Shortfall
Dividend Constr.
Interest Coverage
YES
YES
1,425
6,204
(-2.27)
(-4.08)
(-20.84)
-0.020*
-0.018**
-0.035**
Ln(Tenure)
(-0.01)
(7.12)
(17.83)
-0.000
0.026**
0.013**
(1.48)
(8.85)
(66.51)
0.026
0.026**
0.034**
Ln(Sales)
(25.18)
(2)
(1)
VARIABLES
0.126**
[2010-2014]
[2000-2001]
Market-to-Book
N ON
ALL
AC C + N ON
Panel A
0.02
0.10
0.25
0.048
0.001
-0.017
-0.013
0.030**
0.008
YES
YES
796
1,505
(1.88)
0.003
(18.76)
0.117**
(24.15)
0.183**
(7.51)
0.027**
(4.37)
0.000**
(-22.08)
-0.043**
(15.69)
0.013**
(51.55)
(3)
Diff.
[2000-2001]
AC C + N ON
YES
YES
796
3,573
(-1.32)
-0.015
(-0.35)
-0.008
(-0.02)
-0.001
(-3.35)
-0.033**
(-1.64)
-0.000
(-3.58)
-0.019**
(4.95)
0.023**
(5.85)
0.020**
(4)
[2010-2014]
N ON
BALANCED
0.018
0.125
0.184
0.06
0.000
-0.024
-0.01
0.01
Diff.
YES
YES
1,386
3,859
(-2.81)
-0.032**
(1.58)
0.035
(-0.33)
-0.022
(-2.89)
-0.028**
(-2.94)
-0.001**
(-4.74)
-0.027**
(6.72)
0.035**
(7.49)
0.027**
YES
YES
1,256
2,345
(0.49)
0.007
(-0.90)
-0.027
(-0.53)
-0.051
(-2.01)
-0.027*
(-0.56)
-0.000
(-0.62)
-0.004
(4.58)
0.022**
(5.08)
0.024**
(6)
[2010-2012]
(5)
N ON P LAC
[2013-2014]
AC C + N ON P LAC
PLACEBO
-0.039
0.062
0.029
-0.001
-0.001
-0.023
0.013
0.003
Diff.
Panel B presents the descriptive statistics of the variables used in the estimations presented in the first two columns of Panel A ( Sample ALL). Panel C presents average, median and standard deviation of Perceived Accounting
Benefits calculated using the estimates from the three different samples; and panel D presents their pairwise correlations. All variables are defined in Appendix A. ** denotes significance at the 1% level, ** at the 5%.
Then,
?,ω
bω · Ci js ; ω = {AC C + N ON , N ON } , s = {2000, 2001}
(SOG/T C )i js = Θ
Table 1 is restricted to ExecuComp firms with Compustat and CRSP coverage. In panel A, I present the tobit estimates of model 4 using 3 different samples. In sample ALL and sample BALANCE the AC C + N ON period is
defined by (firm-year) observations between 2000-2001; and the N ON period by observations between 2010-2014. These two samples differ from each other on that the first use all data available —limited by the availability
of CEO compensation data (ExecuComp)—while the second considers only firms found in both AC C + N ON and N ON periods. In sample PLACEBO the AC C + N ON period is defined by observations between 2010-2012;
and the N ON period by observations between 2013-2014. To calculate Perceived Accounting Benefits, I first calculate the expected optimal proportion of option-based compensation in the presence, (SOG/T C )?,AC C+N ON , and
absence, (SOG/T C )?,N ON , of accounting motivations, using the estimates in Panel A. I calculate these variables for years 2000 and 2001 to construct a pre-scandal measure of perceived accounting incentives.
Table 1: Quantifying the Perceived Accounting Benefits of using CEO Stock Option Grants
23
0.14
0.72
0
0.048
0.77
.00237
0.70
0.22
0.10
0.46
19.1
0.88
2.4
1.57
0.32
Std. Dev.
0.16
0.20
0.36
-0.02
0.38
20.7
1.88
1.66
7.52
0.16
0.31
-0.02
0
13.3
1.95
1.31
7.42
0.09
Median
0.44
0.26
0.08
0.48
19.6
0.84
1.21
1.62
0.20
Std. Dev.
POST (N = 6, 204)
Mean
0.235
ALL
0.242
Median
0.131
Std. Dev.
0.228
Mean
0.237
Median
0.133
Std. Dev.
BALANCED
0.94
0.08
AB P LAC EBO
1
AB AL L
AB BALAN C E D
AB AL L
0.14
1
AB BALAN
1
AB P LAC
Panel D:Correlation of different calculations of Perceived Accounting Benefits (AB)
AB
Mean
0.05
Mean
Panel C: Descriptive Statistics of different calculations of Perceived Accounting Benefits (AB)
Stock Return
.0119
CF Shortfall
2
F σRetur
n
0.305
Dividend Constr.
8.05
1.79
1.71
14.9
Ln(Tenure)
1.35
2.03
Market-to-Book
Interest Coverage
0.42
6.96
0.42
7.07
Ln(Sales)
Median
PRE (N = 2, 255)
SOG/T C
Mean
Panel B: Descriptive Statistics of variables used in Panel A (Sample ALL)
0.04
Median
PLACEBO
0.059
Std. Dev.
Figure 3 shows the distribution of the computed prescandal perceived accounting benefits of using options,
AB, in the sample described in Table (1). Aside from the concentration of firms (approximately 14% of them)
with zero accounting benefits, the distribution resembles a bell-shaped curve with mean and median of 0.23 and
0.4
CDF
0.6
10
0
0
0.2
5
Frequency (%)
0.8
1
15
0.24, respectively.56
0
0.2
0.4
0.6
0.8
Accounting Incentives
Percent
CDF
Figure 3: Distribution of perceived accounting benefits of using stock option grants
This figure shows the distribution of perceived accounting benefits of using option grants measured in 2001 before treatment in the sample described in table (1).
Finally, note that to calculate perceived accounting benefits, ABi , I use the pre-scandal expected (modeled)
proportion of option-based compensation, using the estimates from model (5), instead of the observed proportion.
I choose to construct ABi this way to avoid including idiosyncratic temporary shock in this variable, which is
critically important in this context since AB defines treatment intensity in my empirical setting. If AB contains
idiosyncratic temporary shocks unrelated to accounting motivation to use options, then the DiD estimator in (3)
and (4) is likely to overestimate the effect that the revision of accounting benefits of using options (treatment)
had on CEO compensation. I discuss this issue in more detail in section 5.3.
4
Sample Selection
My sample is defined by the availability of CEO compensation data between years 2000 and 2006, obtained from
the S&P ExecuComp database. I restrict my attention to firm-manager duplets for which I observe at least one
observation of data in both the prescandal period (2000–2001) and the postscandal period (2002–2006).57 I
56
The kurtosis of the distribution with and without zero values is 2.79 and 3.49, respectively.
I drop the observations after CEO exits because the value of stock option grants awarded to the incoming CEO are likely to respond to
motivations other than the favorable accounting of options.
57
24
merge CEO compensation data with different sources of data—including S&P Compustat, RiskMetrics Directors,
and Thomson Reuters Insider Filings—and the resulting sample, described in Table 2, consists of 942 firms (and
thus managers) that span 40 of the 43 industries in the Fama-French classification after excluding firms in the
financial and utility sectors, amounting to a total of 5,202 firm-manager-year observations.58 The source and
construction of all variables used in this study are discussed in Appendix A.
Panel A of Table 2 presents summary statistics of the variables used in this study for the pooled sample. The
Black-Scholes value of annual CEO stock option grants, the main dependent variable here, displays an average
value of $2.2 million and a median value of $0.8 million, which is consistent with the widely documented positive
skewness of CEO compensation (see Frydman and Jenter, 2010). Perceived accounting benefits, ABi ∈ [0, 1], the
metric capturing the differing intensity with which the scandals revised firms’ perceptions about the cost of option awards—constructed in section 3.2—exhibits similar average and median values, 0.24 and 0.23, respectively.
There is thus no reason to suspect that the results presented below (in section 5) are driven by a small group of
firms. Moreover, as can be seen in Figure 3, this measure exhibits a considerable dispersion of 0.13, which is
the feature of this metric that permits the identifications of the role of accounting in the use of CEO stock option
grants. Panel A also displays other firm characteristics included in the empirical models estimated in section 5.
This set of firm characteristics, measured in 2001, is calculated using different sources of data. For instance, I use
financial statement data from S&P Compustat to calculate variables used to control for CEO compensation dynamics. And to examine the heterogeneity of the baseline results, I sort firms with data coming from the Thomson
Financial 13F Institutional Holdings database (institutional ownership), Institutional Brokers’ Estimate System
(I/B/E/S) database (coverage of financial analysts), RiskMetrics Directors database (boards’ average age and
board independence), and classifications found in the finance and accounting literature (institutional investors’
investment horizons and corporate governance quality).59
Panel B of Table 1 report averages of the main dependent variable, CEO stock option grants, in separate
columns for the prescandal (2000-2001) and postscandal period (2003-2004), after excluding year 2002 from
the sample. In detail, I calculate the firm-level means over the pre- and postscandal period separately and then
average this calculation across firms. I present averages for the full sample and as well as averages by terciles of
perceived accounting benefits in order to appreciate the differing response to the treatment. All the differences
presented in Panel B are negative and statistically significant.
From the prescandal years to the postscandal period, the average annual CEO option grant decreases from
$3 million to $1.57 million, and the proportion of option-based compensation decreases on average from 44% to
58
Fama-French industries are collections of four-digit SIC industries constructed to represent broader industry categories
(http://mba.tuck.dartmouth.edu/pages/faculty/ken.french/ftp/Siccodes48.zip). My sample does not include firms that fall under the following classifications: agriculture, coal, and other.
59
Information on corporate directors comes from RiskMetrics, which was bought by Institutional Shareholder Services.
25
Table 2: Summary statistics for CEO stock option grants and its determinants
Panel A: Descriptive statistics (pooled sample)?
N
Mean
Std. Dev.
25 th Perc.
Median
75 th Perc.
938
0.23
0.13
0.15
0.24
0.31
CEO Annual Compensation - Levels ($M)
Stock Option Grants
5202
Restricted Stock Grants
5202
Cash (Salary + Bonus)
5202
Other
5202
Total Pay
5202
2.18
0.54
1.41
0.44
4.61
3.93
1.86
1.64
1.40
5.63
0
0
0.59
0.01
1.19
0.78
0
0.96
0.052
2.57
2.47
0.11
1.64
0.29
5.39
35.8
29.8
15.6
28.6
17.8
0
0
59.7
0.49
34.8
0
42.1
2.5
67.8
8.9
12497
14711
1.6
443
435
0.95
1060
1078
1.38
3120
3066
2.09
7.6
19.4
0.46
0.09
0.29
0.63
4
4.5
0
-0.05
0.29
-0.15
6
10.2
0
-0.01
0.54
0.09
11
30.6
1
0.03
0.79
0.36
0.17
0.14
4.1
7
2.6
1.2
0.55
0.23
55
4
8
2
0.68
0.32
59
7
9
3
0.79
0.42
61
13
11
3
Perceived Accounting Benefits
CEO Annual Compensation - Percentage (%)
Stock Option Grants
5202
Restricted Stock Grants
5202
Cash (Salary + Bonus)
5202
Other
5202
7.6
47.9
9.1
Firm and Manager Characteristics - Control Variables
Assets ($M)
5202
4352
Sales ($M)
5202
4403
Market-to-Book
5202
1.8
CEO Tenure (years)
5202
8.6
Interest Coverage
5202
17.7
Dividend Constrained
5202
0.30
Cash Flow Shortfall
5202
-0.01
2
F σRetur
5202
0.53
n
Stock Return
5202
0.17
Firm Characteristics - Sorting Variables
Institutional Ownership
688
Transient Ownership
688
Board’s Average Age
688
Analyst Coverage
688
G-Index
626
E-Index
626
0.66
0.33
58
9.2
9.4
2.6
26
25
5.2
Panel B: Pre- and Post- Scandals Averages†
Pre-Scandal
Stock Option Grants ($M)
Stock Option Grants (%)
3.0
43.2
Post-Scandal
Difference
1.57
28.9
-1.44**
-14.3**
Low Accounting Benefits ( AB LOW = 0.09; 274 firms)
Stock Option Grants ($M)
1.48
0.96
Stock Option Grants (%)
24.9
20.8
Medium Perceived Accounting Benefits ( AB M E D = 0.24; 273 firms)
Stock Option Grants ($M)
2.92
1.54
Stock Option Grants (%)
48
30
High Perceived Accounting Benefits ( AB H I GH = 0.37; 265 firms)
Stock Option Grants ($M)
4.66
2.22
Stock Option Grants (%)
57.2
36.5
?
†
-0.52**
-4.1**
-1.37**
-17.8**
-2.44**
-20.7**
All variables measured in monetary amounts are converted to 2002-constant U.S. dollars using the Consumer Price Index.
I calculate the firm-level means over the pre- and postscandal period separately and then average this calculation across firms.
See Appendix A for variable definitions.
29%. Moreover, the magnitude of these differences is increasing with the level of perceived accounting benefits,
which I explore in detail below.
5
Empirical Results
5.1
Main results
The main hypothesis examined in this paper is whether there exists a revision of firms’ perceived accounting
benefits of using options following the accounting scandals of 2001–2002 that leads to an adjustment of firms’
compensation practices, which in turn would imply that accounting considerations played a role in the design of
CEO compensation as suggested in the perceived cost hypothesis. Moreover, building on the reasoning of section
2, if there exists such a revision following the accounting scandals, then firms should modify their use of option
awards as a function of their prescandal level of perceived accounting benefits. I explore this joint hypothesis
using models (3) and (4). Note that (3) allows us to assess not only whether there exists a revision of firms’
perceived accounting benefits of using options but also, if so, when it takes place. The estimates of these models
are found in Table 3. In Panel A, I model the use of CEO option grants using the logs of their Black-Scholes value,
and in Panel B, I model their use using the proportion of annual CEO compensation paid in option grants. In
both panels, Columns 1–3 contain the estimates of the model in (4) and Columns 4–6 contain the estimates for
its dynamic version in (3).
For the estimates presented in columns 1-3 of Panel A (as well as Panel B), I exclude year 2002 from the
27
sample.60 The estimates reflect that, after controlling for firm-manager effects and industry-year dynamics, the
use of stock option grants, on a firm with a typical pretreatment level of perceived accounting benefits, decreased
by approximately 45% on average in the postscandal period. Column 2 adds a set of controls measured in the
2001
pretreatment period, X i,t
, just barely affecting the estimates of Column 1. These estimates also remain largely
unchanged with the inclusion of the levels of this set of control variables, X i,t , which I argued against in section
3.1 (see the discussion on “bad controls” in Angrist and Pischke, 2008). The coefficients in Column 5 of Panel A
indicate that after controlling for firm-manager effects and industry-year dynamics, the use of stock option grants,
on a firm with a typical pretreatment level of perceived accounting benefits, decreased by approximately 44% in
2002, following the accounting scandals (treatment); the use of option grants kept decreasing during the following
years, yet at a slower rate, to reach a total decrease of approximately 65% by 2006. Panel B depicts a qualitatively
similar picture to panel A. In short, had there been no treatment at all, a firm with a typical pretreatment level of
perceived accounting benefits would have exhibited a proportion of option-based compensation 23% larger than
that observed in the postscandal period (column 2).
In both panels, column 7 presents a test of the validity of the identifying assumption of the empirical setting.
The availability of more than one period of data before treatment permits testing whether the use of CEO stock
option awards was already changing—as a function of accounting benefits—before treatment by freely modeling
“lead” effects, (β2000 , β2001 ). I consider 2001 as the baseline of the comparison and is, thus, omitted. A nonzero
lead effect in the year 2000 have to be interpreted either as an anticipated revision of the perceived accounting
benefits of using options, perhaps catalyzed by the dot-com collapse; or as evidence that firms’ use of CEO option
grants do not share temporal trends (identifying assumption). In fact, an anticipated revision of the perceived
accounting benefits of using options is the critic made in this paper to extant studies analyzing the effect of
FAS123R on the use of CEO stock option grants. As can be appreciated, estimates of this lead effect are close
to zero and not statistically significant. This result seems consistent with the causal interpretation I give to the
results presented in columns 1-6 of Panels A and B.
These estimates suggest that the real inflection point in the temporal dynamics of CEO stock option grants is
the year 2002. As conjectured above in section 2, this is consistent with firms losing the ability to mislead their
investors by using option grants to boost reported earnings. The further decrease observed from 2005 to 2006 is
consistent with firms further and decisively revising perceived accounting benefits following the implementation
60
I exclude the year 2002 from the sample to avoid making unintelligible comparisons because variables are measured at fiscal year-end
but firms do not share the same fiscal year-end. Thus, depending on the fiscal year-end, some firms might have been more affected in 2002
than others because equity-based awards are known to follow a temporal pattern across the fiscal year—over 70% of equity-based awards are
granted within the first fiscal quarter.
Although this presumed differential response in 2002 represents an opportunity to exploit exogenous variation in changes of CEO compensation practices, there exists a major hurdle to such an analysis. Exploiting this variation demands knowing the date of the generalized revision
of perceived accounting benefits, which is difficult to determine given the multitude of events related to the scandals. A study investigating
the market reactions to this multitude of events is beyond the scope of this paper and, moreover, unnecessary to show that firms revise their
perceptions about the benefits of using options well in advance of the implementation of FAS123R.
28
Table 3: Changes in the use of CEO Stock Option Grants as a function of Accounting Benefits
Columns 1-6 of this table presents estimates of the following model
SOGi j t = αi + γ j,t +
X
2001
βk · ABi · 1 t=k + Γ · X i,t
+ "i j t
k≥2002
where, αi , represent firm-manager fixed effects that accounts for firm and manager time-invariant traits; γ j,t , industry-year fixed effects that account for industry
wide dynamics interplaying in the determination of executive compensation; ABi is the firm-specific perceived accounting benefits of using options, as calculated
2001
in section 3.2 ; X i,t
= X i,2001 · 1 t≥2002 is a set of firm controls—sales, market-to-book, CEO tenure, interest coverage, cash-flow shortfall, dividend constrained
dummy, annual stock return, and return variance— suggested in the literature as important determinants of the use of CEO stock option grants also measured in
2001 (before treatment).
Columns 7 presents estimates of the following “extended” model:
X
2001
SOGi j t = αi + γ j,t +
βk · ABi · 1 t=k + Γ · X i,t
+ "i j t
k6=2001
The availability of more than one period before treatment permits to assess the validity of the assumption that in the absence of treatment firms’ use of CEO
option awards follow “common trends”. Because the prescandal period (2000-2001) is defined by the absence of treatment, β2000 = β2001 = 0 represents a test
of the validity of the identifying assumption. Therefore, a non-zero effect would have to be interpreted as a violation of the common trend assumption. Note that
in this extended model β2001 = 0 by construction, since I consider 2001 values as the baseline of the comparison and ommit that dummy correspondingly.
All specifications include firm-manager and industry-year fixed effects. Standard errors presented in parenthesis are clustered at the firm’manager level. **
denotes significance at the 5% level, ** at the 10%.
Panel A : Dependent Variable: Ln(1 + S t ock Opt ion Gr ants)
AB POST
(1)
(2)
(3)
(4)
-0.455**
-0.464**
-0.458**
(-9.36)
(-8.58)
(-8.42)
(5)
(6)
(7)
-0.051
AB2000
(-0.85)
AB2002
-0.408**
-0.443**
-0.464**
-0.465**
(-6.37)
(-6.26)
(-6.37)
(-6.26)
AB2003
-0.410**
-0.440**
-0.437**
-0.462**
(-7.23)
(-7.17)
(-7.01)
(-7.26)
AB2004
-0.420**
-0.450**
-0.456**
-0.472**
(-6.22)
(-6.25)
(-6.22)
(-6.69)
AB2005
-0.525**
-0.556**
-0.559**
-0.577**
(-8.43)
(-8.70)
(-8.72)
(-8.78)
AB2006
-0.632**
-0.663**
-0.652**
-0.685**
(-9.72)
(-9.74)
(-9.42)
(-9.86)
YES
2001
Controls X i,t
NO
YES
YES
NO
YES
YES
Controls X i,t
NO
NO
YES
NO
NO
YES
NO
Observations
4,241
4,241
4,241
5,202
5,202
5,202
5,202
Adj. R2
0.567
0.572
0.588
0.568
0.572
0.587
0.572
(1)
(2)
(3)
(4)
(5)
(6)
(7)
-0.210**
-0.232**
-0.236**
(-10.27)
(-10.76)
(-10.61)
Panel B : Dependent Variable:
AB POST
SOG/T C
-0.028
AB2000
(-1.27)
AB2002
-0.162**
(-6.62)
(-7.01)
(-7.39)
(-7.41)
AB2003
-0.174**
-0.200**
-0.205**
-0.217**
(-7.07)
(-7.74)
(-7.72)
(-7.93)
AB2004
-0.196**
-0.222**
-0.226**
-0.239**
(-7.51)
(-8.25)
(-8.32)
(-8.70)
AB2005
-0.253**
-0.280**
-0.281**
-0.297**
(-8.95)
(-9.93)
(-9.83)
(-10.35)
AB2006
-0.294**
-0.319**
-0.318**
-0.337**
(-10.72)
(-11.38)
(-11.29)
(-11.85)
NO
YES
YES
NO
2001
Controls X i,t
NO
YES
YES
29
-0.188**
-0.204**
-0.206**
Controls X i,t
NO
NO
YES
NO
NO
YES
NO
Observations
4,241
4,241
4,241
5,202
5,202
5,202
5,202
Adj. R2
0.477
0.484
0.492
0.484
0.488
0.495
0.488
of FAS123R, consistent with the findings of Hayes et al. (2012), Skantz (2012), and Ferri and Li (2016). One
explanation for this progressive revision is related to the ability to reduce the probability of violating covenants
by boosting reported earnings (Watts and Zimmerman, 1990). Because contracts are generally written in floating generally accepted accounting principles (GAAP) (Fields et al., 2001), firms were likely to keep substituting
alternative forms of compensation for option-based awards until the implementation of FAS123-R to avoid violating earnings-based debt covenants. Also, if “other stakeholders”—such as current and potential employees,
customers, and suppliers—are less sophisticated than investors and remain fixated to reported earnings following
the scandals, then firms might still have perceived the ability to benefit from using options until the implementation of FAS123R (Bowen et al., 1995). The study of these rationales as motivations to alter reported earnings
remains largely unexplored due to the unavailability of good proxies for earnings-based covenants and for transactions terms with “other stakeholders.”
5.2
Results Heterogeneity
To gain further insight into why the accounting scandals resulted in such a substantial decrease in the use of CEO
stock option grants, I analyze the heterogeneity in firms’ responses to the scandals (treatment). Specifically, I
calculate separate treatment effects across subsamples designed to assess the relevance of different explanations
motivating firms’ use of option awards to boost reported earnings.
To accommodate the calculation of different treatment effects in my sample, I estimate the following “stacked”
model:
SOGi j t m = αi + γ j,t +
X
2001
2001
β POST,m · ABi · θm
· 1 t≥2002 + Γ · X i,t
+ "i j t m
(6)
m
where, in addition to variables already defined, I include (dummy) variables that identify different subsamples
based on firm characteristics measured in 2001 before treatment, θm2001 .61 Again, to account for potential serial
correlation that could lead to downward-biased standard errors, I cluster errors at the firm-manager level. This
“stacked” model easily permits testing for differences in responses across subsamples, which allows me to assess
whether the patterns observed in the data are consistent with the motivation in question.
Investor Base Composition
If firms’ ability to mislead investors through inflated reported earnings is a function of investors’ capacity to unscramble the cash-flow implications of financial statements, then firms with a large proportion of unsophisticated
investors (proxied by low institutional ownership) were more likely to use option awards with the objective of
misleading and thus more likely to modify their compensation practices after the scandals enhanced investors’
61
I define the subsamples using firm characteristics measured the year before treatment, t = 2001, to avoid endogeneity concerns. Also,
2001
note that θm
is subsumed in αi .
30
focus on the quality of financial reporting practices. Similarly, I also hypothesize that firms with a large proportion
of institutional investors with short-term investment horizon (or transient investors, according to the classification
in Bushee, 2001) were likely pressured to boost earnings and thus likely to avoid recognizing the compensation
expense by using option awards. If this is the case, we should also observe a more pronounced decrease in the
use of options in this type of firm following the market-wide correction of investors’ attention. Based on this conjecture, I divide my sample into three mutually exclusive subsamples: (1) firms with low institutional ownership;
(2) firms with high institutional ownership and high transient ownership; and (3) firms with high institutional
ownership and low transient ownership.
Estimates of 6 accommodated to inspect differencial responses in terms of firms’ investor base composition
are presented in table 4.
The results are consistent with the view that firms with an investor base largely dominated by either unsophisticated or transient investors relied more heavily on the use of option grants during the prescandal period
and thus had to reduce their use of option grants significantly following investors’ increased focus on the quality
of firms’ financial reporting—catalyzed by the scandals. This differential response is consistent with Yu (2013),
who presents evidence that institutional ownership affects the value relevance of disclosed (but not recognized)
information.
Boards’ Reputational Concerns
The accounting scandals brought directors’ compensation practices into the limelight, in particular their use of
stock option grants. Throughout 2002, influential regulatory voices and the business media severely criticized
what they deemed an excessive use of option grants and went as far as to blame the scandals in part on their
use. The intense scrutiny may have led some corporate boards to decrease more emphatically the use of CEO
option grants to reduce their risk of being portrayed negatively in the media for the use of a then-controversial
form of compensation. In particular, if negative media exposure represents a credible threat for directors and can
influence corporate governance decisions by pressuring boards to behave according to societal norms (Dyck and
Zingales, 2002), then boards of directors with strong reputational concerns that view as credible the threat of
negative exposure should implement the most pronounced reductions in the use of stock option grants.
To examine this conjecture, I sort the data using proxies for boards’ reputational concerns and firms’ likeliness
of being covered by the business media. I proxy a board’s reputational concerns using the average age of its
directors. And I use the number of financial analysts covering the firm to proxy for the firm’s probability of being
covered by the business media (see Brown and Caylor, 2005), which in turn defines the relevance of the threat
of having their reputation tarnished from the board’s perspective. Using these proxies, I divide the data into four
31
32
ln SOGi j t m = αi + γ j,t +
m={LI,H I−H T,H I−LT }
X
m
2001
2001
β POST
· ABi · θi,m
· 1 t≥2002 + Γ · X i,t
+ "i j t m
0.562
0.562
YES
3,816
YES
0.562
3,816
YES
(-3.98)
3,816
-0.41**
(-7.42)
-0.49**
(-2.09)
(-2.75)
(-6.87)
-0.52**
-0.23**
-0.26**
(-7.48)
(-2.76)
(-6.90)
-0.49**
-0.29**
-0.52**
Observations
Adj. R2
(-7.66)
(3)
(2)
Test
T = 0.25
IO = 0.65
T = 0.20
2001
Controls X i,t
H I−LT
AB POST
H I−H T
AB POST
LI
AB POST
AB POST
Baseline
(1)
-0.46**
(-0.94)
-0.08
(-1.16)
-0.11
Test
0.562
3,816
YES
(-2.62)
-0.28**
(-7.60)
-0.50**
(-6.56)
-0.50**
(4)
T = 0.20
(-2.48)
-0.22**
(-2.28)
-0.22**
Test
0.562
3,816
YES
(-4.10)
-0.40**
(-7.55)
-0.50**
(-6.56)
-0.50**
(5)
T = 0.25
IO = 0.60
(-1.22)
-0.1
(-1.11)
-0.1
Test
2001
where in addition to the variables described above; θ LI
is a dummy that takes values of 1 when institutional ownership, IO, is below IO in year 2001; θH2001
I−H T is a dummy that takes values of 1 when institutional ownership
is above IO and transient ownership (Bushee, 2001), T , is above T in year 2001; and, θH2001
I−LT is a dummy that takes values of 1 when institutional ownership, IO, is above IO and transient ownership is below T in year
2001. All specifications include firm-manager and industry-year fixed effects. Standard errors presented in parenthesis are clustered at the firm’manager level. ** denotes significance at the 5% level, ** at the 10%.
H I−H T
H I−LT
H I−H T
H I−LT
I present tests of equality of coefficients —β POST
= β POST
and β POST
= β POST
—to the right of each specification in columns 2-5.
This table presents estimates of the following “stacked” model
Table 4: Changes in the use of CEO stock option grants as a function of ABi by investors’ sophistication and investment horizon
Table 5: Changes in the use of CEO stock option grants as a function of ABi by the relevance of boards’ reputational
concerns
This table presents estimates of the following “stacked” model
ln SOGi j t m = αi + γ j,t +
X
m
2001
2001
β POST
· ABi · θi,m
· 1 t≥2002 + Γ · X i,t
+ "i j t m
m={H−Y,H−S,L−Y,L−S}
2001
where in addition to the variables described above; θH−Y
is a dummy that takes values of 1 when analyst coverage, AC, is above the sample median and the
2001
is a dummy that takes values of 1 when analyst coverage, AC, is above the sample median and the board’s
board’s average age is below AGE in year 2001; θH−S
2001
average age is above AGE in year 2001; θ L−Y
is a dummy that takes values of 1 when analyst coverage, AC, is below the sample median, and the board’s average
2001
age is below AGE in year 2001; and θH−Y
is a dummy that takes values of 1 when analyst coverage, AC, is below the sample median and the board’s average age
is above AGE in year 2001. All specifications include firm-manager and industry-year fixed effects. Standard errors presented in parenthesis are clustered at the
firm’manager level. ** denotes significance at the 5% level, ** at the 10%.
H−Y
H−S
L−Y
L−S
I present tests of equality of coefficients —β POST
= β POST
and β POST
= β POST
—to the right of each specification in columns 2-4.
Ý=7
AC
AB POST
Baseline
(1)
-0.46**
AG E < 54
(2)
AGE < 57
Test
(3)
AGE < 60
Test
(4)
Test
(-7.66)
H−Y
AB POST
H−S
AB POST
L−Y
AB POST
L−S
AB POST
-0.85**
-0.68**
(-4.20)
(-4.49)
-0.60**
(-4.68)
-0.53**
0.32*
-0.54**
0.14
-0.56**
0.04
(-7.24)
-1.66
(-7.50)
-1.04
(-7.66)
-0.33
-0.34**
-0.34**
(-3.55)
(-3.76)
-0.37**
(-4.93)
-0.39**
-0.05
-0.39**
-0.05
-0.39**
-0.02
(-6.72)
(-0.64)
(-6.77)
(-0.68)
(-6.63)
(-0.42)
2001
Controls X i,t
YES
YES
YES
YES
Observations
Adj. R2
3,816
3,816
3,816
3,816
0.562
0.562
0.562
0.562
mutually exclusive subsamples and then estimate the differential response across subsamples using specification
6. The results are presented in Table 5.
Consistent with the notion that media can influence boards’ decisions, in this case CEO compensation policy,
I find that firms with younger boards reduce their use of stock option grants more than firms with more senior
boards when the threat of being exposed in the business media is nonmarginal (high analyst following). As the
definition of a “young” board becomes more restrictive, the differential response becomes even more pronounced.
In Column 2, we can observe that younger boards (average age below 54) respond 60% more than more senior
boards. Also, interestingly, the results suggest that for boards of firms with a low probability of being covered by
the media, reputational concerns play no role in response to increased public scrutiny.
These results suggest that shaming may be an effective mechanism for policy makers to bring about desired
changes in corporate governance practices. Moreover, it seems that policy makers are well aware of the potential
effectiveness of shaming mechanisms, judging by the recent introduction of a rule that requires public companies
to disclose regularly the ratio of the CEO’s pay to that of the median pay figure for all other employees—which
33
many have interpreted as an attempt to shame firms into paying CEOs less. Also related to the effectiveness of
shaming mechanisms, Joe et al. (2009) document that the inclusion of a firm on Business Week’s worst board list
improves the independence of its board and reduces its level of entrenchment.
5.3
Placebo Tests
Because the construction of the proposed proxy for accounting benefits of options (AB) is based on CEO option
grants, there exists the concern that it captures prescandal temporary shocks unrelated to accounting motivations
to use options that would lead to an overestimation of the DiD estimates presented above. To address this concern,
I propose a placebo test in which I pretend that the revision of firms’ perceived accounting benefits of using options
take place just before 2000 instead of 2002. A nonzero effect on the treatment intensity variable in the postplacebo
treatment date should be interpreted as evidence that the proposed proxy for accounting benefits, AB, includes
transitory shocks. I estimate the following model to implement the placebo test:
SOGi j t = αi + γ j,t +
X
ς
1999
βk · ABi · 1 t=k + Γ · X i,t
+ "i j t
(7)
k≥2000
1999
where in addition to the variables described above; I include X i,t
= X i,1999 · 1 t≥1999 , which is similar to the set of
ς
firm controls introduced above with the difference that it is measured in 1999 before placebo treatment; and ABi
represents perceived accounting benefits accordingly measured before placebo treatment (1998-1999) using the
procedure presented in Table 1. Note that under the validity of the assumption that perceived accounting beneftis
(AB) is a time-invariant (or slow-moving) variable, this choice of “placebo treatment date” should still allow me
to capture the treatment effect from 2002 onwards if I estimate a dynamic model like the one proposed in 7. Also
note that this test is related to that proposed in column 7 of Table 3, which provides evidence consistent with (i)
no anticipated revision of perceived accounting benefits, and (ii) no violation of the common trend assumption.
Therefore a nonzero effect would have to be attributed to the inclusion of transitory shocks in the measure of
perceived accounting benefits.
ς
Table 6 presents the estimates of model (7), where placebo perceived accounting benefits (ABi ) have been
calculated under different assumptions (see the header of Table 6). The estimates in columns 2-4, reject the hypothesis that the methodology followed to construct our proxy of AB includes preplacebo (1998-1999) transitory
shocks that lead our estimations to “identify” a false effect in years 2000. This result represents additional evidence that the use of CEO stock option grants did not change as a function of AB until 2002, consistent with AB
being affected after the arrival of the wave of scandals not before. In column 1, I present estimate of the baseline
specification (model 3) for comparison purposes since the sample analyzed is considerably different from the
one analyzed in previous analyses. Data is restricted by the number of firms for which I can calculate ABi in the
34
Table 6: Changes in the use of CEO Stock Option Grants as a function of Accounting Benefits - Placebo Test
This table presents estimates of the following placebo model (columns 2-4), in which we pretend that treatment takes place in 2000 (instead of 2002):
X
ς
1999
SOGi j t = αi + γ j,t +
βk · ABi · 1 t=k + Γ · X i,t
+ "i j t
k≥2000
1999
where in addition to the variables described above; I include, X i,t
= X i,1999 · 1 t≥1999 , which is similar to the set of firm controls introduced above with the
ς
difference that it is measured in 1999 before placebo treatment; and ABi represents perceived accounting benefits measured before placebo treatment (1998ς
1999), using the procedure presented in Table 1. Each column of this table uses a different measure of ABi . These measures differ from one another in the
samples used to capture both accounting and nonaccounting motivations to use option grants, AC C + N ON ; the sample used to capture only nonaccounting
motivations to use options, N ON , remains unchanged across measures, 2010-2014. In column 2, the AC C + N ON sample uses data between years 1998-1999,
ς
ς
just before placebo treatment, to calculate ABi . In column 3, AC C + N ON sample uses data between years 2000-2001 to calculate ABi , under the logic that
ς
perceived accounting benefits are time-invariant. In column 4, the AC C + N ON sample uses data between years 1998-2001 to calculate ABi .
This test is designed to ameliorate concerns that the proxy for perceived accounting incentives, ABi , (calculated in Table 1) contains idiosyncratic temporary
shocks. If so, the identifying assumption is violated since firms with high ABi are expected to reduce the use of option grants even in the absence of treatement
leading to an overestimation of the treatment effect (see Ashenfelter, 1978). In this case, β2000 6= 0. In performing this test, I provide additional evidence that
CEO compensation practices do not change as a function of perceived accounting benefits until 2002 after treatment. In this sense, this test is similar to the one
conducted using the extended model estimated Table 3, column 7.
Column 1 presents estimates of the baseline specification for comparison purposes since the sample analyzed in this table contains a significantly smaller number
ς
of firms (559) relative to the sample analyzed in Table 3 (959). The sample is restricted by the availability of firms for which I can calculate ABi in column 2.
The data analyzed cover the period between 1998 and 2006. All specifications include firm-manager and industry-year fixed effects. Standard errors presented
in parenthesis are clustered at the firm-manager level. ** denotes significance at the 5% level, ** at the 10%. See Appendix A for variable definitions.
Panel A : Dependent Variable: Ln(1 + S t ock Opt ion Gr ants)
Baseline
(1)
AB2000
ACC+NON
ACC+NON
ACC+NON
[1998-1999]
[2000-2001]
[1998-2001]
(2)
(3)
(4)
-0.02
-0.09
-0.05
(-0.25)
(-1.17)
(-0.72)
0.07
0.01
0.04
(1.09)
(0.16)
(0.58)
-0.23**
-0.20**
-0.23**
-0.22**
(-2.11)
(-2.49)
(-2.71)
(-2.58)
AB2003
-0.24**
-0.17**
-0.20**
-0.19**
(-2.71)
(-2.42)
(-2.63)
(-2.38)
AB2004
-0.31**
-0.18**
-0.19**
-0.18**
(-3.55)
(-2.49)
(-2.60)
(-2.42)
AB2005
-0.42**
-0.29**
-0.33**
-0.31**
(-4.49)
(-3.75)
(-3.90)
(-3.70)
AB2006
-0.48**
-0.39**
-0.47**
-0.44**
(-4.98)
(-4.70)
(-6.07)
(-5.68)
AB2001
AB2002
1999
Controls X i,t
Observations
Adj. R2
YES
YES
YES
YES
4,052
0.541
4,052
0.540
4,052
0.540
4,052
0.540
35
preplacebo period (1998-1999), and also observe data after the real treatment date. As can be appreciated, the
estimates are significantly lower than those presented in Table 3 likely due to sample composition; and also partly
because the lower number of firms hinders my ability to control for industry-wide dynamics as precisely as done
above with the larger sample of firms.
5.4
Robustness Checks
In this section, I explore the robustness of the results presented above using different definitions of the control
variables employed, excluding firms that have been documented to experience changes on their use of CEO stock
option grants on ways that could be confounding my results. In addition, I explore whether the improved boardrelated rules imposed by the NYSE and Nasdaq or two requirements mandated by SOX can explain the above
documented changes in the use of CEO stock option grants.
Table 7 presents the results of these robustness checks. In column 1, I present the results of estimating (4), the
main specification, for comparison purposes. In column 2, I present estimates of a model that does not control for
industry-wide dynamics and only includes firm-manager and year fixed effects. Although the coefficient presented
in column 2 is considerably smaller than the coefficient in the baseline specification (column 1), it remains both
economically and statistically significant. I decide to include industry-wide dynamics in my main model because
the effect is estimated more precisely. Moreover, the inclusion of industry-wide dynamics ameliorates the concerns
that the estimated effect is driven by shocks to any particular industry. Column 3 and 4 modify the industry
definition to consider a more disaggregated level of industry classification. The results in these columns suggest
that there are no considerable gains from disaggregating SIC 2-digit further.62
In column 5, I exclude firms targeted with shareholder proposals to expense stock options between 2003 and
2004 in order to verify that shareholder activism does not explain my findings. The coefficient in column 5 is
barely affected relative to that presented in column 1, implying that the influence of shareholder proposals cannot
explain the effect documented above. An additional concern is that the results are driven by the a reduction of
the use of stock option awards in firms competing in the technology and telecommunications industries following
the dot-com collapse (see note 49). Although all the specifications proposed control for industry-wide dynamics,
these controls might not be sufficient. Estimates in column 6 reveal that the effect is still largely economically
and statistically significant after excluding these firms, suggesting that these firms’ compensation practices cannot
explain my results. In column 7, I include firm specific trends. The rationale for their inclusion is that these trends
might be correlated with the firm’s perceived accounting benefits and my estimate could be only reflecting this
relation. For instance, one might think that firms with higher perceived accounting benefits of using options in
62
The differences in sample size reflect the exclusion of industries in which I only observe one firm in my sample.
36
the prescandal period were already experimenting a decline. However, as can be appreciated, the inclusion of
firm-specific trends do not kill our coefficient, which remains highly significant.
In February 2002, Mr. Harvey Pitt, then SEC Chairman, publicly urged the NYSE and the Nasdaq to strengthen
their governance-related listing standards. In response, the exchanges proposed rulings that were approved with
minor revisions by the SEC.63 To mitigate concerns that the new board requirements of the major U.S. stock
exchanges drive my results, I control for the presumably differing response of firms that were not complying with
these requirements and firms that were already complying with them. Following Chhaochharia and Grinstein
(2009), I measure the level of compliance using three board structure variables affected by these rules: the
requirement for a majority of independent directors, and the requirements for fully independent compensation
and nominating committees. Specifically, I include in my specification the board structure variables interacted
with dummy variables for whether the year belongs to the period before the new rules or after the new rules.64
Table 7, columns 8 and 9, show the results of estimating the model that includes the three measures of compliance
where the announcement year is set to 2002 and 2003, respectively. I explore both scenarios since Chhaochharia
and Grinstein (2009) document that the structure of boards changed significantly as early as 2002. Regardless of
the choice of announcement year, the estimate of the accounting benefits remains virtually unchanged, suggesting
that my results are not driven by the new exchange rules. Moreover, the estimates of the levels of compliance
(untabulated) indicate that only the requirement to have an independent compensation committee had an effect
on the use of CEO stock option grants. This requirement seems to have increased the use of CEO option grants
on firms that were not complying before the announcement of the new rules, which is consistent with the results
documented by Guthrie et al. (2012).
The Sarbanes-Oxley Act: option backdating and section 404
One remaining concern is that the consequences of the regulatory changes induced by the enactment of the
Sarbanes-Oxley Act in 2002 might have curbed firms’ incentives to use CEO stock option grants for reasons different to the negation of the accounting benefits of using options. I identify one provision mandated by SOX
that could be obscuring my findings: the requirement to expense option grants within two business days. This
provision largely limited firms’ ability to backdate option grants vis-à-vis the pre-SOX period, in which firms had
up to 45 days after the company’s fiscal year-end to report option awards (see Heron and Lie, 2007). If firms
employed CEO option grants primarily either because it permitted extraction of compensation rents or to reduce
the cost of dilution to shareholders, then I should find that firms less likely to backdate options should not exhibit
63
NYSE submitted their proposed rule in August 2002, while Nasdaq submitted theirs in October of the same year. See the SEC press release
34-48745, November 4, 2003; https://www.sec.gov/rules/sro/34-48745.htm.
64
I limit my attention to U.S. firms that are members of the NYSE or Nasdaq. Following, Guthrie et al. (2012), I exclude from my sample observations of Apple Inc., Fossil Group Inc, and Oracle Corp. The results remain unchanged if I do not exclude these observations
(untabulated).
37
38
2001
SOGi j t = αi + γ j t + β POST · ABi · 1 t≥2002 + Γ · X i,t
+ "i j t
Adj. R2
Observations
Nasdaq enhanced requirements
5,202
0.574
5,202
0.571
YES
NO
YES
2001
Controls: X i,t
NO
YES
NO
NO
Controls: Compliance with NYSE and
YES
(-7.87)
(-9.58)
Year FE
-0.31**
-0.49**
Industry-Year FE
AB POST
(2)
Dynamics
No Industry
(1)
Baseline
4,828
0.541
NO
YES
NO
YES
(-8.26)
-0.48**
(3)
3-digit
SIC
4,502
0.521
NO
YES
NO
YES
(-7.00)
-0.47**
(4)
4-digit
SIC
4,806
0.557
NO
YES
NO
YES
(-9.28)
-0.47**
(5)
Activism
No Shareholder
4,355
0.587
NO
YES
NO
YES
(-7.73)
-0.40**
(6)
Communications
No Tech. and
5,202
0.627
NO
YES
NO
YES
(-3.61)
-0.34**
(7)
trend
Firm-specific
4,060
0.576
YES
YES
NO
YES
(-8.49)
-0.48**
(8)
4,041
0.568
YES
YES
NO
YES
(-7.87)
-0.49**
(9)
Requirements
Enhanced Exchanges
were announced. Column 8 considers 2002 as the announcement date, whereas columns 9 considers 2003 as the announcemen date.
All specifications include firm-manager fixed effects. Standard errors presented in parenthesis are clustered at the firm-manager level. Levels of significance: ** denotes significance at the 5% level, ** at the 10%. See
Appendix A for variable definition.
interactions of variables that define the level of compliance with the enhanced rules proposed by NYSE and Nasdaq and dummy variables for whether the year belongs to the period before the new rules or after the new rules
the fair-value methodology. Column 6 drops firms that belong to the Technology, and Communications industries, as identified in Murphy (2003). Column 7 includes firm-specific linear trends. Column 8 and 9 include
SIC 3-digit classification; whereas column 4 identifies industries using the SIC 4-digit classification. Column 5 drops observations I identify as being targeted by shareholders with the proposals to expense options following
where all variables are defined as above. Column 1 presents estimates of (4), the main specification, for comparison purposes. Column 2 drops industry control, that is, γ j t = γ t , ∀ j. Columns 3 identifies industries using the
Columns 1-6 of this table presents estimates of the following model
Table 7: Changes in the use of CEO Stock Option Grants - Robustness Checks
a significant reduction in their use of CEO option awards after 2002.65 I explore this hypothesis using the stacked
model proposed in 6, where the subsamples are defined by the likeliness of backdating options. Following Aboody
and Kasznik (2000), I consider firms with a fixed schedule for awarding options—i.e., firms whose options awards
have nearly identical dates every year—as unlikely to backdate options. The remainder of my sample is considered firms with a flexible schedule. I present the estimates of the treatment effect in both samples in Panel A of
Table 8.
Estimates in columns 2 and 3 present estimates under two different definitions of firms with fixed schedules.
I classify a firm as having a fixed schedule when it grants options under a fixed schedule every year between
1999 and 2001. In column 2 (3), I classify a CEO option award as being granted under a fixed schedule when
the grant date differs as much as by 5 (2) days relative to the grant date of the previous year’s grant. As can be
appreciated from columns 2-3 in Panel A, both types of firms exhibit both economically and statistically significant
reductions in their use of CEO option awards regardless of the classification employed. Although firms with a
flexible schedule grant reduced their use of options significantly more than those with a fixed schedule, I can
comfortably reject the hypothesis that backdating options is the only explanation behind my results.
In Panel B, I further explore the changes in the regulatory changes induced by SOX exploiting the fact that
section 404 of the Act was not required for all public firms. Section 404 required firms’ managements to present a
report on internal controls over financial reporting in their 10K filing and accompany it with an external auditor
attestation to their findings. Compliance of section 404 could have resulted in differential changes in the use
of CEO option awards for two reasons: (i) heightened oversight of the financial reporting process might have
disincentivized the use this form of compensation to mislead investors–a rationale in line with the elimination
of perceived accounting benefits of using options–, or (ii) increased compliance costs imposed by section 404
of SOX (Iliev, 2010)—whose provisions were considered by many experts the most costly to implement—could
have resulted in a general equilibrium reduction in the level of CEO option grants. Although finding a significant
difference in the modifications implemented by firms affected and nonaffected by section 404 would not allow
me to discriminate between these two explanations, not finding a statistical difference between these two types of
firms would provide further support to the rationales documented above—namely, an improvement of investors’
focus on corporate financial reporting, and pay-setters’ reputation concerns.
I test the effect of the provisions of section 404 on the use of CEO option awards using 6 after dividing
my sample into two groups defined by their need to comply with section 404. Both U.S. firms not classified as
accelerated filers—firms with a public float below $75 million—and foreign firms were exempted from complying
65
One might view the practice of backdating option grants as the result of managerial entrenchment. Under this view, powerful CEOs attempt
to extract compensation rents through difficult-to-assess forms of compensation or tactics. Alternatively, one might think of backdating as a
strategy to issue the least number of options to convey a pre-defined level of compensation (see Murphy, 2013).
39
Table 8: Changes in the use of CEO Stock Option Grants & the Sarbanes-Oxley Act
This table presents estimates of models that explore whether firms changed their use of CEO stock option grants in reaction to two features of SOX: the requirement
to report stock option grants within two business days; and the implementation of Section 404, which required public firms’ managements to present a report on
internal controls over financial reporting in their 10K filing, and accompany it with an outside auditor attestation to their findings.
To examine whether the results discussed till this point can be explained by the effective elimination of option backdating—a consequence of the requirement to
report option grant within two business days—, I estimate and compare treatment effects in two subsamples defined by the likeliness of backdating option grants.
I divide the sample into firms with fixed award schedules and firms with flexible award schedules; under the premise that firms that grant options the same date
every fiscal year, are less likely to manipulate the grant-date of the option award (see Aboody and Kasznik, 2000). These estimations are presented in Panel A.
Panel B explores whether either the sole need of having to design and report on mechanisms ensuring the accuracy of financial reporting, or the increase in
compliance costs associated with Section 404 of SOX affected the use of CEO stock option grants. To do so, I exploit the fact that Section 404 never affected firms
not classified as accelerated filers and only affected foreign firms starting on 2007. In Panel B, I focus on data from periods between 2000 and 2004 to procure a
See Appendix A for the definition of firms with fixed and flexible award schedules, and firms classified as accelerated and nonaccelerated filers. All specifications
include firm-manager and industry-year fixed effects. Standard errors presented in parenthesis are clustered at the firm’manager level. ** denotes significance at
the 5% level, ** at the 10%.
F LEX
FIX
AC C E L
N ON −AC C E L
I present tests of equality of coefficients —β POST
= β POST
or β POST
= β POST
—to the right of each specification in columns 2 and 3.
Panel A : Elimination of Option Backdating
(1)
AB POST
(2)
Test
(3)
Test
-0.49**
(-9.58)
F LEX
AB POST
-0.50**
-0.50**
(-9.89)
FIX
AB POST
2001
Controls X i,t
Observations
Adj. R2
(-9.90)
-0.39**
0.11**
-0.38**
0.12**
(-6.59)
(2.53)
(-6.30)
(2.54)
YES
YES
YES
5,202
0.571
5,202
5,202
0.572
0.572
Panel B : Implementation of Section 404 of SOX
(1)
AB POST
(2)
Test
(3)
Test
-0.48**
(-9.21)
AC C E L
AB POST
-0.48**
-0.49**
(-9.20)
N ON −AC C E L
AB POST
2001
Controls X i,t
(-9.30)
-0.61**
-0.13
-0.61**
-0.12
(-4.51)
(-1.06)
(-4.53)
(-0.99)
YES
YES
YES
Observations
4,944
4,944
5,031
Adj. R2
0.572
0.572
0.571
40
with section 404 at least until 2006.66 In column 2 of Panel B, I limit my attention to U.S. accelerated and
nonaccelerated filers. Both types of firms exhibit economically significant reductions in the use of CEO option
grants that are not statistically different. In column 3, I include foreign firms to the sample of nonaccelerated
filers, and my results remain practically unchanged. Considering only foreign firms in the nonaccelerated filers
sample yields similar results (untabulated). This evidence suggests that section 404 did not have a major role
in the declining popularity of CEO stock option grants in the post-SOX period. Admittedly, even when I am not
able to find a significant effect of section 404 on firms’ compensation practices, any of the remaining sections of
SOX could have contributed to disincentivizing the use of CEO option grants. However, because these sections
did not exonerate any firm, it is complicated to assess their effect. One point worth emphasizing is that increased
potential legal liabilities imposed by SOX to the firm’s management unlikely affected the use of stock option grants
since avoiding a compensation charge against reported earnings under FAS123 was legal, as discussed above.
6
CEO Compensation Design as part of the Firm’s Earnings Management
Strategy
To this point I have argued that, before 2002, firms used stock option grants, in addition to other more extensively
documented tactics, such as accrual-based maneuvers and real actions earnings management, to boost reported
earnings to obtain higher valuation by shareholders. In this section I document this behavior and the subsequent
decline in the use of this form of compensation to boost reported earnings following the scandals of 2002 without
relying on the construct of accounting incentives proposed above.
The analyses presented here are based on a strand of the earnings management literature that interprets
the discontinuities in earnings distributions at key benchmarks as consistent with the view that managers act to
meet or beat earnings benchmarks. Hayn (1995) interprets the discontinuity around the distribution of reported
earnings as evidence that firms engage in earnings manipulations to help them cross the “red line,” and subsequent
studies examine the distributional properties of reported earnings around key earnings benchmarks, including:
loss avoidance (Burgstahler and DIchev, 1997; Degeorge et al., 1999; Leuz et al., 2003), prior performance
(Burgstahler and DIchev, 1997; Degeorge et al., 1999; Donelson et al., 2013), and analysts’ consensus forecasts
Degeorge et al., 1999; Burgstahler and Eames, 2006; Donelson et al., 2013). In addition, several studies present
survey evidence that complements this interpretation. For instance, Dichev et al. (2013) report that 99.4% of
66
U.S. firms with a public float–common stock held by nonaffiliates–below $75 million, also known as nonaccelerated filers, were exempted
to file both the management report and the external auditor report until their fiscal year 2006. For their fiscal year 2007, nonaccelerated
filers had to include the management report with their 10-K, but remained exempted from accompanying this report with an external auditor
attestation. Similarly, “foreign private issuers” (foregin firms) were also exempted from filing both reports until their fiscal year 2005. From
fiscal year 2006 onward, foreign firms with a public float above $700 million had to file both reports, while foreign firms with a public float
under $700 million remained exempted from filing the external auditor report until fiscal year 2007. No foreign firm in my sample has a
public float under $700 million for all years between 2002 and 2006.
41
CFOs surveyed believe that other CFOs manipulate earnings, and Graham et al. (2005) report that managers
are willing even to sacrifice economic value to manage financial reporting perceptions and avoid the negative
economic consequences of missing an earnings benchmark. These empirical patterns are economically rational
when firms stand to benefit significantly from meeting or beating an earnings benchmark or, conversely, when
firms stand to lose from missing just one of these benchmarks.
I conjecture that if firms perceived the ability to meet or beat earnings benchmarks by substituting other
forms of compensation with stock option grants, then firms close to missing an earnings target are likely to
exhibit greater proportions of option-based compensation. In addition, since I argue that firms lose the ability
to manage financial reporting perceptions using stock option grants following the scandals, I expect to observe a
differential use of option grants before and after 2002 (scandals) on firms close to missing an earnings benchmark.
Specifically, before 2002, I expect that firms with premanaged earnings just below the earnings benchmark should
rely heavily on option grants to compensate their CEOs. In contrast, after 2002, firms with premanaged earnings
just below the earnings benchmark should not be able to use option grants as part of their earnings management
strategy.
In this analysis, I focus on the incremental incentives to manipulate earnings around the prior performance
earnings benchmark.67 I find that, before the accounting scandals, firm-year observations suspected of manipulating earnings to meet or beat this benchmark feature high levels of option usage, consistent with the view
that firms used stock option grants as part of an earnings management strategy. This differentiated use of option
grants vanishes in the postscandal period. In contrast, in untabulated results, I document that the two other
benchmarks considered by the literature to encourage earnings management, namely, analysts’ consensus forecast and loss avoidance, have no significant effect on the design of CEO compensation. The absence of an effect
around these two other benchmarks comes as no big surprise for two reasons. First, boards design CEO compensation long before analysts issue their final forecasts, and analysts continuously revise their forecasts until shortly
before earnings are announced. Second, unprofitable firms likely perceive lower benefits of meeting an earnings
target and face higher scrutiny of their reporting practices, which increases the costs of manipulating earnings.68
In fact, in untabulated results, I find that compensation practices of unprofitable firms are comparable to those of
poorly performing firms (see Gilson and Vetsuypens, 1993; Kaplan, 1994; and John and John, 1993).
67
Studies that document the economic significance of this earnings benchmark include Burgstahler and DIchev (1997), who report evidence
suggesting that 8% to 12% of the firms with small premanaged earnings decreases exercise discretion to report earnings increases; (Degeorge
et al., 1999) also report empirical evidence that prior performance benchmark is one of the main drivers of earnings management; and Graham
et al. (2005) present survey evidence that CFOs consider this benchmark the most important when reporting earnings.
68
Burgstahler and Chuk (2015) elaborate a similar argument about firms in financial difficulty. Moreover, consistent with this view,
Burgstahler and Dichev (1997) document that there is less evidence of discontinuities in earnings levels and changes among firms with a
weaker record of recent profitability.
42
Figure 4: Distribution of Earnings Innovations, ∆E PS t = E PS t − E PS t−1
Relative frequency of firm-year observations by earnings innovations intervals of width $0.05 over the range -2.5 to +2.5. Earnings per share, EPS, is calculated
as the ratio of income before extraordinary items adjusted for common stock equivalents over fully diluted common shares after adjusting for stock splits (see
Appendix A). Suspect firm-years is a subset of the observations grouped in the first interval immediately to the right of the red vertical line. The figure is truncated
at the two ends to include approximately 90% of the sample.
The distribution of earnings innovations
Under the presumption that firms have no ability to manipulate earnings, it seems reasonable to expect that, for
every type of firm, the probability of lying just below the prior performance benchmark is equal to the probability
of lying just above that same earnings benchmark. This assumption implies that the conditional density function
of premanaged earnings per share (EPS), f E PS|T Y P E (eps|t y pe), is continuous in EPS and further implies that the
unconditional density of earnings innovations, f E PS (eps), is also a continuous function.
To test for the (dis)continuity of the density function of earnings innovations around the prior performance
benchmark, I borrow from the program evaluation literature methods developed to evaluate the validity of regression discontinuity designs (RDD)—namely, the continuity of the density of the “assignment” variable at the
discontinuity point.69
In so doing, I diverge from the method traditionally employed in the earnings management literature that
relies on tests of smoothness in histogram-based distributions such as the one depicted in Figure 4.70 Figure 4
plots the distribution of earnings innovations, ∆E PS, measured using fully diluted split-adjusted EPS.71,72 The
tests proposed here represent a significant improvement in comparison to the smoothness criterion typically tested
in this strand of the literature.73
In Figure 5, I present a graphical inspection of the (dis)continuity of the density function around the prior
69
See Imbens and Lemieux (2008), Lee and Lemieux (2010)for references to tests for discontinuity in distributions and other significant
contributions in this field.
70
One notable exception is Bird et al. (2016), who employ techniques similar to those proposed here to analyze the discontinuity of earnings
surprises relative to analysts’ consensus forecasts.
71
his distribution is similar to those previously reported in the literature. See, e.g., Burgstahler and Dichev (1997), fig. 1, and Degeorge et
al. (1999), fig. 6.
72
See Appendix A for the definition of earnings per sahre (EPS).
73
Since Burgstahler and Dichev (1997), the vast majority of studies assess the statistical significance of the discontinuity using a t-statistic
that equals the difference between the actual and expected proportion of firms around the pertinent earnings benchmark divided by the
estimated standard deviation of this difference.
43
(a) Pre-2002
(b) Post-2002
Figure 5: Distribution of 4E PS - McCrary Discontinuity Tests
This figure present the densities of earnings innovations estimated using the methodology proposed by McCrary (2008). Panel A depicts the density using data
between years 1999-2001. Panel B depicts the density using data between years 2003-2006. The two densities have been estimated using data between the 10th
and 90th percentile.
performance before and after 2002 as proposed by McCrary (2008).74 Panel A of Figure 5 succinctly reveals the
sharp discontinuity in the density function in the pre-2002 period and permits the rejection of the null that the
density functions are equal to the left and right of the prior performance benchmark. In contrast, Panel B does not
provide statistical evidence of the discontinuity in the post-2002 period, providing further support to the view that
firms lose the ability to manage earnings. This evindence is consistent with the results of Gilliam et al. (2014),
who document that the discontinuity around the loss avoidance benchmark dissapears soon after the passage of
SOX in 2002.
Stock Option Grants Around the Prior Performance Benchmark
Having documented evidence suggesting that firms manipulate earnings to appropriate the benefits associated
with meeting or beating the prior performance benchmark, I now investigate the role of the design of CEO compensation as a tactic employed to achieve this objective. Simultaneously, I attempt to corroborate the effect
of accounting considerations on CEO compensation and the subsequent elimination of this effect following the
scandals in 2002.
To better understand the connection between the notion of accounting incentives developed above and the
prior performance benchmark, let us think about the benchmark in terms of accounting incentives, as defined in
section 2. In my setting, from the standpoint of a firm with premanaged earnings just below the prior performance
benchmark, the benefits that accrue to the firm from meeting or beating the benchmark correspond to a scenario
74
The graphical test shown in Figure 5 operates in two steps. First, I plot an undersmoothed histogram. Second, I smooth the histogram
using separate local linear regressions on both sides of the earnings benchmark, in which the normalized frequency of observations falling
into the bins are the outcome variable and the midpoints of the histograms are the regressors. I follow the procedure suggested in McCrary
(2008) to choose the bin size needed to plot the histogram and the bandwidth and kernel function required to implement the local linear
regression. I use the code provided in Justin McCrary’s website to generate Figure 2 (http://eml.berkeley.edu/~jmccrary/DCdensity/).
44
in which the firm faces a high level of accounting incentives when designing compensation. To quantify the effect
that these “benchmark-related accounting incentives” had on the design of CEO compensation, I need to compare
the compensation practices of firms that presumably beat the benchmark using stock option grants under two
different accounting-incentives-related regimes. The first is a regime in which the firm suspected of beating the
benchmark using stock option grants faced benchmark-related accounting incentives. The second regime is one
in which the firm faced no benchmark-related accounting incentives, which is unobserved. I approximate this
counterfactual scenario using firms whose premanaged earnings would have exceeded the prior performance
benchmark and controlling for different firm characteristics that are likely to determine both CEO compensation
and the cost of manipulating earnings.
I restrict my attention to firms facing a more favorable earnings scenario because I expect both firms with
benchmark-related accounting incentives and those without benchmark-related accounting incentives to coexist
just below the earnings benchmark, therefore hindering my ability to identify the benchmark-related incentives
to use CEO stock option grants.
It is straightforward to rationalize the existence of firms without incentives to beat the prior performance
benchmark. As argued above, heterogeneity on the ability to boost reported earnings must have made it prohibitively costly for some firms to meet or beat the benchmark. Otherwise, all firms just below the benchmark
would have found it profitable to beat the prior performance benchmark, and the previous analysis would have
revealed zero density immediately to the left of the benchmark, which is clearly not the case, as shown in Figures
4 and 5. The argument needed to rationalize the existence of firms with benchmark-related accounting incentives
to the left of the benchmark, however, is subtler. It relies on the fact that earnings manipulation using CEO compensation design required forecasting end-of-the-year premanaged earnings at the beginning of the fiscal period,
when the bulk of CEO compensation is determined.75 Under these conditions, it is to be expected that some firms
with benchmark-related accounting incentives were forced to miss a benchmark because of unexpected negative
shocks in their earnings process, even after boosting reported earnings by option grants. Ultimately, the problem with the coexistence of these two types of firms is that negative shocks are unobservable, which makes it
impossible to discriminate among them.
To minimize the possibility of misclassifying suspect firms, I develop EPS measures considering the firmreported fair value of employee stock option grants. Specifically, to classify a firm-year observation as suspect,
first, I require that ∆E PS ≥ 0 and ∆E PS < ¢, where ¢ is a threshold measured in cents of USD.76 And second,
75
An examination of Thomson Reuters Insider Filings database reveals that, on average, firms award over 70% of the value of equity-based
awards within the first quarter of the fiscal year.
76
I consider several thresholds within the range of 1 to 5 cents of USD. As expected, the lower the threshold the higher the magnitude of
the estimated coefficients.
45
Table 9: The use of CEO stock option grants as a tactic to meet-or-beat the recent performance benchmark
This table presents estimates of the following model:
SOGi t = αi + θi t + γ j t + δ0 SUSPi t + δ post−2002 SUSPi t · 1 t≥2002 + Γ · X i,t + +"i t
where SUSP EC T takes values of one for those firm-year observations classified as suspect, and zero otherwise; αi and θi t, represent a firm-manager fixed
effect and a firm-manager time trend, respectively; γ t , represents a industry-year fixed effect; and the set of controls is the same as described before. In
this model, δ0 captures the pre-2002 incentives to use option-based compensation as part of an earnings management strategy to sustain recent performance,
∆E PS = E PS t − E PS t−1 ≥ 0. The post-2002 incentives are given by the sum of δ post−2002 + δ0 . The analyzed sample is composed only of suspect and non-suspect
observations; there are 45 firm-observations classifed as suspect for ¢ = 1.5; 60 firm-observations classifed as suspect for ¢ = 2; and there are 78 firm-observations
classifed as suspect for ¢ = 2.5. Standard errors presented in parenthesis are clustered at the firm-manager level. Levels of significance: ** denotes significance
at the 5% level, ** at the 10%.
SOG/T C
Ln(SOG)
SUS P EC T0
SUS P EC Tpost−2002
Controls
¢ = 1.5
¢=2
¢ = 2.5
¢ = 1.5
¢=2
¢ = 2.5
(1)
(2)
(3)
(4)
(5)
(6)
0.98**
0.84**
0.80**
0.32**
0.31**
0.29**
(2.49)
(2.50)
(2.56)
(2.91)
(3.11)
(3.45)
-0.97**
-0.81**
-0.54
-0.23*
-0.20
-0.14
(-2.24)
(-2.02)
(-1.60)
(-1.67)
(-1.60)
(-1.40)
YES
YES
YES
YES
YES
YES
Observations
2,209
2,201
2,197
2,209
2,201
2,197
Adj. R2
0.532
0.546
0.540
0.462
0.475
0.472
that
E PS t − Opt ions C ost t/Outst anding shar es t < E PS t−1
(8)
which implies that this firm would have missed the target if it had to take a charge against earnings for stock option
grants77 Otherwise, my classification would include firms that would have beaten the target even if accounting
for options expense.78
To define the nonsuspect sample, as reasoned above, I limit my attention to firm-year observations with earnings innovations above ¢, ∆E PS > ¢, that would have reported EPS higher than the previous year even if expensing the fair value of employee stock options. This additional requirement is represented by (9) below.
E PS t − Opt ions C ost t/Outst anding shar es t > E PS t−1
(9)
By doing this, I avoid classifying as no-suspect firms that used option-based compensation to dramatically inflate
their EPS reported figures.79
77
I use fully diluted EPS. Several studies findfind evidence suggesting that investors place more weight on diluted EPS as a measure of
performance (see Core et al., 2002b) and that executives’ decisions are driven by diluted EPS but not basic EPS (see Bens et al., 2003).
78
Admittedly, this requirement potentially drops valid suspect observations because I am unable to distinguish between firms with no targetrelated incentives and those firms that opportunistically manipulate the inputs used to calculate the reported fair value of stock option grants
(see Aboody et al., 2006; and Hodder et al., 2006) to beat a prior performance target after accounting for options as an expense. This specific
type of opportunistic behavior does not seem to be a frequent phenomenon in my sample.
79
Botosan and Plumlee (2001) document that the effect of expensing stock options on EPS would have been substantial. In their sample they
find that it would have decreased EPS 14% on average. Dichev et al. (2016) report survey evidence that the magnitude of misrepresentation
in earnings may be as large as 10% of the realized earnings.
46
Having defined my sample—suspect and nonssupect observations— I proceed to assess the effect of the recent
performance benchmark on the use of CEO stock option grants, before and after 2002. I do so using the following
model:
SOGi t = αi + θi t + γ t + δ0 SUSP EC Ti t + δ post−2002 SUSP EC Ti t · 1 t≥2002 + "i t
(10)
where SUSP EC T takes values of one for those firm-year observations classified as suspect, and zero otherwise;
αi and θi t, represent a firm-manager fixed effect that accounts for firm and manager time-invariant traits and a
firm-manager time trend, respectively; and γ t , represents an industry-year fixed effect that accounts for industrywide dynamics interplaying in the determination of executive compensation. In this model, δ0 captures the pre2002 incentives to use option-based compensation as part of an earnings management strategy to sustain recent
performance, ∆E PS = E PS t − E PS t−1 ≥ 0. The post-2002 incentives are given by the sum of δ post−2002 + δ0 .
Table 9 presents estimates of (10). In column 1, when the threshold defining suspect firm-year observations
is 1.5 cents of USD, after controlling for a rich set of fixed effects and control variables, I observe that before 2002
firms suspect of meeting or beating the recent performance benchmark almost double their use of stock option
grants relative to their use of CEO stock options had they been comfortably above the threshold. In contrast,
this effect disappears from 2002 onwards. The result attenuates slightly as the threshold becomes less strict and
incorporate more firms as suspect (columns 2 and 3). Columns 4-6 present estimates when the dependent variable
is the proportion of CEO compensation paid in stock option grants. These coefficients follow a similar pattern,
although they are estimated with less precision; in particular, the reversal of the effect. This evidence suggests
that acute boards, aware of the negative consequences of missing the recent performance target, decided to rely
more heavily on stock option grants to improve the firm’s chances of meeting the target before 2002. It seems
boards lose this ability in the period following the accounting scandals, consistent with the evidence documented
in section 5.
7
Conclusions
Several papers have analyzed the intriguing shift away from stock option grants as a form of CEO compensation during the first half of the 2000s through the lens of the implementation of SFAS No. 123-R (FAS123R) in
2005. In contrast, this paper provides evidence that the most significant changes in firms’ use of CEO stock option
grants between 2000 and 2006 took place in 2002 following the wave of accounting scandals that rocked corporate America, well in advance of the implementation of this modifying rule, thus underscoring the importance of
considering the events that led to the implementation of FAS123R when analyzing changes in firms’ CEO compensation policy. Similar to extant studies, this paper provides evidence consistent with the idea that accounting
47
considerations play an important role in the design of CEO compensation. I document that changes in CEO stock
option grants are negatively related to the proposed construct that proxies for firms’ perceived accounting benefits
of using this form of compensation following the accounting scandals, which largely negated these benefits. In
addition, I find that changes in compensation practices are more pronounced in firms with a large proportion of
transient and unsophisticated investors, which supports the notion that the value-relevance of disclosed (but not
recognized) information is a function of the composition of the firm’s investor base. I also find that the magnitude of the reduction in firms’ use of CEO stock option grants increases as a function of their boards’ reputational
concerns for those firms with a higher probability of being covered by the business media. This evidence associates firms’ reduction in the use of option grants to the intense media criticism of boards’ compensation practices
throughout 2002, and from a policy perspective, it suggests the potential of shaming mechanisms to bring about
desired changes in corporate governance.
As a complementary test of the role of investors’ improved focus in the modification of CEO compensation
practices, I use the discontinuity of earnings distributions around key earnings benchmarks (in this case, the
recent performance benchmark) as an alternative proxy for firms’ motivation to boost reported earnings. I find
that before the accounting scandals firms close to missing the recent performance benchmark nearly double their
use of CEO stock option grants relative to their use of options when facing a “favorable earnings scenario.” I also
find that this differentiated response vanishes for firms close to missing the prior performance benchmark after
the accounting scandals. Together these results corroborate the conclusions reached using the proposed construct
of perceived accounting benefits that firms lose their ability to manage financial reporting perception using CEO
compensation design after the advent of the accounting scandals.
The results presented here accommodate existing evidence on the effect of SFAS 123-R on the use of option
grants. Firms certainly retained some motivations to boost reporting earnings—via option grants—until the implementation of FAS213R, for example, the ability to reduce the probability of violating earnings-based covenants
that are generally written in terms of floating GAAP. Overall, however, the evidence suggests that investors’ improved focus on the quality of firms’ financial reporting—catalyzed by the accounting scandals—was behind the
most significant changes in firms’ perceived accounting benefits of using stock options and thus in the decrease
in their use of this form of compensation.
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58
Other
Other CEO
compensation
ExecuComp
ExecuComp
ExecuComp
Compustat
Compustat
Compustat
Compustat
Market-to-book
Interest
coverage
Dividend
constrained
Cash flow
shortfall
CF shortfall
Compustat
ln(Sales)
Execucomp
Cash
Salary and
bonus∗
FIX /
FLEX
RSG
CEO restricted
stock grants∗
ExecuComp
Firm with a fixed/
flexible granting
schedule
SOG, ln(SOG)
CEO stock
option grants∗
ExecuComp
ExecuComp
TC
Total CEO
compensation∗
Dataset
ln(Tenur e)
AB
Notation
Perceived accounting
benefits of options
Variable
(continued on next page)
The three-year average of {cash dividends ([19]+[21]) - investing cash flow [311] - operating cash
flow [308]}/{total assets [6]};
Dummy variable that takes values of 1 when the firm classifies as dividend constrained in any of the three
previous years. A firm falls in this classification when {retained earnings[36] + cash dividends ([19]+[21])
+ stock repurchases[115]}/{prior year sum of cash dividends and stock repurchases} < 2. This dummy
variable also takes values of 1 when the denominator of the ratio is zero for the three previous years;
Operating income [13] divided by interest expense [15]. Following Yermack (1995), for values above
50 and below -50, I truncate the ratio to lie between these two values; and assign a zero value to those
observations with missing interest expense and long-term debt equal to zero;
The sum of total liabilities ([9] + [34]) and market value of equity ([10] + [199]*[25]) divided by book
value of total assets [6];
The natural log of total sales [12];
Firms that grant CEO option grants under a fixed schedule between 1999 and 2001. A CEO option
grant falls under this classification when the grant date vary up to 2 (or 5) days relative to the grant
date of the previous year’s grant. All other firms are classified as having a flexible award schedule.
The natural log of CEO tenure measured in years;
The sum of tong-term incentives LTIP, and other compensation (OTHANN + OTHCOMP) when
available before 2007; otherwise the sum of NONEQ_INCENT, PENSION_CHG, and OTHCOMP;
The sum of SALARY and BONUS
Grant-date value of restricted stock grants RSTKGRNT when available before 2007; fair-value of
stock awards as reported in the Plan Based Awards table OPTION_AWARDS_FV otherwise∗ ;
Black-Scholes grant-date value of stock option grants OPTION_AWARDS_BLK_VALUE when
available before 2007; fair-value of option awards as reported in the Plan Based Awards table
OPTION_AWARDS_FV otherwise∗ ;
Total Compensation TDC1 before 2007; TOTAL_ALT1 after 2006∗ ;
See Table 1.
Calculation†
APPENDIX A: Variable Definitions
59
AGE
Board’s
average age
I collect data on public float from firm’s 10-K annual filings when their market capitalization ranged between $75 and $175 million. A market capitalization under $75 million implies that
the firm is a smaller reporting firm
This definition accommodates changes in the reporting requirements under FAS123R.
‡
The number of analysts following a firm at fiscal year-end.
I include 3 different variables to measure level of compliance with the enhanced board requirements proposed
by NYSE and Nasdaq at the announcement date. First, a dummy that takes values of 1 when the majority
of directors are not classified as independent. Second, a dummy that takes values of 1 when the boards’
compensation committee is not fully composed of independent directors. Third, a dummy that takes values
of 1 when the boards’ nominating committee is not fully composed of independent directors.
Average age of the board members of a firm at fiscal-year end; and
I merge the institutional investors classification proposed in Bushee (2001) with Thomson Reuters 13F
and then calculate the percentage of common shares held by transient investors;
Percentage of common shares held by institutional investors at the end of the calendar quarter closest
to firms’ fiscal year-end;
Compustat item numbers are presented in square brackets. All monetary values are converted to 2002-constant USD using the consumer price index.
I/B/E/S
RiskMetrics
(ISS)
RiskMetrics
(ISS)
Bushee (2001)
Thomson
Reuters 13F
Following Aggarwal and Samwick (1999), I calculate the cumulative distribution function of the variance
of returns for firms in the sample;
Variance of cumulative weekly returns over the fiscal year;
Cumulative 12-month returns including dividends over the fiscal year (RET);
This variable takes values of 1 for U.S. firms with a public float under $75 million for all years between 2002
and 2006. And values of 0 for U.S. firms with a public float above $75 million for all years between 2002 and
2006.
∗
CRSP
CRSP
Compustat &
SEC Edgar
This variable takes values of 1 for U.S. firms with a public float above $75 million for all years between 2002
and 2006. And values of 0 for U.S. firms with a public float under $75 million for all years between 2002 and
2006.
Firms with incorporation code (Compustat FIC) different to USA.
The ratio of income before extraordinary items adjusted for common stock equivalents [20] over fully
diluted common shares [171] after adjusting for stock splits [27];
Calculation†
APPENDIX A (continued)
†
Analyst
coverage
AC
T
Transient
ownership
Compliance with
exchanges enhanced
requirments
IO
2
F σRetur
n
2
σRetur
n
NON-ACC
Institutional
ownership
CDF of the variance
of returns
Return variance
Stock return
Nonaccelerated
filer‡
Compustat &
SEC Edgar
ACC
Accelerated
filer‡
Compustat
Dataset
Compustat
EPS
Notation
Foreign firm
Earnings per share
Variable