Do private equity investors take firms private for
different reasons?∗
Jana P. Fidrmuc†
Peter Roosenboom‡
Warwick Business School
RSM Erasmus University
Dick van Dijk§
Econometric Institute
Erasmus University Rotterdam
April 2007
Abstract
In recent years, the going private market experienced a considerable boom in
size and also became more interesting for private equity investors. This paper
shows that the higher involvement of private equity investors affects the going
private market as these investors approach firms with different characteristics
relative to the traditional management buy-outs. Our results on a sample of
212 UK going private transactions completed in the period 1997-2003 suggest
that private equity backed deals differ from management sponsored deals without any backing of private equity investors in four ways. First, even though
both types of deals suffer from market undervaluation, the mis-pricing is larger
for management sponsored deals. Second, it is only management sponsored
deals that suffer low financial visibility as their stock’s analyst coverage and
frequency of trading are low. Third, Jensen’s free cash flow hypothesis seems
not to apply to private equity backed deals as they have shortage of cash, low
debt levels, and pay high dividends. Finally, the two types of deals differ in
ownership structure. Private equity backed deals seem to have higher ownership by financial institutions and their ownership is less concentrated.
Keywords: Going Private Transactions, Corporate Governance, Private Equity
JEL Classification: G32, G34
∗
We would like to thank participants of the Conference on Mergers and Acquisitions in Exeter
and research seminars at the Warwick Business School, Manchester Business School, XFi Centre
for Finance and Investment, University of Exeter, RSM Erasmus University and University of
Antwerp. Also, we are grateful to Ian Garrett, Abe de Jong and Ian Tonks for useful comments on
an earlier draft of the paper and Shanaz Raja, Huiyan Xu and Hans van der Weijden for valuable
research assistance.
†
Warwick Business School, University of Warwick, Coventry CV4 7AL, United Kingdom, Email: [email protected] (corresponding author)
‡
Department of Financial Management, RSM Erasmus University, P.O. Box 1738, NL-3000 DR
Rotterdam, The Netherlands, e-mail: [email protected]
§
Econometric Institute, Erasmus University Rotterdam, P.O. Box 1738, NL-3000 DR Rotterdam, The Netherlands, e-mail: [email protected]
1
Introduction
We investigate whether private equity investors take firms private for different reasons than the firm’s managers. The paper builds on the heterogeneity hypothesis of
Halpern et al. (1999) who argue that “the population of LBOs is heterogeneous and
understanding the differences among LBOs is important in understanding both why
each type of firm engaged in this transaction and what they did after the transaction” (page 282). We use this idea of heterogeneity to highlight the involvement of
private equity funds. While undervaluation, low financial visibility, high managerial
ownership and free cash flow are key factors motivating management to take their
firm private, we examine whether private equity investors pursue similar goals or
may have other incentives to be involved in a public to private transaction.
Several empirical papers have examined the reasons for why publicly listed firms
decide to go private (see Maupin et al., 1984; Lehn and Poulsen, 1989; Denis, 1992;
Opler and Titman, 1993; Halpern et al., 1999 for US and Weir et al., 2004, 2005
for the UK). However, most of this empirical evidence investigates the going private
deals of the 1980s, focuses on testing of Jensen’s (1986) free cash flow hypothesis
and does not consider the effect of private equity involvement. Recently, private
equity houses have considerably increased their investment and at the same time
changed their strategy concerning the going private market. In the 1980s, private
equity investors often engaged in highly leveraged transactions, many of which were
seen as hostile by incumbent management. Nowadays, private equity investors are
often looking to partner with management. They are interested in sound strategic
reasons to justify any going private deal and low valuation on its own is not enough.
In fact, private equity investors are after fundamentally strong businesses. Their
typical target is a long-term player, has leading market share within their defined
niche, and has a strong customer base, good growth potential and good margins (The
Deal, 24 November 2003). Operational improvements have become more important
than they were before in the 1980s (Business Week, 27 February 2006).
In this paper, we conjecture that going private deals that have backing of a private
equity house have different characteristics and go private for different reasons than
1
management buy-outs that do not have the backing of a private equity house. In
other words, the population of going private firms is not homogeneous with respect to
the reasons for the deal and pooling all deals together may result in biased results.
The population of public to private (PtP) deals can be partitioned into separate
groups: (i) private equity backed going private transactions and (ii) management
buyouts where management initiates the deal without any direct involvement of
private equity funds (we refer to this group as ‘management sponsored deals’). We
show that these two groups have their own unique characteristics and their own
reasons for going private. To cover the whole population of PtP deals, our analysis
also includes a third type of ‘other’ going private transactions that are initiated by
different parties such as non-executive directors, wealthy private individual investors,
industrial firms or financial investors that are not private equity investors. This
group is included for completeness of our model. That is, we want to be sure that
any differences found between the private equity backed deals and management
sponsored deals are not arising due to omitting the remainder of the going private
firms.
We consider four different aspects that may motivate management sponsored
deals and argue that private equity sponsored deals do not share these incentives
as they target firms with shortage of cash that can be turned around and make
the investment to pay off (Gompers and Lerner, 2001). The first of our hypotheses
is that market undervaluation may motivate managers to take their firms private.
However, information asymmetry and undervaluation may be less important for
private equity backed deals because private equity investors are primarily interested
in a strategic justification for their transactions (Gompers and Lerner, 2001). The
second hypothesis concerns the recently recognized proposition that low financial
visibility plays an important role in going private decisions as it undermines the
main benefits of public ownership (see Boot et al., 2006 and Mehran and Peristiani,
2006). Again, we suggest that management sponsored deals are motivated by low
financial visibility of their stock because firm managers weight the costs and benefits
of the public listing. In contrast, private equity investors profit from the strategic
2
advantages of the deal and so financial visibility may be less important for them.
Third, we explore the free cash flow idea of Jensen (1986) separately for the two
groups of going private transactions. In particular, we propose that private equity
backed firms may have less agency problems associated with free cash flows as, in
fact suggested by Gompers and Lerner (2001), they are more likely to have low cash
balances. Finally, we highlight the differences in pre-buyout ownership structure
between the two types of firms. Elitzur at al. (1998) propose that managers who
have a large amount of their personal wealth invested in the company may be more
likely to decide in favor of a going private deal. In contrast, we argue that private
equity backed deals may have high institutional ownership as institutional investors
may be willing to support their deal. We note that several of the above hypotheses
have not been explored in this context so far, which constitutes another contribution
of this paper.
In our empirical analysis, we examine a sample of 212 UK going private transactions that occurred during the period 1997-2003 and compare them to randomly
selected UK firms that remained listed on the London Stock Exchange.1 Using
multinomial logistic models, we find support for our conjecture that going private
firms are heterogeneous with respect to the private equity involvement in the deal
and, at the same time, the two groups of going private firms are different from
the firms that remained public. First, the results suggest that even though both
types of deals suffer market undervaluation, the mis-pricing is significantly larger
for the management sponsored deals. Second, we show that it is only management
sponsored deals that suffer from low financial visibility as they fail to attract sufficient investor interest. The private equity backed deals are not distinguishable in
this respect from the control group of firms that remain public. Third, our results
suggest that the free cash flow hypothesis does not apply to private equity backed
deals. As expected, they seem to have lower cash balances and, at the same time,
pay higher dividends than management buyouts. Finally, the two types of deals are
1
In the US, the total value of deals went up to $65 billion over the period 1997-2002 (Renneboog
and Simons, 2005). The UK deals total £30 billion over the period 1997-2003. This makes the UK
going private market second only to that in the US.
3
heterogeneous in pre-buyout ownership structure. Our results show that firms in
private equity backed deals have higher ownership by financial institutions and their
ownership is less concentrated relative to management sponsored deals.
The plan of the paper is as follows. In Section 2, we put forward our four
heterogeneity hypotheses. Section 3 describes our dataset, methodology and provides
basic descriptive statistics of our sample. Section 4 presents our empirical results.
Section 5 concludes.
2
Heterogeneity in going private transactions
In this paper, we propose that private equity involvement in going private transactions is associated with different firm characteristics compared to transactions initiated and led solely by managers. Thus, we argue that in order to get a better picture
of the reasons and characteristics of going private transactions, we should not pool
all going private firms together but rather separately evaluate the reasons for going
private in two different types of deals: (i) going private transactions backed by private equity investors, and (ii) going private transactions led by the firms’ executive
directors without any backing of private equity investors (management sponsored
deals).
As active private equity involvement in going private transactions is a relatively
new phenomenon, the literature explaining the motives of private equity investors
to engage in these deals is very limited.2 Therefore, it is not easy to come up
with rigorous hypotheses backed up with theoretical models or previous empirical
findings. The literature, however, is quite useful in providing explanations for the
motives of managers (as opposed to private equity investors) engaging in the buyout deals. In fact, we form four relatively broad hypotheses explaining why firms
go private: (i) undervaluation of the going private firms relative to their peers in
the market; (ii) their low visibility in the form of illiquidity of their stock and low
analyst coverage; (iii) high free cash flows and tax shields; and (iv) high managerial
ownership. At the same time, we suggest that these four motivations may be less
2
Gompers and Lerner (2001) is one of the few references.
4
applicable to private equity backed deals as the general perception of private equity
investors’ activities suggests that private equity houses are interested in turning
companies around within a three to five year horizon and making decent returns
on their investment (Business Week, 27 February 2006). Also Gompers and Lerner
(2001) support this view. The following subsections explain the four possible ways
in which the two types of going private transaction may differ.
2.1
Heterogeneity in undervaluation
Due to information asymmetry between management and outside shareholders, it
is possible that the market value of a company is above or below its fundamental
value (Merton, 1987). In going private transactions, the perceived undervaluation
may play an important role as it potentially limits management’s ability to use the
benefits available to public companies as, for example, the accessibility of funds
required to finance new investment projects or acquisitions (Allen and Gale, 1999;
Pagano et al., 1998). A survey among US managers who underwent a going private
transaction indicates that the perceived undervaluation is indeed one of the primary
reasons for managers to take their firm private (Maupin et al., 1984). Moreover,
undervalued firms are more likely to attract hostile takeover interest (Lehn and
Poulsen, 1989) that may lead to managers losing their jobs (Lowenstein, 1985).
Therefore, we conjecture that undervaluation of firms increases the likelihood of
managers taking their firms private as this allows the managers to keep control over
the firm at relatively low cost and avoid (hostile) takeovers.
Outside investors may also value a company more than the current market value
because they perceive unexploited growth opportunities. In particular, private equity
investors may target firms because of poor management that has not been taking full
advantage of the firms’ growth potential possibly due to a lack of cash (DeAngelo,
1990). Gompers and Lerner (2001) argue that private equity investors look for
companies that have the potential to evolve in ways that create value and at the
same time face problems in raising funds via regular bank debt or public equity.
This aspect may be more important in the UK versus the US since the UK venture
capital and buyout markets have traditionally been more closely linked (Toms and
5
Wright, 2005). In addition, private equity investors may also follow a ‘buy-andbuild’ strategy in which they focus on growth opportunities within an industry and
hire new managers with industry experience as needed. In such a buy-and-build
strategy the private equity investor acts as an industry consolidator taking over a
number of smaller rival firms. In this case, the value of the going private transaction
to the investor depends more on the investor’s previous portfolio investments rather
than the firm being undervalued (Smit and DeMaeseneire, 2005). Moreover, it seems
plausible that private equity investors have less superior information relative to the
management concerning the undervaluation of the firm.
In short, we predict that undervaluation plays a role in both types of going private transactions but to a different degree. Undervaluation is expected to be more
important in management sponsored private equity transactions. In contrast, undervaluation is less important in transactions backed by private equity because these
active investors could create additional value using their unique resources such as
their network and industry expertise and do not have private information advantage.
2.2
Heterogeneity in visibility
If a firm’s shares are thinly traded, being public may not be worth the cost (Bolton
and von Thadden, 1998). Boot et al. (2006) highlight liquidity and low cost of capital
as important benefits of public versus private ownership. Furthermore, thinly traded
stocks have lower analyst coverage in general and are at risk of being neglected by
investors when taking their investment decisions (Merton, 1987). Thus, firms that
are not able to attract adequate level of investor recognition have to bear the high
cost of stock exchange listing while not taking enough advantage of the benefits of
being a public company (Mehran and Peristiani, 2006). Therefore, we propose that
low financial visibility may motivate managers to take their firm private.
In contrast, the predominant and most important economic activity for private
equity investors is the restructuring of their targets. Often, it is argued that it is
easier to fix a private company rather than a public one. This is because public
shareholders are strongly focused on quarterly results, which often is in sharp conflict with long term strategic goals of multiyear restructuring efforts. Thus, going
6
private firms with private equity backing may actually want to hide from relatively
high visibility and scrutiny of the public market. At the same time, private equity
investors prefer projects where monitoring and selection costs are relatively low and
where the information asymmetry costs are less severe (Amit et al., 1998). So, relative to the management sponsored PtP firms, we expect private equity backed deals
to be associated with less problems in attracting financial visibility and investor
recognition.
2.3
Heterogeneity in free cash flows
Most of the empirical evidence concerning going private transactions so far is based
on Jensen’s free cash flow hypothesis. Jensen (1986) proposes that debt-financed
going private transactions may provide a solution to firms in cash-rich, slow growth
and declining industries. The main argument maintains that firms with large cash
balances but low growth prospects are vulnerable to conflicts of interest between
managers and shareholders over payout and investment policies. To mitigate this
problem, managers can increase dividends and thus provide payout of current cash
that would otherwise be invested in low-return projects or be wasted. However, a
promise of increased dividend is not credible as managers can easily reduce dividends
in the future at low cost. In contrast, issuing debt in exchange for stock is a credible
strategy to limit free cash flows because unpaid interests lead to bankruptcy.
The empirical studies testing the free cash flow hypothesis in the context of going
private transactions provide mixed results. Lehn and Poulsen (1989) document that
undistributed cash flow is a significant determinant of a firm’s decision to go private.
Also Opler and Titman (1993) show that high cash flow firms that also have a low
Tobin’s q (which they use to measure future growth prospects) are more likely to
undertake an LBO. In contrast, Kieschnick (1998) argues that after accounting for
choice based sampling, outliers in the data, and potentially misspecified variables,
prior growth rate and level of free cash flows are not significant determinants of the
odds of going private for the Lehn and Poulsen (1989) dataset. Furthermore, Weir
et al. (2004) also fail to find significance of free cash flows in determining the odds
of going private for UK firms. All the empirical papers, however, pool all going
7
private firms together and treat them as a single homogeneous group. Halpern et
al. (1999), in contrast, distinguish two groups of going private firms according to
the level of management ownership before the transaction but do not test the free
cash flow hypothesis on the two groups separately.3
We propose that private equity backed deals, as opposed to management sponsored deals, do not match the profile of Jensen’s cash-rich firms. Intuitively, firms
that search for private equity backing are more likely to suffer low cash balances and
high growth opportunities. (Gompers and Lerner, 2001). In contrast, managers of
cash-rich firms with low growth prospects may see the potential of leveraged transactions as they may be ware of their ill-specified incentives and use the excess cash to
fund the going private transaction or to service new debt and gain control over their
firms (Fox and Marcus, 1992). Moreover, the leveraged transaction allows managers
to avoid the prospect of hostile takeover and/or shared control with an active private
equity investor who typically demands board representation and a say in the firm’s
long-term strategy (Cotter and Peck, 2001). Therefore, we predict that large cash
balances may motivate management sponsored going private transactions but not
private equity backed deals.
2.4
Heterogeneity in pre-buyout ownership structure
Halpern et al. (1999) and Elitzur at al. (1998) argue that managers who have a
large amount of their personal wealth invested in the company may prefer to diversify
their portfolio while keeping or increasing their control over the firm. Managers with
large equity stakes therefore have incentives to initiate a leveraged buyout and use
new debt to decrease their wealth invested in the firm. It is relatively inexpensive
for these managers to acquire a majority of the votes and force the leveraged buyout
given their large equity stakes and any potential information asymmetry (Elitzur et
al., 1998). Thus, we expect that managers with high ownership stakes in their own
firm are more likely to take their firm private and do not seek backing by private
3
In particular, they propose that for the group of low management ownership firms, third-party
takeover pressures force management to consider a third party led buyout or face the prospect
of hostile takeover. In the case of high prior management ownership group, managers hold large
undiversified portfolios and so have an incentive to take cash out of their firm.
8
equity investors.
Private equity investors typically look for support of a number of incumbent
blockholders before they take a firm private. This increases their chances of the deal
being successful. In fact, the private equity investor usually contacts the existing
blockholders in order to receive irrevocable undertakings wherein the existing blockholders promise to accept the private equity investor’s offer (Wright et al., 2007).
After receiving the support of several blockholders, the private equity investor makes
a public offer for the remaining shares at the same price. These irrevocable undertakings are easier to obtain when ownership is concentrated in the hands of a small
number of outside shareholders. Institutional investors may be of special importance
as most of them are passive and not interested in monitoring management closely
themselves (Faccio and Lasfer, 2000). They are likely to sell their shares in case they
are able to negotiate a premium price and earn a return on their otherwise illiquid
investment. Therefore, we expect that high institutional ownership increases the
likelihood of a going private transaction backed by private equity.
3
3.1
Data and methodology
Sample selection
Our original sample consists of 221 non financial firms that have gone private in the
United Kingdom during the years 1997-2003. We identify these public-to-private
(PtP) transactions from the database of the Centre for Management Buyout Research (CMBOR). For all the PtP firms, we also obtain the offer documents accompanying the going private transaction from Thomson Research. We use these
documents to determine whether the deal is backed by private equity investors or not.
This is our primary classification criterion to decide upon whether a deal falls in the
private-equity backed group. In case the transaction is not backed by a private equity
house we further examine whether any of the firm’s executive directors are involved
in the deal. If this is the case, the transaction is coded as a management-sponsored
deal. In all other cases, the transaction is classified into the ‘other’ category. This
category includes deals backed by non-executive directors, wealthy families or insti9
tutional investors other than private equity houses and is included for completeness
of our model.4
We do not have data for 9 PtP firms and, therefore, our final sample of going
private transactions consists of 212 firms that are relatively evenly distributed between 90 private equity backed deals, 69 deals led by management and not backed
by private equity investors and 53 ‘other’ deals led by other parties.
In order to get a more detailed picture of the characteristics of the going private
transactions in general and the three individual types of going private transactions
in particular, we contrast the going private firms with firms that remained publicly
listed. We opt for a random sample of control firms that remained public: in each
of the years of the sample period of 1997-2003 we randomly select 200 control firms
from a population of around 1200 firms that continue to be publicly traded in a
given year. The sampling procedure allows for a control firm to be included in
the sample more than once. In total, we collect data on 1400 control firm-years
that cover 960 different control firms. For both the PtP firms and control firms,
we get market prices from Datastream, financial statement data from Worldscope,
and hand-collect ownership structure and board composition information from Price
Waterhouse Coopers’ Corporate Register (various issues).
In the going private literature it is common to use a matched control sample
of firms that remained public. Firm size and industry classification are the most
commonly used matching criteria. In general, the sampling method – whether we
use a random sample as in this paper or a matched sample – does not pose any
advantages or disadvantages except for the case where the characteristics used for
the matching process are important determinants of the going private process. We
prefer random to matched sampling because we believe that firm size is a relevant
factor influencing the decision to go private. Using a matched sample would not
allow us to test for this possibility.
4
Note that the private equity backed deals may be both with and without management involvement. In fact, private equity funds often look for management involvement in their deal. We believe
it is the presence of private equity involvement that matters for the deal characteristics. Private
equity investors are attracted to certain type of firms and managers behave differently conditional
on private equity interest.
10
3.2
Descriptive statistics
Table 1 lists our variable definitions and Table 2 shows summary statistics for all
PtP firms together, the three types of PtP deals separately, and the control firms.
All variables are trimmed at the 1st and 99th percentiles, except the ownership
and illiquidity variables. We test for differences in means and medians between the
control firms and PtP firms and among the three types of PtP firms. We use a t-test
for equal means allowing for unequal variances and a Mann-Whitney U-test for equal
medians. Below we first discuss the differences between the complete sample of PtP
firms versus the control firms and then we focus on the differences between the two
main types of going private firms. The latter comparison highlights the heterogeneity
idea proposed in this paper. We only discuss the statistically significant differences
for both the mean and the median.
- insert Tables 1 and 2 about here Firms that go private are valued less than control firms as indicated by their
lower market to book ratios. They also experience more takeover rumours. PtP
firms’ shares are traded less actively than the shares of the control firms and are
followed on average by fewer analysts. Moreover, the going private firms have higher
equity stakes held by executive directors and financial institutions than the control
firms. As a result, ownership concentration, as measured by the Herfindahl index,
is higher in PtP firms than in control firms. In addition, firms that go private have
less cash and marketable securities, lower sales growth, higher payout ratio and are
more profitable than control firms.
Turning to the differences between the different groups of going private firms, we
observe that the market to book ratio of the private equity backed deals (that we use
as a measure of stock market valuation) is higher indicating that firms that go private
with the help of private equity investors are less undervalued than the management
sponsored deals. They also have higher analyst following and their shares are more
actively traded. At the same time, the private equity backed deals are larger relative
to the deals sponsored by management. Moreover, the private equity backed PtP
11
deals have significantly lower executive ownership and higher ownership by financial
institutions relative to the management sponsored transactions. Also, their pretransaction ownership concentration is lower.
Finally, we compute the free cash flow proxy of Lehn and Poulsen (1989) used in
the going private literature so far. Table 2 shows that private equity backed deals
have higher free cash flow in the previous year. However, DeAngelo and DeAngelo
(2006) argue that the firm’s current cash flow is not a suitable measure of managerial
opportunism as it does not reflect the stock of resources at managers’ disposal. What
matters is managers’ access to the stock of liquid assets at all points in time which is
better reflected in the firm’s cash and marketable securities. In fact, free cash flow is
highly correlated with the firm’s profitability which indicates that it is restricted to
affect managerial incentives only at ’a distribution point’ (DeAngelo and DeAngelo,
2006) and, thus, it may not be enough to encompass all resources at managers’
disposal and therefore is not able to generate empirically valid predictions. This
suggests that the firm’s cash level is a more suitable proxy for company free cash
at the disposal and discretion of the managers. Table 2 reports that private equity
backed deals have less cash on their balance sheet than the management sponsored
deals.
3.3
Model
We employ multinomial logistic regression (MNLR) models to examine the heterogeneity hypotheses developed in the previous section. As mentioned before, we
divide our sample of UK firms into four different groups: (1) private equity backed
PtP deals, (2) management sponsored PtPs, (3) other PtPs, and (4) non-PtPs. We
denote the observed group for firm i by the variable yi , which can take the discrete
values 1, 2, . . . , M, where M = 4 in our case. In the MNLR model the probability
that firm i will belong to group m, conditional on the (k × 1) vector of explanatory
variables xi consisting of a constant and firm characteristics, is given by
′
exp(βm
xi )
P [yi = m|xi ] = PM
,
′
l=1 exp(βl xi )
12
for m = 1, . . . , M.
(1)
For identification purposes, we set the coefficients for the non-PtP group of firms
equal to 0, that is β4 = 0.
Estimation of the coefficients in the MNLR model in (1) is straightforward by
means of maximum likelihood, except for the following caveat. Our data set is not
a random sample from the population of all firms. In particular, while we include
all known PtP deals during the period 1997-2003, each year we only sample 200 of
the firms that remain listed, which in total equal 1200, on average. This implies
that PtP firms are considerably overrepresented in our sample compared to the
underlying population of firms. Not accounting for this selective sampling would lead
to biased estimates of the intercepts and incorrect standard errors for all estimated
coefficients; see Kieschnick (1998) and Fok and Franses (2002) for detailed analysis
of selective sampling in the context of binary and ordered logit models, respectively.
The problem can be remedied by defining modified probabilities as
γm P [yi = m|xi ]
P̃ [yi = m|xi ] = PM
,
l=1 γl P [yi = l|xi ]
for m = 1, . . . , M ,
(2)
where γm is the fraction of firms in group m that is included in the sample. Hence,
in our case γ1 = γ2 = γ3 = 1 while γ4 = 1/6. The correct likelihood function, which
is used for parameter estimation then makes use of these corrected probabilities.
The effects of the firm characteristics xi on the probabilities that a firm engages
in the different types of PtP deals is a nonlinear function of the model parameters
βm , such that interpretation of these parameters is not straightforward. For interpretation of the model, it is useful to consider the log-odds ratio of group m versus
group l, defined as
log
P [yi = m|xi ]
P [yi = l|xi ]
= x′i (βm − βl ).
(3)
This shows that firms with a larger value for xi,j more likely belongs to group m
than to group l if (βm,j − βl,j ) > 0, where xi,j indicates the j-th element of xi , and
βm,j and βl,j are the corresponding coefficients. Note that this does not necessarily
imply that the probability that firm i belongs to group m increases with xi,j , as the
the odds ratios of group m versus the other categories also change. The net marginal
effect of a change in xi,j on the group probability follows from the partial derivative
13
of P [yi = m|xi ] with respect to xi,j , which is given by
!
M
X
∂P [yi = m|xi ]
= P [yi = m|xi ] βm,j −
βl,j P [yi = l|xi ] .
∂xi,j
l=1
(4)
The sign of this derivative depends on the sign of the term between brackets, which
may be positive or negative depending on the value of xi . Hence, the sign of the
marginal effect of xi,j on P [yi = m|xi ] will not always correspond with the sign of
the coefficient βm,j . Also note that the marginal effect depends on the values of the
other explanatory variables in xi , denoted as xi,−j . In order to obtain a clear view on
∂P [y =m|x
]
i
i,j
=
the effect of the variable of interest xi,j one should therefore consider
∂xi,j
R
∂P [yi =m|xi ]
dxi,−j , integrating out the effects of these other explanatory variables.
∂xi
xi,−j
In practice this can be done by averaging (4) across all realizations of xi,−j in the
sample for each value of xi,j . In the empirical analysis presented in the next section
we follow a more direct approach by considering the group probabilities P [yi =
m|xi,j ] themselves to gauge the effects of the different firm characteristics on the
probabilities of the different types of PtP deals. We do average out the effects of other
P
explanatory variables by computing P [yi = m|xi,j ] = N1 N
n=1 P [yn = m|xi,j , xn,−j ],
where N is the sample size.
An important assumption underlying the MNLR model in (1) is independence of
irrelevant alternatives, meaning to say that the odds ratio of, for example, PE-backed
and management-led PtP deals does not depend on the inclusion of the third group
of PtP transactions, which can also be seen from (3). We examine the validity of this
assumption by means of the specification test developed by Hausman and McFadden
(1984).
Our heterogeneity hypothesis of PE-backed and management-led PtP transactions implies that certain firm characteristics such as free cash flow and management
ownership affect the relative probabilities of a firm belonging to the different groups.
Put differently, in the MNLR model the coefficients βm,j should differ across groups
m. For an individual variable, xi,j say, the null hypothesis of no heterogeneity across
groups m and l can easily be tested by means of a likelihood ratio test of the restriction βm,j = βl,j . The same holds for a given sub-set of the explanatory variables
included in the model. Testing whether there is no heterogeneity at all is slightly
14
more involved, and effectively boils down to testing whether two groups can be combined into one. This is done by means of the likelihood ratio test developed by
Cramer and Ridder (1991). Finally, we should remark that the above likelihood
ratio test statistics also enable us to assess in which respects the third group of
PtP transactions is similar to the private equity backed and management sponsored
deals.
4
Results
The main contribution of this paper is to show that the population of the PtP firms
is heterogeneous and that the reasons for going private are different for private equity
backed deals versus management sponsored deals. To examine this hypothesis, we
estimate multinomial logistic regressions with private equity backed, management
sponsored and other PtP deals analyzed against the control firms that remained
public.
Table 3 reports the estimation results. To account for industry and time effects,
all regressors are taken in deviation from industry median and annual median values.
Our model treats the non-PtP control firms as the omitted category. Hence, Table
3 reports coefficients for private equity backed, management sponsored and other
deals separately in subsequent columns. These coefficients show how the explanatory variables affect the probability of going private through the particular type of
transaction relative to the probability of remaining public. The last three columns
in Table 3 show p-values for a likelihood ratio test of equal parameters among the
three types of PtP deals. Thus, these columns show significance of pair-wise coefficient differences among the going private types. Finally, the last two lines of Table
3 show p-values for the likelihood ratio test of Cramer and Ridder (1991) that all
parameters except the intercept are equal for the corresponding two groups and the
independence of irrelevant alternatives test, respectively.
- insert Table 3 about here Overall, the results suggest that going private transactions are indeed heterogeneous. The no heterogeneity test of Cramer and Ridder (1991) suggests that all
15
three going private groups have different deal characteristics from the non-PtP firms
as well as from each other. Moreover, the independence of irrelevant alternatives
test reported in the last row indicates that the choice of the particular type of going
private deal (whether the deal is indeed supported by a private equity house or is
fully led by the management or is sponsored through another party) is fully independent. Thus, the decision to go private is made at the same time as the decision
about the type of the deal. In other words, it is not the case that the firm decides
about going private first and then looks for possible methods how to realize it. The
independence of irrelevant alternatives test also shows that the multinomial logistic
regression is the preferred estimation method and that this method fits the setting
of going private decision better than a nested logit model, for example.
4.1
Undervaluation
The results in Table 3 suggest that perceived undervaluation plays an important
role in management sponsored deals, but less so in private equity backed deals. The
coefficient for the market to book ratio is significantly negative for both types of
PtP deals showing that both the private equity backed and management sponsored
deals are on average undervalued relative to the firms that remain public. However,
the management sponsored deals are also significantly more undervalued (at the five
percent level) relative to the private equity backed deals. These results show that
low levels of market to book ratio increase the probability of going private but do
not necessarily support the link between market to book ratio and undervaluation.
Therefore, we perform several sensitivity checks that in our view provide additional
evidence that undervaluation is an important factor motivating (management sponsored) PtP deals.
First, we account for the possibility that the low market-to-book ratio reflects
low growth prospects or generally low performance of the PtP firms. Therefore,
we include average sales growth over the last three years and return on assets as
control variables in the regression in Table 3. The two variables are, however, not
statistically significant. It seems that neither growth prospects nor profitability
distinguish the going private firms from each other and from the firms that remained
16
public and suggests that market to book may indeed capture undervaluation of the
firms.
Second, we take advantage of the fact that takeover interest is often associated
with undervalued firms. The positive and significant (at the one percent level)
coefficients for rumours in Table 3 indicate that both PtP types experience relatively
high takeover interest in the period before the deal compared to the control group
of firms that remained publicly listed. Even though the two coefficients are not
statistically different from each other, a closer analysis of the relationship between
takeover interest and market to book ratio reveals interesting differences between the
management sponsored versus private equity backed deals. The literature suggests
that undervaluation of buyout firms increases chances of takeover bids from third
parties (Lehn and Poulsen, 1989) and then this jointly motivates management to
take their firms private. Our data confirm this conjecture for the management
sponsored deals: The frequency of rumours conditional on market to book being
below the sample median is 39%, 48%, and 23% for the management sponsored deals,
private equity backed deals and non-PtP firms, respectively, whereas the frequency
for firms with market to book above the sample median is 0%, 39%, and 18%,
respectively. So, the management sponsored PtP deals experience a very strong
negative correlation between market to book and takeover rumours. In other words,
the management sponsored deals are attractive for third parties only in case they
have relatively low market to book ratio and we believe this result supports the
conjecture that undervaluation together with takeover threat play a very important
role in motivating the management sponsored deals. This effect is not present for
the private equity backed deals which then indicates that undervaluation does not
play such a primary role for the latter type.
To confirm this conjecture in a regression setting, we augment our basic regression
model from Table 3 with an interaction term between market to book and rumours.
Results are reported in Table 4 and support a strong complementary effect between
market to book and rumours for the management sponsored deals: The coefficient
for the interaction term is, as expected, negative and significant and, moreover,
17
the rumour coefficient becomes statistically insignificant. The private equity backed
deals are not affected in the same way. In short, our data show that high takeover
interest strengthens the effect of low market to book ratio on the probability of
management sponsored PtP deals and supports our hypothesis on undervaluation.
- insert Table 4 about here Third, insider trading patterns in firms in our sample reported in Table 4 provide
further evidence supporting the view that our results on market to book ratio pick
up the effect of private information. If undervaluation is indeed one of the reasons
for going private, we could expect that managers’ trading in advance of the event
may partially reveal the importance of that information. In fact, Harlow and Howe
(1993) document significant increase in trading by insiders prior to the announcement
of US management-led buyouts over the period from 1980 to 1989. They show,
however, that this abnormal pattern arises not from increases in purchases but from
abnormally low levels of stock sales. Harlow and Howe (1993) argue that this passive
insider trading strategy is preferred by managers as it reduces their liability risk. In
line with this existing evidence, our management sponsored deals should experience
abnormally low insider sales relative to the private equity backed deals and non-PtP
firms.
In order to show this, Table 4 includes two dummy variables that reflect the
insider purchase and sale patterns of executive directors of firms in our sample. In
particular, the executive director purchase (sales) dummy is set to one in case an
executive director purchased (sold) some shares of his/her own firm in the calendar
year prior to the announcement and set to zero otherwise. Our results confirm that
managers of the management sponsored deals tend to sell their shares significantly
less often than their counterparts in non-PtP firms and private equity backed deals.
We do not see any significant differences for the purchase patterns. Thus, our results
are in line with Harlow and Howe (1993) and support our conclusion that firm
undervaluation is more important in motivating management sponsored deals versus
private equity backed deals.
18
Figures 1a to 1d show the probability that a firm belongs to a particular type of
going private transactions as a function of market-to-book, rumours, sales growth,
and ROA, respectively. These graphs provide additional intuition for the results
as the coefficient estimates in Tables 3 and 4 indicate only the sign but not the
functional form of the relation. It is important to note that the variable on the
x-axis is measured as deviation from the year and industry specific medians. In
addition, the effect of the remaining explanatory variables from the regression is
averaged out. Figure 1a supports our result from Table 3 that undervaluation tends
to be associated with the management sponsored deals and to a lesser extent with
the private equity backed deals. In fact, Figure 1a shows that the probability of
a management sponsored deal sharply increases as market to book falls whereas
this is not the case for the private equity backed deals. In Figure 1b we see that the
probability of both private equity backed and management sponsored deals increases
with the number of takeover rumours but the two functions are quite close to each
other.
In summary, our results suggest that market undervaluation plays an important
role for the management sponsored PtP transactions that have significantly lower
market-to-book ratio relative to the control firms that remained publicly listed.
In contrast, for private equity investors, undervaluation of their target is not so
important. On average, private equity targets are significantly less undervalued
than the management sponsored deals.
4.2
Visibility
The results in Table 3 also support our second conjecture that low financial visibility
may increase chances of management sponsored deals whereas it is not important
for private equity backed deals. The most widely accepted empirical measure of firm
visibility is the number of analysts following a firm (O’Brien and Bhushan, 1990).
Analyst reports are documented to be the primary source of information for most
buy-side investors (Baker et al., 2002). In Table 3, the coefficient for the number of
analysts following a firm is negative and significant at the ten percent level for the
management sponsored deals whereas it is positive but not significant for the private
19
equity backed deals. Importantly, the two coefficients are significantly different at
the ten percent level indicating that management sponsored deals suffer significantly
lower financial visibility relative to private equity backed deals.
We also employ an alternative measure of investor interest that is available for
all firms in our sample and is based on frequency of price changes. It is intuitive to
expect that stocks with low investor interest would be traded relatively infrequently.
Low trading frequency or days with no trading are then associated with no price
movements and zero daily returns. In contrast, stocks that attract investor interest are traded relatively frequently and experience frequent (small) price changes
that reflect constant price discovery. So, our measure of thin trading measures the
frequency of zero daily returns in the previous calendar year and high levels indicate low trading frequency, investor interest and financial visibility (Fidrmuc et al.,
2006). The results in Table 3 show that thin trading (high fraction of zero returns)
is associated with higher probability of a management sponsored deal relative to
both the non-PtP control firms as well as the private equity backed deals (both statistically significant at the one percent level), which confirms our financial visibility
hypothesis.
Firm size may also be closely associated with financial visibility of a firm (O’Brien
and Bhushan, 1990). The coefficient estimates for size (log of total book value of
assets), however, do not support this claim. In fact, the coefficient for management
sponsored deals is positive and significant indicating that management sponsored
deals on average are larger. Inspecting the correlation matrix, however, we find high
positive correlation between size and thin trading. This means that inclusion of the
thin trading variable in the regression strongly affects the coefficients for size. As
both variables are important and their correlation does not affect other coefficients,
we opt to include both total assets and thin trading in our model. However, the
coefficients for size should be interpreted with caution as excluding the thin trading
variable from the regression results in size being negative and statistically significant
(at the one percent level) for the management sponsored deals and not significant
for the private equity backed deals.
20
Another important issue is that our results for financial visibility may be driven
by size. Put differently, it may be the case that our PtP firms, especially the management sponsored deals, are relatively small and then of course, they are thinly
traded and not followed intensively by analysts. To make a stronger case for our
visibility hypothesis, we would like to see that the PtP firms are distributed relatively evenly across the size spectrum of our sample and show lower levels of analyst
coverage and higher levels of thin trading across all sizes. In order to check this,
Table 5 shows mean values of our analyst coverage and thin trading measures as
well as frequencies of private equity backed and management sponsored deals across
size deciles (measured by total assets). Table 5 confirms that analyst coverage and
thin trading are indeed positively correlated with size. However, it also shows that
even though the going private firms are significantly underrepresented among the
smallest and largest firms, PtP firms are relatively evenly spread across the remaining 8 middle size deciles. Thus, this indicates that the association between financial
visibility and probability of going private is not due to the size effect.
- insert Tables 5 and 6 about here To push this argument further, Table 6 reports the same statistics by size deciles
separately for the non-PtP firms, private equity backed firms and management sponsored firms, respectively. Panel A, reporting the means for the non-PtP firms, confirms the overall trend that analyst coverage is increasing and thin trading falling
with size. The same pattern is reflected in Panel B for the private equity backed
firms. Overall, the private equity backed deals seem to be slightly less frequently
traded but equally monitored by analysts relative to the non-PtP firms across all
size deciles. In contrast, Panel C shows sharply lower analyst coverage and thinner
trading for management sponsored deals relative to the non-PtP firms across all size
deciles. Thus, this shows that the management sponsored deals suffer lower market
visibility relative to both private equity backed deals as well as the non-PtP deals.
Moreover, this effect is clearly present across all size deciles and, thus, is not driven
by size.
21
Another closely related argument is that the going private firms might be more
likely to be listed on the Alternative Investment Market (AIM) with lower listing
requirements which in turn would drive the visibility result. The last column in
Tables 5 and 6 reports AIM listing frequency among our firms and does not detect
any significant trend. Also, in an unreported regression, we include an AIM dummy
as an additional regressor. As the coefficients are not significant and the other results
remain unaffected we conclude that AIM listing does not drive our results.
Figures 1e and 1f provide visual intuition and further support for our findings.
They show clearly that the probability of management sponsored deals is sharply
increasing with low analyst coverage and thin trading. In contrast, this is not the
case at all for the private equity backed firms. Overall, our results support the
hypothesis that the going private firms are heterogeneous with respect to financial
visibility. Management sponsored deals seem to suffer both low analyst coverage and
infrequent trading and therefore have less reasons to remain publicly listed whereas
this is not the case for the private equity backed deals.
4.3
Free cash flows
Our third hypothesis conjectures that more cash rich firms are more likely to go
private via a management sponsored deal whereas low cash levels increase the probability of a private equity backed deal. The coefficients for our cash variable in Table
3 support the hypothesis as they show that, relative to the non-PtP control firms, the
management sponsored deals and the private equity backed deals have significantly
higher and lower cash levels, respectively. The coefficients are significant at the ten
and one percent level, respectively and are statistically different from each other at
the one percent level. Still, it is important to control for sales growth as Jensen
(1986) suggests that the agency problems of free cash flow concern mature firms
with low growth prospects. The sales growth variable, however, is not significant
implying that growth opportunities do not affect the probability of going private.
Thus, controlling for growth opportunities, our results suggest a sharp difference in
the effect of cash levels: management sponsored deals seem to be cash rich while
private equity backed deals suffer very low cash levels.
22
An alternative interpretation of the low cash levels of private equity backed deals
might be firm insolvency and inability to pay interest payments. To account for
this possibility, we check average interest coverage across deciles of cash levels of the
private equity backed firms. This exercise, however, shows that interest coverage
is not related to cash level of the private equity backed deals. Moreover, including
interest coverage in the regression does not result in a significant coefficient for the
private equity backed deals nor does this affect the cash coefficient.5 Therefore, we
conclude that this alternative explanation is not plausible for our setting.
As a sensitivity check for our measure of managerial opportunism, we also estimate excess cash recently proposed in the literature on determinants of cash reserves level (Opler et al., 1999 and Dittmar and Mahrt-Smith, 2007). In line with
this methodology, we estimate a cash regression that should determine normal cash
levels used in companies to cover their liquidity needs.6 Residuals of this regression then measure cash reserves held in excess of those needed for operations and
investments. These resources are most at risk of being wasted at the managers’
discretion (Dittmar and Mahrt-Smith, 2007). Note also that excess cash satisfies
the stock requirement of DeAngelo and DeAngelo (2006) as discussed in section 3.2
above and therefore is a very good candidate for the proxy that should reflect managerial opportunism in cash rich firms. In Table 4, we partition the cash variable
from Table 3 into two separate variables: normal cash and excess cash representing
the fitted values and residuals from the cash regression, repectively. The results are
equally strong now despite fewer observations due to data availability for the cash
regression. They show that high excess cash levels increase the probability of management sponsored deals whereas low excess cash levels are associated with private
equity backed deals. The regression in Table 4 includes also free cash flow and shows
5
All results are available upon request.
Following Opler et al. (1999) and Dittmar and Mahrt-Smith (2007) we regress the natural
logarithm of cash over net assets (total assets minus cash and marketable securities) on the natural
logarithm of net assets, market to book adjusted for net assets, net working capital over net assets,
capital expenditures over net assets, R&D over net assets, free cash flow over net assets, leverage
and a dividend dummy. All variables are industry and time adjusted. Due to missing data, we are
able to obtain only 1,437 observations for excess cash compared to 1,579 observations in our full
sample.
6
23
that free cash flow do not affect the decision to go private.
Finally, we include the payout ratio as an additional explanatory variable because firm dividend policy is closely related to the free cash flow hypothesis (Jensen,
1986): high dividend payout may mitigate the agency problems associated with free
cash flow. DeAngelo and DeAngelo (2006) provide a different argument. They combine capital structure decisions with payout policy and constraints on liquid assets
(cash) and claim that firms can develop three potential sources of future financial
flexibility: cash accumulation, preservation of debt capacity and reputation for stable and substantial equity payouts. Cash accumulation as argued above increases
managerial opportunism. High leverage reduces the agency costs but it utilizes debt
capacity and therefore also financial flexibility. High ongoing equity payouts control
agency costs without high leverage and thus preserve the firm’s option to borrow,
and they also increase its future equity issuance capability. Therefore, it is important to control for payout as well as leverage when analyzing liquidity position of
firms.
Table 3 shows that the two payout coefficients are positive but only the coefficient for the private equity backed deals is significant (at the ten percent level).
The coefficients for leverage are of opposite signs – private equity backed deals are
less leveraged and management sponsored deals are more leveraged relative to the
non-PtP firms – but neither of the coefficients is statistically significant at a conventional level. However, results in Table 4 are stronger: private equity backed firms
pay significantly more dividends relative to the non-PtP firms and at the same time
have lower leverage relative to both non-PtP firms and management sponsored deals
(all at the five percent level). This suggests that private equity deals pay high dividends that reduce managerial opportunism of excess cash and keep high reputation
to equity holders while, at the same time, preserve firm future borrowing potential.
Clearly, private equity backed deals do not seem to suffer from agency problems and
managerial opportunism. If anything, private equity backed going private transactions experience shortage of cash. Management sponsored deals, in contrast, enjoy
high excess cash levels while at the same time enjoying relatively common payout
24
and leverage. This indicates that management sponsored deals may have problems
with managerial opportunism. However, alternative and perhaps more plausible explanation for the high excess cash levels is that managers build up financial slack to
finance the deal.
Finally, we turn to the functional form of the probability functions shown in
Figures 2a to 2c. Figure 2a shows that as the cash levels increase, the probability
of a private equity backed deal decreases whereas the probability of a management
sponsored deal increases, showing a contrasting characteristic of the private equity
versus management sponsored deals. At the same time, the payout ratio in Figure 2b
has a more dramatic effect on the probability of a private equity backed deal relative
to a management sponsored deal. Hence, the results suggest that the management
sponsored deals are more likely in firms with higher cash levels that may suffer the
agency problems associated with excess cash. Alternatively, availability of high cash
levels may simply make a leveraged buyout more feasible. In contrast, private equity
backed deals are associated with negative excess cash levels, high dividend payments
and low leverage.
4.4
Ownership structure
Our final hypothesis highlights the heterogeneity in pre-transaction ownership and
proposes that high executive ownership may increase chances of a management sponsored deal whereas high ownership by financial institutions increases chances of a
private equity backed deal. Table 3 shows that the coefficient for executive ownership in the management sponsored deals is indeed positive and significant at the one
percent level indicating that high executive ownership increases the probability of a
management sponsored transaction relative to the firm staying public. However, executive ownership also increases the probability of a private equity backed deal. The
corresponding coefficient is positive and significant at the five percent level. This
supports the view that private equity investors look for support from management in
their deals and choose firms with relatively high managerial ownership. Even though
the MS coefficient is slightly larger, the difference between the two coefficients is not
statistically significant. So, higher executive ownership increases the chances of both
25
private equity backed as well as management sponsored deals.
In contrast, we find significant differences with respect to ownership by financial
institutions, confirming the second part of our hypothesis. High ownership by financial institutions increases the probability of a private equity deal relative to both
non-PtP firms as well as management sponsored deals (both statistically significant
at the one percent level). These results are visualized in Figures 2c and 2d. In particular, the probability of management sponsored as well as private equity backed
deals increase with executive ownership. However, only the probability of private
equity backed deals increases with ownership by financial institutions.
We also check for ownership concentration measured by the Herfindahl index with
the expectation that the management sponsored deals have higher concentration of
ownership by the management as other blockholders in these deals are relatively
small. The results suggest that ownership concentration matters: it is significantly
higher for the management sponsored transactions and lower for the private equity
backed deals.7
In summary, our heterogeneity in ownership hypothesis is partially supported.
Our results suggest that a management sponsored deal is more likely when the managers have higher control and concentration of ownership. At the same time, however, also private equity backed deals are more likely in firms with high managerial
ownership suggesting cooperation between private equity investors and managers in
the deals. The heterogeneity in pre-transaction ownership structure stems mainly
from the differences in ownership by financial institutions. Willingness of the institutional investors to accept a deal proposed by a private equity party seems to
increase the success of the transaction and thus motivates private equity investors
to proceed with the deal in the first place. At the same time, very low ownership
by financial institutions in the management sponsored deals may also reflect low
investor interest for these deals that was discussed in section 4.2 above.
In addition to size, our model includes two other control variables: tax and
standard deviation of returns. The argument for taxes is that highly leveraged
7
All results are available upon request.
26
buyout transactions are associated with positive interest tax shield that may further
increase incentives for going private. Our results in Table 3 suggest that taxes
increase the odds of going private for the management sponsored deals but this is
not supported in Table 4. The standard deviation of returns seems to be associated
with a higher probability of a private equity backed deal (significant at the five
percent level). This confirms the Gompers and Lerner (2001) suggestion that targets
of private equity deals are associated with higher uncertainty and riskiness.
5
Concluding remarks
This paper proposes that the recent increase of private equity investors’ involvement
in going private transactions may affect sources of value creation and reasons explaining why firms decide to go private. Our analysis shows that the probability
whether a deal is backed by private equity investment or is purely managed by insiders of the firm depends on different firm characteristics. Drawing upon several
theoretical arguments (including Jensen’s free cash flow hypothesis, ownership structure, and the framework of costs and benefits of being publicly listed versus privately
owned commonly applied in the IPO literature) we derive four testable hypotheses
explaining reasons for management sponsored PtP deals: (i) undervaluation, (ii)
financial visibility, (iii) free cash flow, and (iv) pre-transaction ownership structure.
At the same time, we propose that private equity backed deals do not share the same
characteristics as they are usually backed by strategic reasons and are not able to
raise funds via traditional market sources. Therefore, pooling all PtP firms together
may lead to weak estimation results and fail to detect the important determinants
motivating management sponsored as well as private equity backed deals.
In summary, our empirical results for the UK provide convincing evidence that
the population of going private firms is indeed heterogeneous. We show that private
equity backed and management sponsored deals have different reasons for their decision to take a firm from the stock market. Firms involved in management sponsored
deals are significantly undervalued, not followed by analysts, have relatively high
cash levels, high executive ownership, and ownership concentration. The private
27
equity backed deals, in contrast, have high executive ownership and high ownership
by financial institutions but have lower ownership concentration. These firms have
shortage of cash, low debt levels, and pay high dividends. Thus, our results suggest
that the management sponsored deals fit the characteristics of the Jensen’s cash rich
firms with strong management that are relatively thinly traded in the stock market,
not followed by analysts and are significantly undervalued. This suggests that the
benefits of remaining publicly listed fall short of the costs and the firms are more
likely to go private. In contrast, benefits of stock market listing seem to be larger
for the private equity backed deals as they are more actively traded, followed by
analysts and not so much undervalued. However, they seem to be short of cash.
Private equity backing may provide the necessary financing and extra know-how for
their strategic restructuring.
28
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Realignment, Private Equity and the Decision to Go Private: Public to Private Activity
in the UK,” working paper, Aberdeen Business School.
Wright, M., C. Weir, D. Laing, and A. Burrows, 2007, “Irrevocable Commitments and
Going Private”, European Financial Management, forthcoming.
30
Table 1: Variable Definitions
Total assets
Total debt
ROA
St. dev. of stock returns
Market to book
Rumours
Sales growth
Director buying
Director selling
Analysts following
Thin trading
Ownership of
executives
non-executives
financial inst.
other firms
individuals
Herfindahl index
Cash
Free cash flows
Investment
Payout ratio
Tax
AIM
total assets (in millions)
total debt divided by total assets
net income divided by total assets
standard deviation of stock returns over the period
from January to December of the calendar year before PtP transaction
market capitalization plus total debt divided by total
assets
number of takeover rumours during two calendar
years before PtP transaction
sales growth during 3 financial years before PtP
transaction average
dummy variable that is set to one in case executive
directors were buying shares of their own firm during
January to December of the calendar year before PtP
transaction or in the previous year for the non-PtP
firms and zero otherwise
dummy variable that is set to one in case executive
directors were selling shares of their own firm during
January to December of the calendar year before PtP
transaction or in the previous year for the non-PtP
firms and zero otherwise
number of analysts following the company in December of the calender year before PtP transaction
fraction of days with zero percent return during January to December of the calendar year before PtP
transaction or in the previous year for the non-PtP
firms
Worldscope
Worldscope
Worldscope
Datastream
percentage of shares held by executive directors of
the company
percentage of shares held by non-executive directors
of the company
percentage of shares held by financial institutions
(e.g. pension funds, mutual funds, insurance companies, banks, venture capitalists)
percentage of shares held by industrial firms
percentage of shares held by persons that are not
directors of the company
sum of squared equity stakes held by the individual
blockholders
cash and marketable securities divided by total assets
(ebitda - taxes - interest - cash dividend - stock repurchases) divided by sales
capital expenditures divided by sales
cash dividend divided by the sum of net income and
depreciation
income taxes divided by sales
dummy variable that is set to one in case the firm
is listed on the alternative market with lower listing
requirements
Corporate Register
31
Worldscope
Lexis Nexis and
SDC M&A
Worldscope
Hemmington Scott
Hemmington Scott
IBES
Datastream
Corporate Register
Corporate Register
Corporate Register
Corporate Register
Corporate Register
Worldscope
Worldscope
Worldscope
Worldscope
Worldscope
Corporate Register
Table 2: Comparison of types of firms
Variable
non-PtP
Means
PtP
PE
MS
Oth.
t-test p-values
non-PtP PE
PE
PtP
MS Oth.
MS
Oth.
Panel A
Total assets
Total debt
ROA
St.dev. of return
Market to book
Rumours
Sales growth
Director buying
Director selling
Analysts following
Thin trading
Ownership of
executives
non-executives
financial inst.
other firms
individuals
Herfindahl index
Cash
Free cash flows
Investment
Payout ratio
Tax
Number of obs.
692.4
0.199
−0.019
15.38
1.528
0.322
0.154
0.341
0.169
3.028
0.530
223.3
100.8
300.8
0.190
0.213
0.223
0.031 −0.019 −0.020
15.50
15.46
14.90
1.050
0.702
0.818
0.778
0.493
0.604
0.129
0.126
0.053
0.267
0.377
0.321
0.167
0.072
0.170
3.122
1.913
2.811
0.566
0.733
0.618
0.000
0.303
0.103
0.443
0.000
0.000
0.021
0.231
0.109
0.011
0.000
0.008
0.210
0.087
0.479
0.000
0.041
0.475
0.073
0.032
0.000
0.000
0.211
0.157
0.055
0.203
0.010
0.151
0.046
0.250
0.481
0.215
0.074
0.013
0.396
0.490
0.246
0.052
0.227
0.093
0.261
0.057
0.012
0.001
0.088
0.121 0.098
0.198
0.060
0.032
0.029 0.026
0.039
0.020
0.187
0.243 0.302
0.170
0.237
0.030
0.034 0.020
0.024
0.072
0.124
0.139 0.072
0.165
0.218
0.076
0.106 0.078
0.114
0.143
0.132
0.114 0.085
0.140
0.130
−0.084 −0.036 0.033 −0.130 −0.029
0.063
0.079 0.026
0.095
0.149
0.188
0.195 0.286
0.190
0.045
0.020
0.022 0.022
0.019
0.026
0.006
0.215
0.000
0.287
0.141
0.000
0.069
0.117
0.225
0.417
0.214
0.001
0.122
0.000
0.345
0.000
0.008
0.017
0.054
0.046
0.060
0.293
0.080
0.241
0.039
0.004
0.000
0.004
0.051
0.119
0.028
0.005
0.235
0.000
0.044
0.030
0.008
0.094
0.132
0.389
0.179
0.225
0.066
0.144
1400
202.8
0.206
0.002
15.34
0.879
0.642
0.109
0.316
0.137
2.651
0.633
212
90
69
53
continued on next page
32
Variable
non-PtP
PtP
Medians
PE
MS
Oth.
continued from previous page
U -test p-values
non-PtP PE
PE
MS
PtP
MS Oth. Oth.
Panel B
Total assets
Total debt
ROA
St.dev. of return
Market to book
Rumours
Sales growth
Director buying
Director selling
Analysts following
Thin trading
Ownership of
executives
non-executives
financial inst.
other firms
individuals
Herfindahl index
Cash
Free cash flows
Investment
Payout ratio
Tax
67.01
0.173
0.044
14.54
1.013
0.000
0.071
0.000
0.000
2.000
0.571
65.18
0.177
0.038
14.42
0.764
0.000
0.048
0.000
0.000
2.000
0.663
66.50
0.172
0.057
14.62
0.830
0.000
0.052
0.000
0.000
3.000
0.598
51.05
0.170
0.038
14.64
0.670
0.000
0.028
0.000
0.000
1.000
0.750
125.3
0.187
0.020
13.90
0.769
0.000
0.041
0.000
0.000
2.000
0.663
0.488
0.493
0.349
0.369
0.000
0.000
0.013
0.275
0.228
0.036
0.000
0.015
0.399
0.002
0.288
0.000
0.107
0.238
0.117
0.155
0.000
0.000
0.123
0.293
0.000
0.084
0.040
0.430
0.049
0.295
0.487
0.146
0.013
0.001
0.385
0.054
0.255
0.058
0.160
0.155
0.298
0.179
0.005
0.002
0.015
0.002
0.151
0.000
0.055
0.047
0.070
0.048
0.027
0.180
0.015
0.027
0.001
0.221
0.000
0.064
0.065
0.050
0.041
0.025
0.207
0.013
0.020
0.001
0.306
0.000
0.031
0.050
0.045
0.050
0.024
0.235
0.018
0.122
0.004
0.112
0.000
0.111
0.092
0.076
0.029
0.022
0.208
0.013
0.009
0.002
0.189
0.000
0.110
0.088
0.039
0.046
0.030
0.167
0.008
0.001
0.491
0.000
0.423
0.227
0.000
0.030
0.082
0.215
0.091
0.236
0.000
0.085
0.000
0.304
0.000
0.000
0.060
0.045
0.241
0.083
0.151
0.002
0.406
0.024
0.006
0.001
0.001
0.407
0.242
0.141
0.018
0.218
0.000
0.105
0.054
0.023
0.349
0.493
0.164
0.248
0.401
0.193
0.478
Number of obs.
1400
212
90
69
53
Note: This table shows the means and medians across non-PtP, PtP firms as well as private equity backed (PE), management
sponsored (MS) and other (Oth.) deals. The last four columns show p-values for a t-test for equal means allowing for unequal
variances in Panel 1 and a Mann-Whitney U-test for equal medians in Panel 2. All variables are trimmed at the 1st and 99th
percentiles, except for the ownership and illiquidity variables. See Table 1 for variable definitions.
33
Table 3: Multinomial Logistic Regression Analysis of Factors Influencing the Likelihood of Going Private Transactions
Variable
Market to book
Private equity
backed deals
coeff.
s.e.
−0.293 (0.170)∗
Management
p-value for LR test
sponsored deals
Other deals
of equal parameters
coeff.
s.e.
coeff.
s.e.
PE-MS PE-Oth MS-Oth
−0.887 (0.243)∗∗∗ −0.512 (0.233)∗∗
0.038
0.431
0.251
Rumours
0.414
(0.099)∗∗∗
0.440
(0.138)∗∗∗
Sales growth
0.010
(0.360)
0.282
(0.368)
ROA
1.530
(1.046)
0.419
(0.779)
Analysts following
Thin trading
Cash
Payout ratio
Total debt
34
Executive ownership
Financial inst. own.
Size
0.009
0.268
−2.324
0.565
−1.034
2.044
(0.050)
(0.800)
∗∗
−0.134
4.713
∗
(0.070)
0.321 (0.135)∗∗
0.865
0.537
0.513
−1.085 (0.695)
0.586
0.128
0.054
−0.216 (0.806)
0.375
0.176
0.564
0.027 (0.062)
0.079
0.814
0.071
∗∗∗
3.075 (1.000)
0.000
0.026
0.213
∗
(0.951)
∗∗∗
(1.150)
1.589
(0.854)
1.445 (0.958)
0.004
0.009
0.906
(0.293)∗
0.172
(0.340)
−0.585 (0.256)∗∗
0.372
0.003
0.058
(0.825)
0.718
(0.796)
1.400 (0.868)
0.113
0.038
0.543
(0.805)∗∗
2.676
(0.692)∗∗∗ −1.608 (1.393)
0.528
0.013
0.001
(0.812)
0.094 (0.830)
0.000
0.001
0.814
0.278 (0.146)∗
0.078
0.114
0.908
∗∗∗
3.247
(0.618)
−0.170
−0.017
(0.122)
0.301
(0.140)∗∗
12.060
∗∗
(5.030)
15.715 (4.743)
0.330
0.114
0.573
0.045
(0.034)
−0.012 (0.041)
0.463
0.074
0.260
0.000
0.000
0.000
Tax
5.668
(4.600)
St.dev. of returns
0.079
(0.033)∗∗
∗∗∗
No heterogeneity test
0.000
0.000
0.000
IIA test
1.000
1.000
1.000
Note: The table reports estimation results for the multinomial logistic regression model given in (1), using the non-PtP firms as
reference group. The model is estimated using 212 observations for UK PtP deals over the period 1997-2003. All regressors are
industry and time adjusted as they enter the multinomial logit model as deviations from the industry and year median. The numbers
of observations in the PtP groups are 90 for PE-backed deals, 69 for MS-deals, and 53 for other deals. The non-PtP group consists of
1400 observations. Standard errors are given in parentheses, with ∗∗∗ , ∗∗ , and ∗ indicating significance at the 1%, 5% and 10% level,
respectively. The final three columns show p-values for the LR test of equal parameters across two sub-groups of PtP deals. The line
“No heterogeneity test” reports p-values for the LR test of Cramer and Ridder (1991) that all parameters except the intercepts are
equal for two groups. The first three numbers in this line compare the non-PtP group with one of the PtP groups. The line “IIA test”
reports p-values for the Hausman and McFadden (1984) LR test for the validity of the independence of irrelevant alternatives (IIA)
assumption, omitting the indicated group from the model. Variable definitions are provided in Table 1.
Table 4: Multinomial Logistic Regression Analysis of Factors Influencing the Likelihood of Going Private Transactions
Variable
Market to book
Private equity
backed deals
coeff.
s.e.
−0.339 (0.234)
Management
sponsored deals
Other deals
coeff.
s.e.
coeff.
s.e.
∗∗
−0.664 (0.287)
−0.409 (0.329)
p-value for LR test
of equal parameters
PE-MS PE-Oth MS-Oth
0.367
0.860
0.560
Rumours
0.411
(0.112)∗∗∗ −0.013
(0.307)
0.276 (0.170)
0.144
0.452
0.387
M/B × Rumours
0.051
(0.121)
−1.021
(0.527)∗
0.035 (0.207)
0.024
0.946
0.047
Director buying
−0.199
(0.298)
0.266
−0.320 (0.407)
0.288
0.805
0.254
Director selling
0.048
(0.360)
−1.424
0.342 (0.462)
0.047
0.610
0.028
Sales growth
0.373
(0.346)
0.369
(0.402)
−1.019 (0.832)
0.992
0.074
0.092
−0.323
(1.118)
0.263
(0.803)
−0.608 (1.541)
0.661
0.880
0.626
(0.089)
0.042 (0.076)
0.120
0.596
0.070
(1.167)∗∗∗
2.083 (1.203)∗
0.015
0.232
0.292
∗
ROA
Analysts following
Thin trading
−0.007
(0.056)
−0.164
0.306
(0.904)
3.807
∗
(0.334)
(0.755)∗
∗
∗
35
Excess Cash
−1.249
(0.659)
0.588
(0.318)
0.686 (0.356)
0.006
0.006
0.830
Normal Cash
−0.103
(0.079)
−0.012
(0.093)
0.116 (0.117)
0.444
0.104
0.378
FCF
0.335
(0.740)
0.050
(0.220)
2.377 (1.476)
0.688
0.200
0.050
Payout ratio
0.720
(0.302)∗∗
0.080
(0.381)
−0.628 (0.283)∗∗
0.190
0.002
0.119
−3.122
∗∗
(1.556)
0.988
(1.215)
−0.670 (1.610)
0.031
0.267
0.399
2.638
(0.938)∗∗∗
2.793
(0.824)∗∗∗ −3.380 (2.220)
0.897
0.003
0.001
Financial inst. own.
3.605
∗∗∗
(0.694)
−0.472
(1.038)
0.717 (0.975)
0.001
0.012
0.394
Size
0.105
(0.136)
0.123
(0.170)
0.130 (0.187)
0.931
0.912
0.978
Tax
7.688
(5.242)
8.296
(6.200)
−7.381 (7.501)
0.939
0.091
0.100
St.dev. of returns
0.080
(0.038)∗∗
0.052
(0.041)
0.039 (0.046)
0.610
0.492
0.834
0.000
0.000
0.000
Total debt
Executive ownership
No heterogeneity test
0.000
0.000
0.005
IIA test
1.000
1.000
1.000
Note: The table reports estimation results for the multinomial logistic regression model given in (1), using the non-PtP firms as
reference group. The numbers of observations drop to 70 for PE-backed deals, 49 for MS-deals, and 35 for other deals. The non-PtP
group consists of 1283 observations.Standard errors are given in parentheses, with ∗∗∗ , ∗∗ , and ∗ indicating significance at the 1%, 5%
and 10% level, respectively. See Table 3 for further details.
Table 5: Summary Statistics of Firms Grouped by Size
Decile
1 (small)
2
3
4
5
6
7
8
9
10 (large)
Total
Size
4.17
12.36
21.22
34.78
53.43
84.44
142.15
266.93
731.44
6912.01
PtP(%)
5.0
11.7
13.7
16.7
20.4
16.8
14.2
17.4
14.2
2.5
828.11
13.2
Mean by deciles of size
Thin
PE(%) MS(%) trading
2.5
2.5
0.79
5.6
4.9
0.72
3.1
5.6
0.69
7.4
4.9
0.66
8.6
8.6
0.59
6.8
6.8
0.59
5.6
4.9
0.54
7.5
3.1
0.46
8.0
1.9
0.29
0.6
0.0
0.10
5.6
4.3
0.54
Analysts
following
1.77
2.06
1.96
2.27
2.86
3.27
3.20
3.56
4.40
4.36
2.97
AIM
0.12
0.15
0.08
0.11
0.07
0.13
0.12
0.15
0.13
0.14
0.12
Note: The table reports mean values of several variables across size deciles where size is measured by
total assets. Variable definitions are provided in Table 1.
36
Table 6: Summary Statistics of Firms Grouped by Size
Decile
Size
Mean by deciles of size
Thin
Analysts
trading
following
AIM
Panel A: Non-PtP firms
1 (small)
2
3
4
5
6
7
8
9
10 (large)
4.09
12.32
21.49
34.97
53.42
84.80
141.75
268.83
734.29
7020.87
0.79
0.72
0.68
0.66
0.57
0.58
0.53
0.45
0.27
0.10
1.79
2.10
2.03
2.25
2.95
3.41
3.36
3.62
4.39
4.35
0.15
0.15
0.09
0.11
0.06
0.13
0.13
0.11
0.12
0.13
2.00
2.33
2.40
2.42
2.43
3.09
3.00
4.25
4.69
4.00
0.25
0.11
0.00
0.08
0.07
0.09
0.22
0.17
0.15
0.00
0.75
1.25
0.78
2.00
2.21
2.27
1.88
2.40
4.33
−−
0.25
0.12
0.00
0.12
0.07
0.09
0.00
0.00
0.33
−−
Panel B: Private equity backed PtP deals
1 (small)
2
3
4
5
6
7
8
9
10 (large)
5.72
12.46
19.19
36.41
54.64
82.19
136.51
270.71
769.57
3281.60
0.71
0.72
0.70
0.60
0.52
0.60
0.60
0.45
0.46
0.25
Panel C: Management sponsored PtP deals
1 (small)
2
3
4
5
6
7
8
9
10 (large)
5.37
12.44
19.33
31.73
52.89
80.27
148.03
228.58
825.51
−−
0.91
0.76
0.78
0.79
0.76
0.74
0.64
0.66
0.43
−−
Note: The table reports mean values of thin trading, analysts following
and frequency of AIM listing across size deciles for the non-PtP firms
(Panel A), private equity backed (Panel B), and management sponsored
deals (Panel C). Size is measured by total assets. Variable definitions are
37
provided in Table 1.
.06
.30
PE-backed
Management-sponsored
Other
.05
PE-backed
Management-sponsored
Other
.25
.04
.20
.03
.15
.02
.10
.01
.05
.00
.00
-4
-2
0
2
4
6
8
10
12
0
1
2
3
(a) Market to book
4
5
6
7
8
9
10 11 12
(b) Rumours
.020
.020
PE-backed
Management-sponsored
Other
.016
.016
.012
.012
.008
.008
.004
.004
.000
-0.8
-0.4
0.0
0.4
0.8
1.2
1.6
2.0
.000
-2.0
2.4
(c) Sales growth
PE-backed
Management-sponsored
Other
-1.6
-1.2
-0.8
-0.4
0.0
0.4
(d) ROA
.024
.20
PE-backed
Management-sponsored
Other
.020
.16
PE-backed
Management-sponsored
Other
.016
.12
.012
.08
.008
.04
.004
.00
-0.8 -0.6 -0.4 -0.2
.000
-8
-6
-4
-2
0
2
4
6
8
10
(e) Analysts following
0.0
0.2
0.4
(f) Thin trading
Figure 1: Multinomial Logit probabilities
38
0.6
0.8
1.0
.030
.030
PE-backed
Management-sponsored
Other
.025
.020
.020
.015
.015
.010
.010
.005
.005
.000
-.3 -.2 -.1
.0
.1
.2
.3
PE-backed
Management-sponsored
Other
.025
.4
.5
.6
.7
.000
-3.0 -2.5 -2.0 -1.5 -1.0 -0.5 0.0 0.5 1.0 1.5 2.0
.8
(a) Cash
(b) Payout ratio
.05
.09
.08
PE-backed
Management-sponsored
Other
.04
.07
PE-backed
Management-sponsored
Other
.06
.03
.05
.04
.02
.03
.02
.01
.01
.00
.00
-.1
.0
.1
.2
.3
.4
.5
.6
.7
.8
.9
-.3 -.2 -.1
(c) Exec.own
.0
.1
.2
.3
.4
(d) Fin.inst.own
Figure 2: Multinomial Logit probabilities
39
.5
.6
.7
.8