Journal of Financial Economics 34 (1993) 345-371. North-Holland
The interma
corporate control
Mergers and acquisitions OFU.S. firms
anese firms*
Jun-Koo Kang
The University of Rhode Island, Kingston. RI 02881, USA
RboceivedOctobet 1991, final version received March 1993
Japanese mergers and acquisitions in the U.S. create statistically significant wealth gains for both
Japanese bidders and U.S. targets. Consistent with the literature on foreign direct investment and
the market for corporate control, bidder-specific characteristics and exzhange-rate movements are
useful in explaining the cross-sectional variation in bidder returns: returns to Japanese bidders and
to the portfolio of Japanese bidders and U.S. targets increase with the bidder’s leverage, the bidder’s
ties to financial institutions through borrowings, and the depreciation of the dollar in relation to the
Japanese yen.
Key words: Japanese mergers and acquisitionq Monitoring Exchange-rate movement; Bidder return
1. Introduction
Japanese mergers and acquisitions in the U.S. have surged in recent years,
both in dollar amounts and in the number of transactions. Kester (1991) reports
that in 1984 Japanese firms spent about $1.4 billion to acquire 39 U.S. targets.
This compares with a total cost of over $12.6 billion in 1988 for 132 U.S. targets.
In spite of the large transaction volume and the controversy surrounding
Japanese takeovers of U.S. rms, Japanese merger and acquisition activity in the
Correspondence to: Jun-Koo Kang, Department of Finance and Insurance, College of Business
Administration, The University of Rhode Island, Kingston, RI 02881-0802, USA.
*This paper is based on my Ph.D. discertation at the Ohio State University. I am indebted to ReEl
Stulz, my advisor, for his valuabfe comments. I am also grateful to Warren Bailey, Kenneth
Borokhovich, KC. Chan, Margaret Forster. Michael Jensen (the editor), Carl Kester (the refze).
Francis Langstaff, David Mayers, Ghon Rhee, Anil ShivdasanL and Ralph Wal line for their useful
s ggestions. All errors are my own respansibilrty.
0304405x/93/$06.00
c
1993- Eisevier Science Publishers B.V. All rights reserved
346
.I.-K. Kang , The interm rroual markee fbr corpra te control
U.S. is not fully understood. In particular, there is little evidence on the forces
driving Japanese mergers and acquisitions and their effects on the market value
of Japanese bidders and US. targets.
The literature on foreign direct investment (FDI) and the market for corporate
control suggests that foreign mergers and acquisitions are motivated by several
factors, such as imperfections in product and factor markets [Kindleberger (1969),
Caves (1971), and Hymer (1976)], imperfections and asymmetries in capital
markets [Froot and Stein (199 l)], differences in tax codes [Scholes and Wolfson
(1990)], znd incumbent management that acts in its own interest to the detriment
of shareholders [Jensen (1986)]. Because of the unique financial characteristics of
Japanese firms, the depreciation of the dollar against the Japanese yen, and
tax-law changes in the 198Os,data on mergers and acquisitions of U.S. firms by
Japanese firms provide an opportunity to examine these and other issues related
to FDI. Specifically, I attempt to provide evidence relevant to the following
questions: What drives Japanese mergers and acquisitions in the U.S.? What are
the wealth effects for the Japanese bidders acquiring U.S. firms, and *&hatexplains
the cross-sectional variation in these returns?’ EgoJapanese bidders pay different
premiums for their U.S. targets than U.S. bidders? To address these issues,
I examine a sample of 119 Japanese bidders and 102 U.S. targets during the
1975-1988 period, and a control sample of 119 U.S. bidders and 102 U.S. targets
involved in domestic mergers and acquisitions during the same period.
A key aspect of Japanese financial practices in this context is that some firms
maintain close financial ties with large banks and receive substantial equity
and/or debt financing from them. These links can reduce the agency costs in
a firm’s financial structure and thus influence its investment decisions [Nakatani
t i 984), Prowse (1990), and Hoshi, Kashyap, and Scharfstein (1991)].
Whereias the Glass-Steagall Act of 1933 prevents U.S. banks from owning and
dealing in voting stocks,2 Japanese banks are allowed to invest in equity of other
companies. In 1987, equity ownership by Japanese banks and insurance companies accounted for about 42.2% of the shares iisted on the Tokyo Stock
Exchange. According to Shleifer and Vishny =!1986)and Jensen (1989), this
concentrated equity ownership can provide strr.,ng incentiv/cb for Japanese
financial institutions to monitor the performance of management.3 The large
‘See Jensen and Ruback (1983) and Jarrell, Brickley, and Netter (1988) for discussions of and
evidence on announcement returns in domestic: mergers and acquisitions.
‘Bank trust dep artments are commercial banks’ only direcl: link to equity and no more than 10%
of a bank’s trust fund may be invested in the stocks of any single corporation. See Roe (1990) for
a detailed discussion.
3Jensen ( 1989) argues that the monitoring role of Japaccse banks has changed over time as a result
of several factors. For example, by i9”U7,Japanese banks are required to reduce their equity holdings
of any company to no more than 5% (Article f i cf the Revised Anti-Monopoly Act of 1.~77).This
new legal constraint reduces the bank’s incentive to engage in active monitoring. Furthermore, the
development of capital markc G and the large cash balance held by Japanese firms in rp’: :nt years
have left corporate managers increasingly unconstrained.
J.-K. Kang, The internatronal market for corporate control
347
debt and bank loans can also reduce managerial discretion over the allocation of
the free cash flow [Jensen(1986)] and provide a mechanism for monitoring the
conflict between debtholders and shareholders [Fama (a*990)].As Fama argues,
bank loans allow CC monitoring of the conflict between debtholders and
shareholders to be largely delegated to banks. In addition, some Japanese firms
have one commercial bank among their maior stockholders, called the main
bar:k, w&h provides both short-term and -long-term financing.4 This main
bank lea& the loan consortium when a group of banks extends major Ion-term
credit to the firm, and is responsible for monitoring the firm closely. Since the
main bank holds a large equity and a large debt position in the same firm, it has
a strong incentive to maximize the firm’s value. In this regard, the main bank
acts much like the U.S. leveraged buyout (LBO) partnership [Jensen jl989)].
Thus, I first investigate whether large equity and,/or debt positions by Japanese
financial institutions effectively deter investments that reduce the client firm’s
market value.
Another set of explanations for FDI is based on imperfections and asymmetries in capital markets, Froot and Stein (1991) show that when globally integrated capital markets are subject to informational imperfections, there shou!d
be a link between exchange-rate changes and FDI. Since FDI, such as mergers
and acquisitions, involves significant informational asymmetries about an asset’s payoffs, these asymmetries make external financing more expensive than
internal financing. As a consequence, entrepreneurs find it costly or impossible
to finance the asset solely with externally obtained funds. The more net wealth
firms can invest in mergers and acquisitions, the lower will be their cost of
capital Since Japanese firms hold more of their wealth in yen-denominated
assets, a depreciation of the dollar increases their relative net weaith position
and in turn lowers their relative cost of capital. Therefore, Froot and Stein’s
model predicts that a weajc dollar gives Japanese firms a competitive edge in
purchasing advantages and leads ths U.S. to become an important host for
Japanese FDI. If the stock-price effect of FDI reflects its net present value
(NPV), their model also implies that returns to Japanese bidders increase with
depreciatio:q of the dollar because of cost-of-capital advantages. Furthermore,
the favorabLe production environment resulting from the yen’s appreciation
means that costs for manufacturing, such as labor, parts, and land in the U.S.,
become rela.tively lower for Japanese firms. Since the dollar dropped 42%
against the yen i atween 1976 and 1978, and another sharp increase in the
relati? Q114ue of the yen began in late 1985, the elect of exchange-rate movements on Japanese bidder returns is likely to be substantial.
4For some Japanese firms, this main-bank relationship is part of an industrial group known as the
keiretsu, where the group bGifk plays the leading sole in tLe hnancial activities within the group
[Aoki (1984), Nakatani ( 1984). Hodder and Tschoegl(l98Q Kester ( 1986), and Hoshi, Kashyap, and
Sci-arfstein (1991) 3.
348
J.-K. Kang, The international market for corporate cowrol
Changes in U.S. tax laws in the 1980s may also affect Japanese merger and
(M&A)
activity in the U.S. Scholes 2:ld Wolfson (1990) argue that
while the Economic Recovery Tax Act (ERTA) of 1981 discouraged M&A
w1U.S. sellers and foreign buyers, it increase the demand for
transactions betwemr
transactions among U.S. firms. In particular, accelerated depreciation schedu
and more generous investment tax credits introduced in the ERTA of 1981
increased ta*<incentives for U.S. buyers, but put foreign buyers at a disadvantage. &holes and Wolfson also argue that this disadvantage was removed by the
Tax Reform Act (TRA) of 1986, which reduced the marginal corporate tax rate
in the U.S., making the U.S. a tax haven for many European and Japanese firms
that face higher corporate tax rates in their home countries. The elimination of
investment tax credits and accelerated depreciation schedules further increases
this relative tax advantage for foreign firms. As a result, investors in such
countries may place a higher value on U.S. assets than U.S. investors do.
Consistent with their arguments, Scholes and Wolfson provide evidence that
foreign acquisitions of U.S. firms decrease after 1981 and increase substantially
after the fourth quarter of 1986. Kester (1991j also shows that Japanese mergers
and acquisitions in the U.S. surge after 1986. If tax benefits are reflected in the
takeover gains, Japanese bidder returns are JiLiy to bc bzx in the 1981-1986
period and higher after 1986.
Although the empirical work on the market for corporate control shows that
targets realize significant gains from domestic acquisitions [Jensen and Ruback
(1983) and Jarrell, Brickley, and Netter (1988)], it is not known whether the
wealth gains differ for targets of Japanese and U.S. firms. Since the theory of
FDI posits that imperfections in product markets, factor markets, and capital
markets give multinatior4 firms a competitive advantage over local firms in the
host county [Kindlebergei <1969),Caves (1971), Hymer (1976), and Froot and
Stein (1991)], cross-border acquisitions are likely to create more wealth than
domestic acquisitions. Since targets tend to reap most of the benefits of the
acquisitions, the theory suggests that wealth gains to targets of Japanese firms
are larger than those :o targets of U.S. firms. Harris and Ravenscraft (1991) find
that U.S. targets of foreign buyers have significantly larger wealth gains than do
those of U.Sy firms.
The empirical results indicate that Japanese acquisitions of U.S. firms create
statistically significant wealth gains for both bidders and targets. Consistent
with both Jensen’s (1986) an3 Fama’s (1990) hypotheses, I find that returns to
Japanesebidders and returns realized by the portfolio of Japahiese bidders and
their U.S. targets are positively related to the bidders’ total debt and their
borrowings from financial institutions. I also find a positive and significant
relationship between Japanese bidder gains and the appreciation, of the yen,
which supports Fmot and Stein’s (1991) model. ? do not find, however, that
changesin U.S. tax laws affect the returns to both U.S. and Japanese bidders.
Finally, 1 find that the comparative returns to U.S. targets of 3
acquisition
J.-K. Kang, The international market $w corprate control
349
depend 012the type of acquisition (i.e., acquisition of ;dmajority interest, unit, or
minority interest). U.S. targets of Japanese F-qs realize the greatest differential
returns when they sell a majority interest to Japanese bidders, implying that the
ccntrol of target resources by Japanese bidders is an important determinant of
cross-border takeover premiums. The analyses for subperiods, however, indicate
that the main &dings for both Japanese bidders and U.S. targets are driven by
the last three years of transactions (19861988).
The remainder of the paper is organized as follows. Section 2 describes the
data and methodology. Empirical results are reported in section 3. A summary
and concluding remarks are presented in section 4.
2.1. Data
The sample consists of Japanese bidders and U.S. targets engaged in international mergers and acquisitions between 1975 and 1988. I obtain the sample
from the Foreign Acquisitions Roster of Mergers and Acquisitions
(1975-1988),’ Fxeign Direct Investment in the United States: Completed
Transacti.ons published by the Department of Commerce (3975-1983), and the
Dow Jones News Retrieval Service (1979-1988). The Wall Strzet Journal (WSJ)
is used Lx the announcement date. I confine the sample to those Japanese
bidders and U.S. targets fer which stock-price riata are available on the daily
returns file tram ths Pacific-Basin Capital Markets (PACAP) Research Center6
and the daily returns file from the Center for Research in Security Prices (CRSP),
respectively. This restriction results in a sample of 119 Japanese bidders (93
nonfinancial bidders) and 102 U.S targets. Among these bidders and targets,
daily stock prices are available for 68 matched nonfinancial bidder and target
pairs.
Table 1 summarizes Japanese M&A activity in the U.S., by year and by type.
It also reports the total and average bid prices for transactions that disclose the
price. Each sample is divided into three categories: for Japanese bidders, 30 case?
are mergtxs and tender offers, 30 cases are partial acquisitions (less than 50%) of
iDdependent firms, and 59 cases are acquisitions of units. For U.S. targets, 21
cases are mergers and tender offers, 26 cases are partial sales of independent
‘The sample does not include transactions with a value of less than $1 million and real estate
investments made by Japanese firms. This is because the Foreign Acquisitions Roster of Mergers
and Acquisitions, which is the paper’s primary source, does not list thest: types of transactions. In
addition, the sample does not include acquisitions of nonvoting securities by Japanese firms.
‘Tine PACAP daily returns tape contains daily stock returns of all Japanese firms listed on the
First Section of the Tokyo Stock Excha:r;lge.
J.-K. Kang, The international market for corporate control
350
Table 1
Japanese merger and acquisition activity in the United States, stratified kly year and by type
(1975-1988). The sampie consists of 119 Japanese bidders listed on the Tokyo Stock Exchange and
102 U.S. targets listed on the NYSE, AMEX, or OTC. Total and average bid prices are computed for
the transactions that disclose the price in the Wall Street Journal or Mergers and Acquisitions. The
corresponding values for U.S. targets are in parentheses.
Bid price ($ millions)b
Sample sizea
Year
Total
1975
5
Mergers and
tender offers
(4)
3
(2)
1
(0)
0
(0)
1977
0
(0)
0
(0)
1978
5
(5)
Partial acquisitions
of independent firms
(less than 50%)
0
(0)
Acquisitions
of units
2
(2)
Total
Average
127.50
(127.50)
(31.88)
1.00
(0.00)
31.88
(0)
cf
1.00
(0.00)
0
V-8
0
(0)
0.00
(0.00)
0.00
(0.00)
(;)
2
(2)
2
(2)
112.05
(43.55)
22.41
(12.71)
(6,
(:,
4
(2)
0
(2)
122.50
(148.00)
24.50
(37.00)
(Z)
5
(3)
3
(1)
1
(2)
260.40
(287.00)
43.40
(95.67)
1981
7
(2)
2
(0)
0
(0)
5
(2)
88.00
(7.00)
17.60
(7.00)
1982
8
(4)
(A)
4
(3)
3
(1)
22.70
(9.20)
7.57
14.60)
8
(6)
2
(2)
2
(0)
(:)
748.70
(730.30)
149.74
(182.58)
1984
8
(9)
1
(1)
3
(3)
4
(5)
690.50
(713.80)
98.64
(101.97)
1985
6
(6)
1
(0)
1
(1)
4
(5)
231.25
(154.25)
57.81
(38.56)
1986
16
(15)
3
(1)
5,
9
(9)
1776.00
(1252.90)
161.45
(125.29)
1987
12
(13)
3
(2)
3
(4)
6
(7)
3955.40
(4852.40)
565.04
(441.13)
1988
27
(26)
5
(7)
4
(5)
14
(14)
6448.70
(5428.40)
308.03
(271.42)
Total
119
(102)
30
(21)
30
(26)
59
(55)
14404.70
(13774.30)
175.96
(183.66)
1976
0
- 1979
2930
1983
---
-
--
-
“Since the sample size for bidders includes cases in which their targets are not listed on the stock
exchanges, and the same is R de for targets and their bidders, bidders and targets in the specific year
are not always matched pans.
bBid prices are a ~&ble for 83 Japanese bidders and for 75 U.S. targets. The sample is excluded
from the computation of total and average bid prices if its bid price is unavailable.
J.-K. Kmg, The intemaiioonalmarket for corporate control
351
firms, and 55 cases are divestitures. Both the number of transactions and the
total bid amounts increase substantially after 1985. In fact, 46% of the number
of transactions and 84% of the dollar value are concentrated in the 1986-1988
period.
In analyzing the announcement effects and characteristics of Japanese bidders, I use U.S. bidders involved in domestic mergers and acquisitions as
a control sample. A U.S. control bidder is matched to each Japanese bidder by
type of acquisition (i.e.,merger, tender offer, partial acquisition, or acquisition of
a unit), method of payment, industry (at least to the first two digits of the SIC
.S. bidder whose market value of equity most closely
matches that of a Japanese bidder is selected. If no firm matches by year, a U.S.
firm matched in the previous or following year is used.
I match firms by industry because firms in certain industries, such as oil, tires,
and tobacco, tend to have higher levels of free cash flow than those in other
industries .* Since leverage ratios tend to differ across industries [Bradley,
Jarrell, and Kim (1984)], industry matching also ensures that any documented
differences in the capital structure of Japanese and U.S. bidders do not merely
reflect industry factors. The U.S. and Japanese bidders are matched by year
because Japanese leverage ratios and proportions of bank loans ?rt the capital
structure have changed substantially over the last ten years [French and
Poterba (199 l)]..
Since the industrial relatedness between Japanese bidders and their U.S.
targets takes many forms, 1,
‘f ic
=lidifhcult to find matched U.S. bidder and target
pairs in the same industries as the Japanese bidder and target pairs. To analyze
target gains and characteristics more precisely, 1[construct a sample of control
targets that are matched to targets of Japanese firms by type of acquisition,
method of pa? sent, industry, year, and market value of equity.
I obtain data on Japanese financial and ownership structure from the PACAP
Research Center data base, lni=anational Annual Report by the Center for
International Financial Analysis and Research, Induefrial Groupings in Japan
bv Dodwell Marketing Consultants, Japan Company Handbook by Keizei
Shinposha, and Analyst’s Guide by Daiwa Institutes of Research Ltd. The
corresponding sources of data for U.S. firms are COMPUSTAT, annual reports,
and proxy statements. The exchange rate (U.S. dollar/Japanese yen) and the
‘The exceptions to type matching are acquisitions of joint ventures and partial acquisitions of
units, which are sometimes not available in domestic mergers and acquisitions. For these types, U.S.
’ -‘:‘;nn) for a few acquisitions made by
firms that acquire entire units are used as control bidders. In aourr.,_
~~G)CP
‘hi,+i~tp
by the first one &gi;it of
Japanese firms in the wholesale industry, I match U.S. and Jal~....-~
YsW_I_C
-A acquisitions by the industry ofJapanese firms shows that
the SIC code. A breakdown of mergers a--the most M&A activity invoives firms classified in manufacturing (74 cases), ‘>ank,insurance, and
securities (21 cases), and whoiesale and retail trade (19 cases). The majority Jf Japanese fm~ use
cash as the form of payment when they acquire U.S. firms.
%ee Jensen (1986) for a detailed discussion.
352
J.-K. Kang, The internationalmarketfor corporate control
consumer price index for the U.S. and Japan are from S&P’s Statisticai Report.
The price-earnings (F-E) ratio for the First Section of the Tokyo Stock
Exchange (TSE) and the TS Price Index (TOPIX) are from the TSE Fact Book
and the TSE Monthly Statistical Report.
2.2. Sample characteristics
Table 2 reports summary statistics for the sample. Panel A shows large
differences in capital and ownership structures between Japanese nonfinancial
bidders and their U.S. counterp: r’s, Leverage (debt/total assets) measured
by book value is roughly 30% higher for the Japanese bidders. When the quasimarket value of the firm (debt plus market value of equity) is used in place
of the book value of total assets to measure the leverage ratio, hC+vevey,
the difference in average leverage is reduced substantially (54.07% v. SO.93YGj.
Panel A also indicates that loans from financial institutions in the Japanese
sample account for about 20% of the market value of the firm. In comparison,
these loans represent only 5.4% of firm value for the U.S. sample. Japanese
banks and insurance companies hold, on average, 48% of the equity in
firms, and all institutions, including banks, insurance companies, investment
trusts, securities companies, other corporations, and foreign institutions, collectively hold about 78% of total equity. It is noteworthy that Japanese bidders
maintain large cash balances and investments in marketable securities as financial slack, almost 12”/0 of the market value of the firm. Consistent with the
finding of French and Poterba (1991), the average F-E ratios for Japanese
bidders are substantially higher than those for U.S. bidders (47.94 v. 15.20
times).g
Panel B of table 2 compares the ownership and capital struct :re of U.S.
targets of Japanese firms with those of targets in the control sample. U.S. targets
of Japanese firms are characterized by higher management ownership than
control-sample targets (15.08% v. 10.41%). The t-statistic for the difference is
1.80. Since Japanese firms tend to undertake friendly takeovers, these results are
not surprising, and may indirectly indicate that Japanese firms acquire U.S.
firms for synergistic rather than disciplinary purposes [Merck, Shleifer, and
Vishny (1988)], Panel B also shows that Japanese bidders hold a relatively larger
fraction of target shares before the bid announcement than U.S. bidders (5.21%
v. 2. TO%), but iius difference is not statistically significant.
‘French and Poterba (1991)show that Japanese P-E ratios increase dramatically in the second
half of the 1980s.Between 1975and 1985,my data indicate that the average P-E ratio for Japanese
bidders was about twice that for U.S. bidders. During 1986-1988,however, the Japanese bidder P-E
ratio doubles from 30.0 to 68.4,while the U.S. bidder ratio increases by only 37%. The increate in she
P-E ratio for Japanese bidders after 1985is dominated by an increase in stock prices, rather than by
a decline in the earnings per share.
J.-K. Kang. The internatiotzal market for corporate control
353
Table 2
Selected summary statistics (means) for a sample of 98 Japanese nonfinancial bidders and 102
U.S. targets in cross-border mergers and acquisitions (1975-1988), and a control sample of 98
U.S. nonfinancial bidders and 102 U.S. targets in domestic mergers and acquisitions. A U.S.
bidder and a U.S. target involved in domestic mergers and acquisitions are matched to each
Japanese bidder and to each U.S. target of a Japanese bidder by type of acquisition, method of
payment, industry, year, and market value of equity. Samples sizes used to compute means are in
parentheses.
Panel A: Nonfinancial bidders
Variables’
Market value of b tder equity ($ millions)
Total assets ($ miC.ons)
Sales ($ millions)
P-E ratios (times)
R&D, sales (%)
Debt/total assets (%)
Debt/market value OSthe firmb (%)
Loans/market value of the firm (%)
Financial slack’/market value of the firm (%)
Management ownershipd (%)
Equity ownership (%)
Bank and insurance cos.
Investment truct & s,:curities cos.
Other corporations
Foreigner
Individual
Japanese firms
U.S. firms
3076 (98)
6032 (98)
12231 (97)
47.94 (93)
2.83 (44)
72.08 (98)
54.07 (98)
20.24 (98)
11.78 (97)
0.63 (91)
2714 (98)
7000 (98)
6769 (98)
15.20 (91)
3.46 (61)
56.26 (98)
50.93 (98)
5.42 (9s)
5.26 (98)
10.2S (92)
48.2 (91)
4.2 (91)
18.9 (91)
7.3 (91)
22.0 (91)
N.A.
N.A.
N.A.
N.A.
N.A.
Panel B: U.S. targets
Variable
Market value of target equity (%millions)
Debt/total assets (%)
Debt/market value of the firm (%)
Bidder ownership (%)
Management ownership’ (%)
Block ownership (more than 5%) (%)
Bank and insurance cos.
Mutual fund and investment cos.
Foundation and corporations
Target of
Japanese firms
Targets of
U.S. firms
1423 (102)
60.57 (102)
56.57 (102)
5.21 (99)
15.08 (98)
1296 (102)
63.94 (102)
56.79 (102)
2.70 (98)
10.41 (98)
2.56 (98)
3.10 (98)
3.42 (98)
3.32 (98)
3.58 (98)
6.19 (98)
“Variables are based on the figure reported at the end of the fiscal year preceding the bid except for
ownership data for U.S. firms, which are based on the figures reported in the proxy statemeE.-s
preceding the bid. Market values of equity, &otaIassets, ana sales for Japanese firms equal the yen
value of each variabk: times the monthly dverage exchange ra!te ($/Y) at the end of the fiscal year
preceding the bid.
bMatrket valu e of Ihe firm = debt + market value of equity.
‘Financial slack = cash + marketable securities.
dFor Japane se bidders, management ownership is the sum of the block ownership (more than 3%)
of bidder shares held by officers and directors. For U S. bidders, it is the sum of tke fraction of bidder
shares held by all officers and directors.
‘The fraction of target shares kcld Qy officers and directors.
354
J.-K Kang, The international
market _for corporate
c,:ltrol
2.3. Methodolsgy
I compute abnormal returns using standard event-study methodology.
Market-model parameters are estimated from 220 ;o 20 trading days before the
first bid announcement. For Japanese firms, i use the PACAP equally-weighted
index as the benchmark, and for U.S. firms. the CRSP equally-weighted index.
Since Japanese initial bid annol~ncennent dates are identified from the WSJ
(U.S. day 0) and trading hours in okyo and New York never overlap, there
is a time-zone difference between the two markets in observing stock-price
reactions. Since stock-price reactions in Tokyo at U.S. day 0 are observed one
busisess day later, Japanese day 0 is defined as one business day after U.S. day 0.
tier single and unrevised offers, the cumulative abnormal return (CAR) from
day tl through t2 is estimated from day tl before Japanese day 0 to day t2
after Japanese day 0. For multiple-bid and revised Xers, the CAR is estimated
from day tl before Japanese day 0 to day t2 after the final revision by the
successful bidder. 1 use t-statistics to test the hypothesis that the zverage
analyze the total synergistic gains created by takeovers, I construct
a ; alue-weighted portfolio of the target and the bidder. Following Bradley,
Desai, and Kim (1988), I measure the total synergistic gain as a value-weighted
average of the target CAR and the bidder CAR, with the weights being the
hmarketvalue of target and bidder equity at the end of the fiscal year preceding
the bid. For the market value of target equity, target shares held by the bidder
are excluded from the number of shares outstan iing to avoid double counting.
3. Empirical results
3.1. Return analyses
Panel A of table 3 repot :s the CARS of 119 Japanese bi&L :s and 119 U.S.
bidders. For Japanese bidders, the CAR( - 1,O) and the CAR( - 1, 1) are
0.59% and 0.51 %, which are statisticalij: significant at the 0.05 and 0.10 level,
respectively. In contrast, U.S. bidders realize insignificant CARS of - 0.29%
and - 0.10% during the same intervals. In particular, the mean difference of
CARS over ( - 1,0) between Japanese and U.S. bidders is significat tly different
from zero at the 0.05 level. These findings support the view that cross-border
acquisitions enable multinational firms to exploit imperfections ;:! product,
factor. and capital markets, and thus create moie gains for their shareholders
[Kindreberger (1969), Caves (1971), Hymer (1976), and Froot and Stein (1991)].
Panel A also presents the CARS of Japanese bidders by subperiod, 1975-1985
and 19864988. I bifurcate the sample in this manner because, as discussed in
section 2.1, about half of the sample of Japanese acquisitions in the U.S. is
concentrated in the final three years of the sample period, which represents 84%
J.-K. Kang. The internatiotzal market for corporate control
5.35
Table 3
Cumw i r~ abnormal returns (CARS) for a sample of 115 Jap rnec ardders and 102 U.S. targets L
cross-border mergers and acquisitions (19751988), a subs? m;i~ &’‘38Japanese nonfinancial bidderu
in cross-border mergers and acquisirions stratified accordrhzgto debt and equity positions held by
a single bank, and a control sample of 119 U.S. bidders and 102 U.S. targets in domestic mergers and
acquisitions. A U.S. bidder and a U.S. target involvec4 in domestic mergers and acquisitions are
matched to each Japanese bidder and to each U.S. target of a Japanese bidder by type of acquisition”.
method of payment, industry, year, and market value of equity. For Japanese firms, day 0 is one
business day after the initial announcement date of the bid identified from the Wall Street Journal. For
U.S. firms, it is the initial announcement date. CAR(t,, t2) is the cumulative abnormal returns from
aay t I before day 0 to day t2 after the final revision by the successful bidder (t-statistics in parentheses).
Panel A: C.4Rs of Japartese and U.S. bidders
Full period: 1975-1988
Days in relation to
initial announcement
date
Japanese firms (A! = 119)
-
U.S. firms (Iv = 119)
Difference
CAR (%)
CAR (%)
CAR (%)
0.59b
(2.41)
- 0.29
( - 1.05)
0.51’
(1.85)
( - 0.33)
0.88b
(2.40)
0.61
(1.47)
0.51
(1.55)
( - 0.54)
0.17
(0.42)
( - 0.30)
-ltoO
- 1 to 1
-2tol
-5tol
- 5 to 5
- 0.10
- 0.15
- 0.05
- 0.06
( - 0.09)
( - 0.09)
0.66
(0.50)
0.03
(0.03)
- 20 to 20
0.70
(1.46)
0.32
(0.50:
- 0.19
0.0:
(0.01‘3
0.63
(0.36)
Subperiods: 1975-l 985 and 1986-l 988
Days in relation to
initial announcement
date
. :%I0
- 1 to 1
__2 to 1
- 5 to 1
- 5 to 5
- 20 to zc
U.S. firms
Japanese firms
-
1975-85
(N = 64)
1986-88
(N = 55) Difference
--
1975-85
1986-88
(IV = 64)
(N = 55)
DiEere:xe
CAR (%)
CAR 4%)
CAR(%)
CAR (%)
CAR (%)
- 0.77’
0.90
(1.63)
0.62b
(2.23)
0.51
(1.59)
0.56
(1.31)
CAR (%)
0.06
(0.13)
0.69b
(2.27)
0.52
(1.09)
0.30
(0.48)
0.22
(0.56)
0.12
(0.15)
0.10
(0.12)
( - 0.18)
0.01
(0.01)
( - 0.10)
0.22
(0.19)
1.18
(0.45)
- 0.96
( - 0.34)
- 0.10
- 0.01
1 - 0.02)
0.39
(0.57)
- 0.11
0.13
(0.38)
0.32
(0.7 ,)
0.40
(0.87)
0.59
(0.83)
0.89
( H.07)
.62
C&37)
( - 1.79)
- 0.59
( - 1.22)
- 0.88
( - 1.65)
- 1.02
0.91
(1.45)
1.28
(1.63)
( -* 1.34)
1.61
(1.60)
2.05’
(170)
- 0.65
- 0.40)
1.27
(0.54)
( - 1.45)
- 1.16
J.-K. Kang, The international market for corporate control
356
Table 3 (continued)
strat@d
Pane! B: CA Rs of Japanese nonfinancial bidders,
according to the debt and equity positions held by a sing/e bank
Full period: 1975-1988
Days in relation to
initial announcement
date
Firms where the same
bank is the largest
shareholder and the
largest loan provider
(N = 33)
Other firms (N = 65)
Difference
CAR (%)
CAR (%)
CAR (%)
1.46”
(2.91)
0.24
(0.70)
1.22b
(2.03)
0.07
(0.19)
1.48b
(2.18)
(2.30)
0.02
(0.05)
1.48’
(1.87)
0.52
(0.62)
c.09
\‘u16)
0.43
(0.42)
0.26
(0.38)
0.21
(0.16)
1.08
(0.59)
- 0.18
( - 0.05)
- I to 0
-
1.5Sb
1 to I
(2.69)
-2to
4
-St0
l.SOb
I
0.47
(0.44)
-St05
- 20
to 20
=
0.90
(0.31)
Panel C: CARS of U.S. targets
Full period: 1975-I 988
Days in relation !o
iniM announcement
date
Targ ts of Japanese
fir ns (N = 102)
:“AR (%)
-
Targets of U.S. firms
(N = 102)
CAR (%)
Difference
--CAP (%)
--It00
9.07a
(-5.69)
6.84”
(4.41)
2.23
(OW
- 1 to 1
9.42a
(4.83)
7.01”
(4.36)
2.41
((‘.96)
-2to
9.63a
(4.94)
8.07”
(4.77)
1.56
(0.61)
-5tol
10.85”
(5.32)
9.&84s 4,
1.37
(0.49)
-5to5
9.92”
(4.63)
9&j”
(G.99)
0.03
(0.01)
12.418
(4.09)
13.71”
(4.87)
- ‘i.30
( - 0.32)
--
1
- 20 to 20
--
J.-K. Kang, The inrem donalntoicketfor corporate control
357
Table 3 (continued)
Subperiods:1975-1985 and J984-1988
Targets of Japanese firms
Days in relation to
initial announcement
date
Targets of U.S. firms
1975-85
(N = 48)
--
198688
1975-85
1986-88
(IV = 54) Difference (A! = 48) (N = 54) Difference
-Cc R (%) CAR (%) CAR (%) CAR (%) CAR (o/o) CAR (%)
-it00
8.42’
(3.45)
9.65”
(3.27)
- 1.23
( - 0.32)
4.99’
(2.85)
8.96”
(3.62)
- 3.97
( - 1.49)
- 1 to 1
8.57’
(3.46)
10.18”
(3.42)
- 1.61
( - 0.42)
5.39”
(2.80)
8.45’
(3.36)
4 - 0.97)
9.07”
(3.61)
10.13’
(3.43)
( - 0.27)
5.85”
(3.32)
10.03’
(3.62)
( -. 1.27)
9.94’
(3.71)
i 1.66’
- 1.72
c3.83)
( - 0.42)
9.26”
(3.21)
10*5?”
(3.33)
- 1.25
( - 0.29)
6.09’
(3.06)
6.76”
(3.12)
12.4Y
(3.99)
12.68’
(3.98)
- 5.92
( - 1.54)
14.55”
(3.20)
10.50b
(2.57)
4.05
(0.66)
10.17”
(‘3.25)
16.86”
(3.73)
( 1 1.22)
-2tol
-St01
-St05
- 20 to 20
- 1.06
a- 3.06
- 4.18
- 6.40’
( - 1.73)
0.69
“Significant at the 0.01 level.
bSignificant at the 0.05 level.
‘Significant at the 0.10 level.
of the dollar value of transactions. Further, Japanese P-E ratios began increasing substantially around 1986. Since the TSE became substantially overvalued
in the 1986-1988 period [French and Poterba (1991)], it may have systematically overestimated the true value of acquisitions by Japanese bidders. The
evidence in panel A, however, indicates that this ‘bubble’ effect is not an
underlying cause of higher CARS for Japanese bidders in the 1975-1988 period.
The CARS for the two subperiods are not significantly diRerent.
Panel B of table 3 reports the CARS af two subsamples of Japanese nonfinancial bidders, stratified according to the debt and equity positions held by a single
bank. Although Japanese multinational firms in which a single bank holds the
largest portion of equ;ty and loans (i.e., serves as a proxy for the presence of
a main bank) experience a positive CAR of i .46% (t = 2.91) over ( - 1, 0), the
other firms realize an insigr;&ant positive CAR of 0.24%. The iflerence in the
CAR between two ~.;-oupsis statistically significant at the 0.05 level. Further,
the number of negative CARS is smaller for the main-bank-related firms than for
the other firms over all selected intervals. lo These rest&s suggest that the
10Specifically, the number of negative CARS for the main-bank-re!ated firms over the event
windows ( - 1, O),( - 1, l), and ( - 2, 1) is 39%, 30%, and 36%, respectively. For t1 s other firms. it
is 43%, 49Oh, and 46%, respectively.
358
J.-K. Kang, The internationalmarketfor corporate contrd
significance of announcement effects for Japanese firms is partly attributable to
firms that have a close relationship with a main bank. They also indicate that
when the same bank is both a firm’s largest shareholder and the largest lonn
provider, it has a greater %XXX% incentive to prevent negative-NP o/ bids.
Panel C oi table 3 reveals that shareholders of U.S. targets of Japanese fifes do
not earn gains statistically distinguishable from those of control sample targets.
3.2. Explanatory variables
In this section, I discuss several explanatory variables that may be related to
Japanese bidder returns. Table 4 summarizes the variables used in the regression
analysis, including definitions and acronyms.
a Monitorirzg variables. I use three variables as instruments for monitoring by
financial institutions (banks and insurance companies): LOAN is the ratio of
loans from financial institutions to the market value of the firm; BANKOWN is
equity ownership by financial institutions; and MAINBANK is a dummy
variable that takes a value of one if the same bank is both the largest shareholder
and the largest loan provider. I expect that equity htihdings and/or loans by
financial institutions create incentives for these institutions to monitor. Thus,
abnormal returns earned by Japanese bidders are expected to be positively
related to LO_4N, BANaKOWN, and MAINBANK.
To examine whether other debtholders or equityholders can influence managements actions, I use three other variables: OTHERDEBTis the ratio of other
debt (total debt minus loars) to the market value of the firm; CORPOWN is
equity ownership by other corporations; and IN VO WN is equity ownership by
investment trusts and securities compsi ies. Finally, to test the free cash flow
hypothesis advanced by Jensen (1986), 1 .nclude LEVERAGE, which is the ratio
of total debt to the market value oi -he firm. Jensen claims that managers
endowed with free cash flow tend to invest it in negative-NPV projects rather
th; 7 pay it out to shareholders. According to this hypothesis, debt creation
heir93 limit the waste of free cash flow and controls managerial discretion.
coqn. _tent with this argument, Maloney and McCormick (1989) find that bidder
retur * are positively related to the bidder’s leverage.
Exchange-rate variable. I measure the strength of the Japanese yen in
relation to the U.S. dollar as the deviation of the real exchange rate (S/S) for the
year of the bid announcement from the average real exchange rate for the
1975-1988 sample period. EZCHRATE is this difference divided by the average
real exchange rate. The real exchange rate is employed to adjust the nominal
exchange rate for differences in inflation rates be
the U.S. ptld Japan.
Therefore, the higher the value of EXCHRATE, the
r the real value of the
dollar.
Table
4
Definitions of the variables used in the
Variables
-._I
LOAN
MAiNBANK
Definitions
-IILoar;sJmarket value of the firm”
Dummy variable that takes the value of one if the same bank is both the largest
shareholder and the largest loan provider
Aguity owmmhip held by Japanese institutionsb:
Sum of the fraction of bidder shares held by
BANK0
CORPOWN
Sum of the fraction of biddershares held by ot
INVOWN
Sum of the fraction of bidder shares held by
companies
(Debt - loans)/market value of the firm
Debt/market value of the firm
Change in the real exchange rate”
Dummy variabie that takes the value of one if the bid is made during the
19814986 period
Dummy variable that takes the value of one if the bid is made after 1986
TA.Y86
BIDMGM TO WN Sum of the block ownership (more than 3%) of bidder shares by Japanese
officers and director@
(Cash + marketable securities)/market value of the firm
FSLACK
TSE P-E ratio for the year of the bit taken/TSE P-E ratio for 1992’
TSEPE
Dummy variable that takes the value of one if the bidder holds target share;
BIDOWNDUM
fore the bid announcement
Log (market value of bidder qaity)
LOGBIDMV
Market value of target equity/market value of bidder equity
M VRATIO
Dummy variable that takes the value of one if the type is a merger or a tender offer
MERGER
Dummy variable that takes the value of one if the type is acquisition cl a t.tnit
UNIT
Dummy variable that takes the value of one if the firm is in the manufacturing
MANUFACT
industry
Dummy variable that takes the value of one if there is more than one bidder or
MULl7PLE
a bid reversion
OTHERDEBT
LEVERAGE
EXCHRATE
TAX81
Distribution qf target ownership:
MGMTOWN
Sum of the fraction of target shares held by officers and directors
BID0 WN
Sum of the fraction of target shares held by the bidder before the bid announcement
BANKBLKOWN
Sum of the block ownership (more than 5%) of target shares by banks and
insui ante companies
CORPBLKO WV Sum of the block ownership (more than 5%) of target shares by foundations and
other (i:orporations
IN VBLKO WN
Sum of the block ownership (more than 5%) of target shares by mutual funds
and investment companies
TARGET OF
JAPANESE
BIDDER
LOGTGTM V
Dummy variable that takes the value of one if a U.S. firm is the target of the
Japanese firm
Log (market value of target equity)
“Market value of the firm = debt + market value of equity.
bFor US. bidders, equity ownership is the sum of the block ownership (more than 5%) of tidder
shares by U.S. institutions.
‘Change in the real exchange rate is the deviation of the real exchange rate (%/U) for the year of the
bid announcement from the average real exchange rate for the sample period 19754988, divided by
the average real exchange rate. If the bid is made after June, the exchange rate for the bid year is
used. If the bid is made before June, the exchange rate from the year before the bid is used.
dFor U.S. bidders, BIDMGM TO WN is the sum of the fractiohi of bidder shares held by all oticers
and directors.
‘The average P-E ratio from January 1992 to October 1992.
360
J.-K. Kang, The internationalmarketfor corporate cvtrol
Tax variables. Two indicator variables are included to measure the eflect of
the U.S. tax changes: TAX81 is a dummy variable that takes the value of one if
the bid is made during the 1981-1986 period: lrAX86 is a dummy variable that
takes the value of one if the bid is made au:. : ‘Y. Tf tax benefits are reflected in
the takeover gains, Japanese bidder returns are likely to be lower in the
1981-1986 period and higher in the 1987-1988 perrod.
managerial equity ownerr) Bidder management Iwp-hip variable.
ship can lead to a better &nment of shareholder tend management interests
[Jensen and Meckling (19?6)]. Therefore, bidder returns are likely to be Fositively related to management ownership. Lewellen, Loderer, and Rosenfeld
(1385) show that bidder returns increase with the frac ion of bidder equity held
by management. To examine the effect of the managerial ownership on the
market value of the Japanese bidder’s equity, I define BZDMGMTO WN as the
block ownership (more thzn 3%) cf ’ -’ 4der shares i.jy officers and directors.’ ’
For U.S. firms, BZDMGM TO WN is t1.r;sum of the fraction of bidder shares held
by all officers and directors.
a Other control variables and target ownership variables. As discussed in section 2.2, Japanese bidders have large cash balances, twice those held by US.
counterparts in the control sample. It is widely known that by the mid-1960s
some Japanese firms deposited cash in the same banks from which they borrowed. To control for this source of financial slack, I include FSLACZIC, which is
the ratio of financial slack [cssh plus marketable securities) to the market value
of the firm. In addition, it could be that the results are affected by the extremely
high valuations of Japanese firms in the second half of the 1980s. For instance,
by acquiring U.S. firms, Japanese firms could transfer the extremely high
valuations of their cash flows to the target cash flows and thereby increase their
value. To control for this eflect, I use as an explanatory variable TSEPE, which
is the ratio of the TSE P-E mu’,5l)le for t:le year of the bid to the TSE P-E
multiple for 1992. Since Japanese P-E ratios declined
f+im 7a
;QI17v9
l nQtb Cc
_ ___---- ..VI\..
, V. 6J LU
BC. S-L@
by the end of October 1992, TSEPE should reflect the extremely high valuations
of Japanese firms in the later part of the sample period.12
The literature shows that several other variables affect bidder returns. To
control for prior bidder ownership af target equity [Mikkelson and Rubac:k
(1985)], the size of the t,idder or the relative size of the target and the bid%ier
[Asquith, Bruner, and Mullins (1987)], the type of acquisition the industry
I1Because of a ;ack or’available Japanese sources, Yam unable to Uti ii;_;r;:,x,&Gment ownership
of all officers and ~Jirectors. To come as close to this SWg-=Fsiblewith the av,ilable information, I use
3% as the cutoff l.oi+
121 thank the rer”ertb for raising this issue and suggesting this approach.
J.-K. Kang, The intmational market for corpwclte control
361
effect [Maloney and McCormick { i989)], and the numbnr of bidders,’ 3 I include
the following seven variables: BIDOIVNDUIM is a duinmy variable that takes
thr: value of one if the bidder holds target shares before t &bid announcement;
LOGBILUJ4V is the logarithm of market value of bidder equity; A4VRATZO is
the relative market vjIalue of target equity to bidder eo-:‘:y; MERGER is
a dummy variable that takes the value of one if the type is a merger or a tender
offer; UNIT is a dum y variable that takes the value of one if the type is
acquisition of a unit; MANUFACT is a c!b, ny variable that takes the value of
one if the bidder is a manufacturing tirm; at, MULTIPLE is a dummy variable
that takes the value of one if there is more than one bidder or a bid revision.
Finally, Stulz, Walkling, and Scng (1990) show that target returns depend on
the distribution of target ownership among management, large shareholders,
and the bidder. Since managers want to avoid a hostile takeover, they may use
their voting rights to decrease the probability of takeover success and hence
increase its cost for the bidder. 0n the other hand, since the bidders as large
shareholders are supportive of the bid, target returns may fall with the bidder’s
holdings. To examine whether target returns are related to owner&hip structkrre,
I employ the following variables: MGMTO WAf is the fraction of target sha.res
held by officers and directors; BID0 WN is the bidder ownership of target equity
WN, CORPBLKO WN, and INVBLKO W-N
before the bid; and BAN
s of more than 5% of the firm’s outstanding
are blockholdings (equity
voting shares) by backs and insurance companies, foundations and other
corporations, and mutual funds and investment companies, respectively.
3.3. The relation between bidder returns and explanatory variabied
For the cross-sectional analysis, I use the CAR( - 2, 1) as the deper lent
variable. I use the ( - ?I, 1) interval as the announcemen: period to t&e into
account that the day 0 identified from the WSJ may not C,. the same reporting
day as in Japan. l4 To correct for possible
heteroskec”asticity in the data,
L
I estimate regressions using weighted least squares. All o*&ervations are divi&zd
by the standard error of the CAR( - 2,1)? Table 5 repctts regression estimates
for Japanese bidders. The first two regressions use c)P’~Srm-specific characteristics as explanatory variables. They show that the bid ier gains increase with the
leverage ratio (LEVERAGE) and the loan ratio (~&4h ). These findings are
consistent with both Jensen’s (1986) free cash flow hypothesis and Fama’s (1990)
13Bradley, Desai, and Kim (1988) sugg&+ that acquirers irn mu?tiple-bid&r contests may ‘be
subject to a winner’s curse and find that bidder rebgJrra;
xc; lower in such contests.
14Since the results shown hold with a narrower event wnndaw, CAR( - 1, I), E report only
regression results using CAR( - 2,l).
~smreroot of the variance of the predic!+n
lSThis standard error is computed as the q
_qtiO
I errors over
The( - 2, k) interval.
.
J.-K. Kang, The international mcrket for corpora?e control
362
Table 5
Regression estimates ofcumulati ge abnormal returns (CARS) of Japanese bidders that acquired U.S.
firms on explanatory variables (1975-1988). Weighted least squares is used for the estimation. Day
0 is one business day after the initial announcement date of the bid identified from the Wall Street
Journal. The dependent variable is the CAR from two days before day 0 to one day after the final
revision by the successful bidder (t-statistics in parentheses).
- -Regressions
Full period:
1975-1988
Variables”
Subperiod:
1975-1985
Subperiod:
1986-1988
(4)
(5)
- 0.148’
( - 2.17)
- 0.090
[ -_ 0.85)
( 1.94)
0.069’
(2.42)
0.03 1
(0.68)
0.177b
(4.17)
o.c57
(1.49)
0.098’
(2.48)
0.035
(0.56)
0.177”
(3.12)
0.007
(0.76)
0.008
(0.88)
0.006
(0.55)
- 0.024
- 0.54)
( - 0.15)
( - 0.57)
- 0.041
- 0.66)
0.027
(0.30)
- 0.090
( - 0.94)
1N L’OW’N
0.042
(0.33)
0.050
(0.39)
0.016
(0.09)
0.140
(0.75)
EXCNRA TE
0.127b
(4.60)
0. I 28b
(4X0)
_
CONSTArt 7
_-- 0.124’
( -- 2.06)
LOAN
(3)
-~------.-----
ia
- 0.124’
( - 2.04)
- 0.138’
i- 2.10)
0.059d
OTHERDEBT
LEVERAGE
o.0595
(2.68)
MAlNBANK
0.010
(0.95)
B.dNKO WN
0.010
(0.92)
- 0.003
- MO3
- 0.025
( - - 0.06)
( - 0.07)
( - 0.57)
-- 0.034
TAX81
- 0.001
( - 0.07)
TAX86
.- 0.068’
1 - 2.49)
- 0.064
0.064
( - 0.67)
TSEPE
0.195
0.007“
( 1.67)
MERGER
UNIT
-
-~
______ _
0.007
( 1.66)
- 0.008
- OX3
( - 0.31)
t-
- 0.118
1.61)
0.1 72b
(5.25)
- O.rJo2
- 0.16)
- 0.070’
- 2 53)
0.944
(1.05)
0.196
(1.40)
- 0.003
( - 0.38)
- 0.038
1 - 0.43)
0.113
(0.77)
- 0.157
( - 1.33)
04
(0.20)
0.006
(0.27)
0.008
(0.25)
- 0.222b
( - 3.70)
-0001
( - U.91)
t - 0.65)
MO
- cr.001
( - 0.86)
LOGB”nMV
- 0.012
( - 0.82)
0.198
(1.63,)
.
B
: I
I - cL:w
-
0.110
(0.86)
0.079b
(3.36)
CO RF0 Ck’N
FSLACK
(6)
- 0.001
- 0.017
( - 0.52)
0.008d
(1.91)
O.OOYd
( 1.96)
0.003
(0.52)
t>00’2d
cm:o;,
0.015
(1.33)
0.017
(1.41)
0.001
(0.05)
0.041’
( 1.94)
- 0.002
- 0.002
( - 0.20)
0.001
(0.07)
0.006
(0.42)
1 - 0.24)
J.-K. Kang, The international nturket jbr crwpomte l*ontro!
Table 5 (continued)
- ~--___-___
---..
__~._______I__
Variables”
-
MANUFACT
-___-
----_
0)
0.02 5’
(2.26)
~.---
-
___~_________
Regressions
________
--
_-_
----_
Full period:
1975-1988
(2)
0.025’
(2.24)
MULTIPLE
F-ratio
1.90d
2.25’
0.07
Adjusted R2
0.08
91
Sample size
91
-__--_____--__
aSee table 4 for definitions‘ of variables.
bSignificant at the 0.01 level.
‘Significant at the 0.05 leve).
dSignificant at the 0.10 level.
-
363
(3)
(4)
0.025’
(2.42)
- 0.069b
( - 3.84)
0.024.
(2.35)
- 0.069b
( - 3.84)
3.68b
3.45b
0.34
0.33
91
91
- -- ---~----_-_--.
_
Subperiod:
1975-1985
____(5)
Subperiod:
1986198g
0.011
(0.69)
0.037’
(2.67)
- 0.078b
( - 3.85)
--..---__
;6!
0.31
5.20b
3.00
0.60
45
46
__ __ _~__ ____.
monitoring hypothesis, and suggest that Jaiyanese bidders who maintain high
leverage or have substantial borrowings from financial institutions are more likely
to be subject to scrutiny than their peers and therefore less likely to pursue
value-decreasing acquisitions. The other monitoring variables, however, including
the proxy for the main-bank relationship (MMNBANbC ), are qot statistically
significant. Thus it appears that efhcient monitoring is related to the Japanese
bidder’s debt or bank loans rather than the main-bank rekltionship per se. The
positive and significant coefficient on the industry &_znly variable (MAAWF..QCT)also indicates that the benefits of acquisitions in the U S. are larger for the
Japanese manufacturing sector than for the nonmanufacturin~ sector.
In regressions (3) and (4), I dd the exchange-rate and t,tx variables, and
contra! for other variables. l6 Both regressions show that the earect of exchange:a:.z movements (EXCHRA TE) on Japanese bidder gains is st; tistically significant and economically large. A13else being equal, a 10% increase in the value of
the yen leads to an increase of nearly 1.3% in Japanese bidder returns. T:IX
variables, however, do not provide a satisfactory ex lanation for t&c varlztion In
Japanese bidder gains. In particular, TAX&3is significantly negatively. related to
bidder gains, which is contrary to the prediction.’ e The nega+it:c!coeficients un
“Controlling for the relative market valwe of target equity tcl bidder ecpity rather than the
logarithm of market value of bidder ec;.lity (J,OGBII?&I~) leaves the qualitative results in regressiori i3) and (4) unchanged.
“Since the dollar’s fall in i986 coincided closely with the enactment af the Tax Reform Act of
1986, including tax and exchange-rat th
_ ~:riables in the same regression: may cause collinearity
problems. To examine this possibility, I \ ,rform the regressions without EXCHRATE. In those
are no longer significant. EXCHRATE, however.
specifications, the negative coeffici:, rlts on ; ,(.&‘,EA
retains its significance even when P drop the , .A variables. -
364
J.-K. Kang, The international market Jar corporate control
the bidder competition dummy variable (MULTIPLE) indicate that Japanese
bidders lose in multiple-bidder contests or when there are bid revisions=
Finally, in regressions (5) and (6), I examine whether the variables used have
any different effects on the CARS realized by Japanese bidder? in the W75--~W5
and 19864988 periods. The results show that the explanatorg power of several
important variables for the overall sample period is drive;- bi the second
subperiod, 19864988. Whereas the results for most variables in the second
similar to those for the overall period, none of the coefficients are
significant in the first subperiod. Since the value of the transactions is much
higher in the second subperiod than in the first, it is not surprising that the effect
of the monitoring variables (LOAN and OTHERDEBT) can be measured more
precisely in the second subperiod; presumably, creditors have stronger incentives to monitor bidders when acquisitions are larger.” The negative coefficient
on TSEPE in regression (6) suggests that high P-E ratios during the second
subperiod lead tc, lawer Japanese bidder retllrns, which is a clltprising result.
However, a correlation test indicates that TSEPE and EXCHRATE have
a correlation coefficient of 0.82, which is significant at the 0.01 level. Because
including such highly correlated variables can lead to biased and inconsistent
estimates. I reestimate regressions (3) through (6) including only EXCHRATE
or TSEPE. While I do not report the results in table 5, they are unchanged,
except that the coefficient on TSEPE in the second subperiod is no longer
significant when EXCHRA TE is omitted from the regression.1g This suggests
that the statistical significance of LEVERAGE and L&IN in the second subperiod is not due to the overvaluation of the TSE.
I estimate the previous regressions using US. bidder CARS over ( - 2, 1) and
report the results in table 6. Since there are restrictions on equity holdings by
U.S. banks, I focus mainly on loans to examine the relationship between bidder
gains and the monitoring variables. Al’& =$ table 3 shows that the other
monitoring variables cannot explain the cross-s;& inal variation of U.S. bidder
returns, the coeficient on block ownership by other corporations (CORPO FWV)
is statistically significant in regressions (3) through (5), suggesting that institutions play different monitoring reles in the U.S. and Japan, possibly because of
differences in institutional arrangements.
‘*For regressions 15) and (6), a test of changes in the slope coefficients across the different
subperiods indicates that the null hypothesis of equal coeffc’eq nts across subperiods is rejected at the
0.05 level (F = 1.95). As a further test, I reestimate regressions (3) and (4) using an additional
interaction gem, EXCHRATE x YWDUM, where YRDUM takes the value of one if the bids are
made in the second subperiod and zero if made in the first subperiod. The results show that the
coe%cicnts on this variable are positive and significant.
’ 9As a robustness check, I use an alternative measure for TSEPE, the deviation of the TSE P-E
ratio for the year of the bid from the average TSE P-E ratio for the period 1975-1992, divided by the
average TSE P-E ratio. The results are the same. The results are also si_milar when the TSE P--E
ratio is replaced with the TOPZX.
Table 6
Regression estimates of cumulative abnormal returns (CARS) of U.S. bidders involved in domestic
mergers and acquisitions on explanatory variables (19751988). Weighted least squares is used for
the estimation. A U.S. bidder is matched to each Japanese bidder by type of acquisition, method of
payment, industry, year, and market value of equity. Day 0 is the initial announcement date of the
bid identified from the Wall Street Journal. The dependent variable is the CAR from two days before
day 0 to one ddy after the final revision by the successful bidder (t-statistics in parentheses).
Regressions
Full period:
19751988
Variables”
CONSTANT
(1)
- 0.001
;t - 0.03)
(2)
Subperiod:
1975-1985
!3)
(4)
- 0.004
- 0.013
- 0.013
( - 0.19)
( - 0.39)
( - 0.39)
Subperiod:
19861988
(5)
-
(6)
--
0.007
(0.18)
( - 0.35)
- 0.021
LOAN
0.053
(0.73)
0.02 1
(0.29)
0.058
(0.54)
0.027
(0.27)
OTHERDEBT
0.009
(0.43)
0.016
(0.46)
- 0.007
( - 0.24)
( - 0.12)
- 0.027
( - 0.30)
0.147
(1 37)
( - 1.32)
O.llld
(1.99)
0.i59
(1.19)
- 0.236d
0.148d
(l-66)
LEVERAGE
0.014
(0.70)
BAhrKOWN
o.oi3
(0.15)
0.016
(0.83)
0.014
(0.16)
CORPO WN
- 0.027
( - 0.30)
0.123’
(2.09)
0.076
(1.23)
‘N VO WN
- 0.003
TAX81
! - 0.24)
- c.014
TAX84
( - 0.94)
BIDMGMTO WN
FSLACK
LOGBIDM V
0.076
(1.23)
f - 1.85)
- 0.003
( - 0.22)
- 0.014
( - 0.93)
ci.329
(0.84)
0.030
(0.84)
- 0.023
0.002
i0.w
0.037
(0.63)
- 0.120
- 0.063
( - 0.98)
( - 0.31)
- 0.022
( - 0.30)
0.028
(0.33)
( - 0.97)
0.001
(0.56)
0.002
(0.70)
0.003
(0.95)
0.003
(0.92)
0.001
(0.2 1)
0.001
(0.97)
- 0.067
- 0.033’
( - 2.39)
- 0.013
IJNIT
( - 1.33)
- 0.022’
( - 2.22)
- 0.021’
i - 2.16)
- 0.009
( - 0.80)
1.56
0.03
92
1.34
0.02
92
“See tyc!e 4 for definitions of variables.
bSig-,tficant at the 0.01 level.
‘Si&ficant at the 0.05 level.
dS1.gnifican t a t the 0.10 level.
- 0.033’
( - 2.35)
- 0.029”
( - 2.12)
- 0.041
(-1.64)
- 0.013
( - 1.31)
- 0.006
( - 0.49)
( - 1.12)
- 0.009
( - 0.79)
- 0.004
( - 0.32)
( - 0.89)
( - 1.63)
- 0.044
( - 1.61)
2.30’
0.16
92
2.11’
0.15
92
- 0.044
M JLTIPLE
F-ratio
Adjusted R 2
Sample size
0.122’
(2d4)
- 0.187
( - 1.05)
MERGER
MANUFACT
- 0.005
- 0.021
- 0.017
--_0.023
( - 0.62)
1.79d
0.16
46
1.66
0.15
46
-
366
J.-K. Kang, The international market for corporate control
To gain further msight into Japanese and domestic mergers and acquisitions,
I examine the relat,ionship between the CAR realized by the portfolio of bidders
and tar;:ts and the explanatory ~=iables. Regression estimates are reported in
table 7. For the portfolio of Japanese bidders and their U.S. targets, the findings
are similar to those reported in table 5: the coefficients on LOAN, LEVERAGE,
and EXCHR A TE are positive and significant.
Table 7
Regression estimates of cumu :*“pp abnormal returns (CARS) realized by the portfolio of Japanese
bidders and their U.S. targsts (19751988) and by the portfolio of U.S. bidders and their U.S. targets
CXIexplanatory variables. Weighted least squares is used for the estimation. A U.S. bidder is matched
td each Japanese bidder by type of acquisition, method of payment, industry, year, and market value
of equity. The CM realized by the portfolio is measured by a value-weighted average of the target
CAR and the bidder CAR, where the weights are the marl et va?ue of target and bidder equity at the
end of the fiscal year preceding the bid. For Japanese firms, day 0 is one ‘business day after the initial
announcement d&e of the bid identified from the Wall Street Journal. For U.S. firms, it is the initial
announcement date. The defxndent variable is the portfolio CAR from two days before day 0 to o:r3e
day after the final, revision by the successful bidder (t-statistics in parentheses).
Regressions
---
Portfolio of U.S. bidders
and their U.S. tzrgcts
Portfolio of Japanese bidders
and their U.S. targets
Variables”
-CONSTANT
-
-
(!,
(2)
(1)
(2)
0.035
(ass)
0.036
(0.58)
0.001
(0.06)
0.003
(0.17)
LOAN
OTliERDEBT
LEVERAGE
Mi44NBANK
BA.YKO WN
COR PO WN
INVOWN
EXCHRATE
TAX81
TAX86
- 0.010
0,073d
(1.79)
( - 0.17)
0.070
(1.26)
0.072’
(2.28)
f 1.019
(1.46)
- 0.109
(-160)
0.034
(1.65)
0.028
(1.53)
.
0.019
(1.42)
- 0.109
- 0.161’
( - 159)
( - 2.12)
- 0.156’
( - &.
3 12)
- 0.102
- 0.102
- 0.040
- 0.034
( - 1.19)
( - 1.17)
( - 0.67)
( - 0.55)
- 0.042
( - 6.22)
- 0.043
( - 0.22) ,
0.112
(1.21)
0.107
(1.05)
0.014
(0.92)
0.012
(0.74)
0.088’
(2.11)
0.002
(0.09)
- 0.026
( - 0.61)
0.08+ “9:
(2.09)
0.0(\3
(0.10)
- 0.026
- 0.004
- 0.005
( - 0.59)
( - 0.29)
( - 0.35)
“--
J.-K. Kang, Tire international market -for corptirate control
_.
Table 7 (continued)
---
-----
-
Regressions
--
Portfolio of Japanese bid&xc
and their U.S. targets
Vdrkblesa
BIDjdGTO WN
FSLACK
TSEPE
M VRA 170
MERGER
UNIT
MANUFACT
MULTIfLit
367
(1)
(2)
- 0.015
( - 0.10)
( - 0.10)
- 0.015
- 0.145
- 0.144
( - 1.35)
( - 1.30)
0.001
(0.03)
0.001
(0.03)
Portfolio of U.S. bidders
and their U.S. tar:.&
(1)
(2)
0.002
(0.10)
0.002
(0.09)
C.100
(1.42)
0.110
(1.61)
- o.ooo
- 0.000
- 0.001
( - 0.22)
( - 0.21)
( - 0.32)
0.034’
(2.11)
0.034’
(2.09)
- 0.005
( - 0.39)
- 0.005
( - O.s8)
0.020
(1.29)
0.020
(1.27)
- O&16
( - 0.70)
F-ratio
1.58d
Adjusted R ’
0.13
Sample size
68
_~
..
‘See table 4~for definitions of variables.
bSignificant at the 0.01 level.
‘Significant iit the 0.05 level.
dSignifisant at the 0.10 level.
--
-- o.GoO
( - 0.17)
- 0.016
- 0.017
( - 1.21)
( - 1.26)
- 0.027b
( - 2.84)
- 0.028b
( - 2.90)
0.004
(0.35)
0.004
(0.34)
- 0.016
- ox)19
- 0.018
( - 0.69)
( - 0.78)
( - 0.74)
1.46
0.11
68
2.07’
0.17
69
1.93’
0.16
69
3.4. D$ere:lces in returns between targets of Japanese and of U.S. firms
The previaus analysis shows that shareholders of U.S. targets of Japanese
bidders do nret earn higher abnormal returns than those of targets in the controe
sample. In this section, I investigate the difference in target returns further by
contr-lling for other explanatory variables. As a first step, Kregress CAR( -- 2,1)
for a pooled sample of 98 targets of Japanese bidders and 98 targets of U.S.
bidders on $1country dummy and ownership variables, defined in table 4.
Regression (1) in table 8 indicates that controlling for ownership variables does
not change the statistical significance of the coe$c ent on the country dummy
variable (TA RGET OF JAPANESE BIDDER).
positive and signifi@ant
coefficient on management ownership (MGM il3
j is c0nsistent with tk
finding of Stu lz, Wallcling, and Song (1990).
,
Table 8
egression estimates of cumulative abnormal returns (CARS) of U.S. targets acquired by Japanese
and U.S. bidders on a country dummy and other explanatory variables (1975-1988).Weighted least
squares is used for the estimation. A target of the U.S. bidder is matched to each target of a Japanese
bidder by ww of acquisition, method of payment, industry, year, and market value of equity. Day
55uai3klb~&
Journal. The dependent
Le 1X.1-1’
@*,=-*
0 is the in1 mnouncement date of the bid identified from tllv
variable is the CAR from two days before day 0 to one day after the final revision by the succe~s~til
bidder (t-statistics in parentheses).
Regressions
Full period:
1975-1988
Variables”
Cl)
(2)
CONSTANT
O&Md
(l.731
0.184b
(3.87)
0.127’
(2.18)
- 0.000
( - 0.68)
0.219d
(1.96)
0’18
(1.59)
0.304’
(2.25)
- 0.021
( - 0.67)
0.001
(0.19)
0.273”
(5.68)
0.003
(0.49)
MGMTOWN
0.203b
(2.89)
BID0 WN
0.000
(0.48)
BANKBLKO
CORPBLKO
0.139
(0.86)
WN
0.102
(0.8 1)
WN
EN VBLKO WN
0.242
(1.25)
TAX81
TAX86
MULTIPLE
LOGTGTMV
TARGET
OF JAPANESE
BIDDER
Subperiod:
1975-1985
Subperiod:
1986-1988
-
(3)
(4)
0.290b
(5.15)
- 0.098
(-1.44)
-I‘
^61)
(-
3.045
(0.28)
0.027
(0.22)
0.291
(1.59)
0.127d
(1.79)
0.254”
(2.82)
- 0.001
( - 0.74)
0.258d
(1.67)
0.166
(1.37)
0.243
(1.30)
0.380b
(4.72)
- 0.016d
( - 1.99)
3.27Gb
(4.18)
0.009
(1.02)
0.011
(0.39)
TARGET OF JAPANESE
E!DDER/
SALE OF MAJORiTY
INTEREST
0.03iP
(2.35)
-0.X*
TARGET OF JAPANESE
BIDDER/
SALE OF UNIT OR MINORITY
INTEREST
( - 590)
TARGET OF U.S. BIDDER/
SALE OF UNIT OF MINORITY
INTEREST
(-
F-ratio
Adjusted R ’
Sample size
“See table 4 for definitions of variables.
bSignificant at the !?01 let el.
‘Significant at the 0.05 level.
dSignificanr dt the 0.10 leaJe1.
- 0.199b
5.48)
1.98d
0.03
196
LO.i32b
c .;:
19b
- 0.000
( - 0.00)
(i
.,.183b
4.61)
- 0.186b
9.75)
(-
12.41b
0.56
92
---
0.131’
(2.05)
- 0.207b
( - 3.39)
- 0.17gb
2.89)
( -
13Mb
0.55
104
J.-K. Kang, The international m&rrkctfor corporate control
359
In regression (2), I a d other variables, including interaction terms between
country and acouisitioJi
,
type dummies. The main finding of regression (2)
is that the sale o! .i majb_+ty interest to Japanese bidders leads to signifi.cantly
higher target rea;lrns than the sale of a majority interest to U.S. bidders.
In particular, this significance is robust to the choice of different event
tvindows such as ( - l,O), ( - 1, l), and ( - 5,l). Therefore, the results
suggest that Japanese bidders pay large premiums for U.S. targets only when
they acquire a majority interest, and that the degree of control in U.S.
target resources is an important determinant of Japanese takeover p
The results for subperiods [regressions (3) and (4)], however, indic
U.S. targets of Japanese bidders earn these large premiums mainly during
the 1986-1988 subperiod. Finally, consistent with Bradley, Desai, and
Kim (19883, all regressions show that target shareholders earn larger
returns from multiple-bidder contests (MULTIPLE)than from single-bidder
offers.
4.Conclusions
I present evidence that Japaeese mergers and acqui; Lens; in the U.S. create
statistically significant wealth gaiE ‘ar both Japanese bidders and U.S. targets.
Consistent with the arguments advanced by Je;,den (1986), Fama (1990), 2-d
Froot and Stein (1991), I find that bidder-sticific characteristics and dollar- yin
exchange-rate movemenrs are useful in explaining the cross-sectional variation
in bidder returns: returns to Japanese bidders and returns realized by the
portfolio of Japanese bidders and U.S. targets increase with bidders’ total debt,
their ties to financial institutions through borrowings. and the depreciation of
the dollar. Tax variables, however, dc5not provide a satisfactory explanation for
the variation in Japanese bidder returns. For targets, I document that, after the
method of payment, industry, and other explanator;] variables are controlled
for, U.S. targets that sell a majority interest to Japanese bidders earn significantly higher abnormal returns than t ose selling to U.S. bidders. Subperiod
results indicate that the main findings for the overall sample period are driven
by the last three years of transactions (19861988).
Although this paper emphasizes some of the variables that may explain
Japanese bidder performance, it does not investigate other potentially relevant
factors. For example, acquisitions of U.S. targets ;by Japanese firms might have
the primary effect of conveying to the stock market information about the
bidd,er’s future investment strategy. Alternatively, acquisitions of U.S. targets
could provide valuable information about the bidder’s abhliiy to circumvent
present or anticipated U.S. trade restrictions. It is also possible that the size
of the U.S. market and its political stability provide substantial incentives r"or
such transactions, and contribute to the abnormal returns earned by Japanese
370
J.-K. Kang, The internationalmarketfor corporate control
bidders. This paper sheds no light on these issues An analysis of their importance represents a valuable area for future research.
References
A&i, Masahiko, 1984, Aspects of the Japanese firm, in: Masahiko Aoki, ed., The economic analysis
of the Japanese firm (North-Holland, Amsterdam) 3-43.
&q&h, Paul,Robert F. Bruner, and David W. Mullins Jr., 1987, Merger returns and the form of
financing, Unpublished paper (Harvard University, Boston. MA).
Bradley, ’ ichael, Anand Desai, and E. Han Kim, 1988,Synergistic gains from corporate acquisitions and their division between the stockholders of target and acquiring firms, Journal of
Financial Economics 21, 340.
Bradley, Michael, Gregg A. Jarrell, and E. Han Kim, 1984,On the existence of an optimal capital
structure: Theory and evidence, Journal of Finance 39,857-878.
Caves, Richard E., 1971,International corporations: The industrial economics of foreing investment,
Economica 38, l-27.
Fama, Eugene F., 1990, Contract costs and financing decisions, Journal of Business 63. 71-91.
French, Kenneth R. and James M. Poterba, 1991,Were Japanese stock prices too high?, Journal of
Financial Economics 29, 337-364.
Froot, Kenneth A. and Jeremy C. Stein, 1991, Exchange rates and foreign direct investment: An
imperfect capital market approach, Quarterly Journal of Economics 106, 1191-1217.
Harris, Robert S. and David Ravenscraft, 1991,The role of acquisitions in foreign direct investment:
Evidence from the U.S. stock market, Journal of Finance 46, 825-844.
Hodder, James E. and Adrian E. Tschoegl, 1985, Some aspects of Japanese corporate finance,
rnllfnat
“V
_. .._Lof Financial and Quantitative Analysis 20, 1~73-131.
Hoshi, Takeo, Anii Kashyap, and David Scharfstein, i991, Corporate structure, liquidity and
investment: Evidence from Japanese indLstria1 groups, Quarterly Journal of Economics i06,
33*.
Hymer, Stephen H., 1976,The international oxrations of national firms: A siudy of direct foreign
investment (MIT Press, Cambridge, MA!.
Jarrell, Gregg A., James A. Brinkley, aad Jeffry M. Netter, 1988,T’hemarket for corporate control:
The empirical evidence since 1980,Journal of Economic Perspectives 2.49-68.
Jensen, Michael C., 1986, Agency ccsts of free cash flows, corporate finance, and takeovers,
American Economic Review 76,323-329.
rgnsen, Michael C., 1989, Eclip8%of the public corporation, Harvard Business Review no. 5,
61-74.
Jensen, Michael C. and William I%Meckling, 1976,Theory of the firm: Managerial behavior, agency
costs and ownership structure, Journal of Financial Economics 3,305-360.
Jensen, Michael C. and Richard S. Ruback, 1983, The market for corporate control: i-he scientific
evidence, Journal of Financial Economics 11, S-50.
Kester, W Carl. 1986, Capital and ownership structure: A comparison of United States and
Japanese manufacturing corporations, Financial Management 15, 5-l 6.
Rester, W” Carl, 1991, Japanese takeovers: The global contest for corporate control (Harvard
Business SC oc Press, Boston, MA).
KindMerger, Gwites P., 1969, American business abroad: Six lectures on direct investment (Yale
University Press, New Haven, CT).
Lewellen, Wilbur, Claudio Loderer, and Ahron Rosenfeld, 1985, Merger decisions and executi\lt
stock ownership in acquiring firms, Journal of Accounting and Economics 7, X9-231.
Lang, Larry, Renk M. Stulz, and Ralph A. Walkling, 1989,Tobins 4 pnd the gp’--”from successful
tender offers, Journal of Financial Economics 24, 137-154.
Mikkelson, Wayne H. and Richard S. Ruback, MU, An empirical analisis of the interfirm equity
investment process, Journal of Financial Economics 14: 523-553.
Maloney, Michad T. and Robert E. h4-C
,rlb~~~mick, 1989, Mergers and capital structure, Working
paper (Clemson University, Clemson, SC).
J.-K. Kang, The internotionalma*ket-for corporate comd
371
Merck, Randall, Andrei Yyleifer,and Ro
skny, 1988,Characteristics of targets of hostile
and friendly takeovers, 1:Alan 1.Aue
Corporate talkeovers:Causes and consequences
(University of Chicago Press, Chicago, IL).
Nakatam, Iwao, 1984, The economic role of financial corporate grouping, in: asahike, Aoki, ed.,
The zca~~onncanalysis of the Japanese firm (North-Holland, Amste~dam~2?7-258.
Prowse, Stephen D., 1990,Institution al investment patterns and corporate fit ~7 -iaj behavior in the
United States and Japan, Journal of Financial Economics 27,43-66.
Roe, Mark J., 1990, Political and legal restraints on ownership and control of public companies,
Journal of Financial Economies 27, 7-4f.
Scholes, Myron S. and Mark A. Wolfson, 1999, The effects of ehanges in tax laws on corporate
reorganization activity, Journal of Business 63, 141-164.
Shleifer, Andrei and Robert W. Vishny, 1986,L.argeshareholders and corporate control, Journal of
Political Economy 94, 461-488.
Stulz, Rene M., Ralph A. Walkling, ancoMoon H. Song, 1990,The distribution of target ownership
and the division of gains in successful tender offers, Journal of Finance 45,817-833.