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Do Investment Banks Matter for M&A Returns?

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Do Investment Banks Matter for M&A Returns?
Jack Bao Fisher College of Business, Ohio State University Alex Edmans Wharton School, University of Pennsylvania and NBER
We document a significant investment bank fixed effect in the announcement returns of M&A deals. The interquartile range of bank fixed effects is 1.26%, compared with a full- sample average return of 0.72%. The results remain significant after controlling for the component of returns attributable to the acquirer. Our findings suggest that investment banks matter for M&A outcomes, and contrast earlier studies that show no positive link between various measures of advisor quality and M&A returns. Differences in average re- turns across banks are also persistent over time and predictable from prior performance. Clients do not chase past returns, which may explain why persistence exists in M&A per- formance while it is absent in mutual funds. (JEL G24, G34)
Mergers and acquisitions (M&A) are among the most critical decisions a CEO can make. Successful mergers can create substantial synergies, while mis- guided acquisitions can lead to misallocation of companies to parents unable to reap their full potential. In addition to these large effects on shareholder value, a bad acquisition also increases the CEO��s risk of being fired (Lehn
and Zhao 2006). A prominent example is the departure of Carly Fiorina from
Hewlett Packard, which was widely attributed to her acquisition of Compaq. The quality of M&A transactions is also of great importance to the economy as a whole. The total value of M&A announced by a U.S. acquirer in 2007 was $2.1 trillion, around 15% of GDP.
We thank an anonymous referee, the Editor (Alexander Ljungqvist), Franklin Allen, Alexander Dyck, Florian Ederer, Vincent Glode, Itay Goldstein, Todd Gormley, Dirk Hackbarth, Kewei Hou, Jeff Jaffe, Dirk Jenter, Adam Kolasinski, Anna Kovner, Gilberto Loureiro, Qingzhong Ma, Stew Myers, Micah Officer, David Pedersen, Nagpurnanand Prabhala, Jun Qian, Michael Roberts, Rob Stambaugh, Ren�� Stulz, Torben Voetmann, Mike Weisbach, Fei Xie, and seminar participants at the 2008 WFA meetings, 2008 FIRS meetings, 2007 EFA meetings, BYU, Exeter, MIT Sloan, New York Fed, Toronto, Vanderbilt, Virginia, and Warwick for valued input. Special thanks to Jerry Hoberg, Wei Jiang, and Luke Taylor for extensive insightful com- ments. We are grateful to Michael Roberts for generously providing Dealscan-Compustat data, and to Cong Wang for data that were used in an earlier draft of this project. This article was previously circu- lated under the titles ��Do Investment Banks Have Skill? Performance Persistence of M&A Advisors�� and ��How Should Acquirers Select Advisors? Persistence in Investment Bank Performance.�� Send correspon- dence to Alex Edmans, 3620 Locust Walk, Philadelphia, PA 19104; telephone: (215) 746-0498. E-mail: aedmans@wharton.upenn.edu. c The Author 2011. Published by Oxford University Press on behalf of The Society for Financial Studies. All rights reserved. For Permissions, please e-mail: journals.permissions@oup.com. doi:10.1093/rfs/hhr014 Advance Access publication May 6, 2011
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Do Investment Banks Matter for M&A Returns?
Since CEOs make M&A decisions rarely, they typically lack experience and seek counsel from investment banks. The skilled-advice hypothesis is that banks help clients identify synergistic targets and negotiate favorable terms. If banks indeed provide valuable advice, it is reasonable to expect that the highest-quality advisors lead to the best outcomes. However, existing research generally fails to find such a relationship. Bowers and Miller (1990) and Michel,
Shaked, and Lee (1991) measure an advisor��s quality by the prestige of its
name and find no link with acquirer returns; Rau (2000) uses market share to measure quality and documents a negative relationship. Servaes and Zenner (1996) find no benefit of hiring any advisor at all, compared with executing the deal in-house.1 These findings instead appear to support the passive-execution hypothesis, that banks do not provide true advice but are simply ��execution houses,�� which undertake deals as instructed by the client. If true, such a con- clusion has troubling implications. The investment banking industry, which consumes a significant proportion of an economy��s talented human capital, is mainly a deadweight loss to society. CEOs�� inexperience in M&A is not mitigated by hiring an advisor, which may explain why so many acquisitions destroy value. This article reaches a different conclusion. Prior studies investigate the im- portance of investment banks for M&A outcomes by hypothesizing a measure of advisor quality, such as market share or name prestige, and correlating it with this measure of quality. Such studies will find significant results only if their chosen measures are truly accurate proxies of ability. We instead em- ploy a fixed-effects analysis. This is a broader approach that examines whether banks exhibit differential deal returns in the first place, without having to spec- ify a measure of advisor quality with which any differential will be correlated. Indeed, we find significant bank fixed effects to a deal��s three-day cumula- tive abnormal return (CAR). Studying all banks that advised on at least ten deals over 1980–2007 and controlling for time effects, the difference between the 25th- and 75th-percentile banks is 1.26%. This difference is economically meaningful applied to the mean bidder size of $10 billion and compared with the mean CAR of 0.72%. An F-test that bank fixed effects are equal is rejected at the 1% significance level. Our results support the skilled-advice hypothesis and contrast prior findings that banks have little effect on M&A outcomes, as predicted by the passive-execution hypothesis. Returns analyses have also been used to evaluate skill in mutual funds, hedge funds, and security analysts. Our setting shares two challenges also faced by studies of stock-picking ability. The first is performance attribution: Returns
1 To our knowledge, only Kale, Kini, and Ryan (2003) find gains to employing market-leading advisors. They
study 324 contested takeovers of public targets, and find that large banks are more likely to withdraw when the price becomes too high. By contrast, both we and Rau (2000) find a negative link between market share and performance across all M&A transactions (over 15,000 in our sample), of which approximately one-third are public. One reason may be that the incentives to act in the client��s interest are far stronger in public situations, where ��honest�� advice to withdraw from a deal is widely observed.
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are not purely the responsibility of the financial intermediary. In an invest- ment setting, returns also depend on the portfolio��s factor loadings and re- alized factor outcomes. Since investment skill depends on how a portfolio performs over the long run, investment studies typically investigate long- horizon returns. Therefore, the results are highly contingent on the benchmark asset-pricing model used (Fama 1998). Benchmark adjustment is less of an issue here, since performance can be measured by short-horizon announcement returns: In an efficient market, they capture the full value impact of an acqui- sition. Our setting faces a different performance attribution challenge—CAR may be the responsibility of either the bank or the client. A bank may be asso- ciated with positive (negative) CARs if it is systematically mandated by high- quality (empire-building) clients. Many prior studies (e.g., Bowers and Miller
1990; Michel, Shaked, and Lee 1991; Rau 2000; Hunter and Jagtiani 2003) do
not tackle performance attribution and assume CAR results entirely from the bank. Others control for deal characteristics (e.g., Servaes and Zenner 1996;
Kale, Kini, and Ryan 2003) but acknowledge that this solution may go too far
the other way, since deal characteristics are often the advisor��s responsibility.2 We control for the component of CAR that can be explained by acquirer char- acteristics that proxy for the likelihood that the client is empire building (such as free cash flow and various governance measures, as used by Masulis, Wang,
and Xie 2007) and high quality (such as stock and operating performance and
Tobin��s Q). We also add acquirer fixed effects to proxy for time-invariant un- observable measures of quality. Even after these controls, the bank fixed effects remain statistically and economically significant, with an interquartile range of 1.23%. A second shared challenge is that average returns depend on not only ability, but also scale. Maksimovic and Phillips (2002) argue that conglomerate firms�� lower productivity arises because they are able to accept all projects with posi- tive NPV, including those with modestly positive NPV, whereas single-segment firms with financing constraints can pursue only those with high NPV. Applied to our setting, this limited-capacity hypothesis posits that banks differ not in ability, but in capacity to accept mandates. Small banks can work only on the highest-return deals; large banks also accept mandates with small (but posi- tive) value and thus exhibit lower average returns. We refute this hypothesis by showing that the banks with the lowest average returns are not the most fre- quent advisors—the correlation between bank fixed effects and market share is an insignificant 0.03 (p = 0.72). Indeed, the most frequent advisors occupy the middle of the fixed-effects distribution. While addressing the limited-capacity hypothesis, the finding that the most positive and most negative fixed effects are associated with infrequent advisors
2 For example, Servaes and Zenner (1996) caveat their conclusion by acknowledging that ��it is not certain that
the [deal characteristics] affecting investment banking choice are exogenous. For example, it is possible that investment banks influence the form of payment or the decision to pursue the acquisition.��
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Do Investment Banks Matter for M&A Returns?
may give rise to two quite different concerns. First, these extreme fixed effects may result from measurement error arising from few observations, rather than a ��true�� dispersion of advisor quality. Second, it may be that there is little dispersion among the major banks that matter more for the M&A market— i.e., the result is not generalizable to the most important advisors. Note that infrequent advisors will not drive the statistical significance of our results: If many banks have imprecisely measured fixed effects, the F-test will have lit- tle power. To address concerns that they drive our economic magnitudes, the interquartile range reported earlier is based on fixed effects weighted by the inverse of their standard errors. Moreover, to further investigate whether our results hold among frequent advisors alone, we then test the equality of bank fixed effects on a subsample of the largest banks that advise on at least 84 deals over the 28-year sample period (i.e., 3 per year). Even though the num- ber of banks falls markedly from 143 to 42, the results stay significant with an interquartile range of 0.74% (controlling for acquirer characteristics and ac- quirer fixed effects) and the F-test of equality is rejected at the 1% level. Thus, there are meaningful differences even among the most active advisors. Having documented that banks are associated with different CARs over the entire period, we then ask whether these differences are predictable based on historical data, and thus can be used by clients to guide their selection of advi- sors. The fixed effect implies a persistent component to a bank��s CAR and thus provides us with a priori motivation to predict future returns using past returns, rather than the market share and prestige measures previously studied. Indeed, we find that performance is persistent: The top quintile of banks based on CAR over the past two years outperforms the bottom quintile by 0.94% over the next two years (significant at 1%). Persistence remains after removing the compo- nent of CAR that can be attributed to client characteristics. Inconsistent with the limited-capacity hypothesis, the low CARs of the bottom-quintile banks do not arise from executing mildly positive transactions, but from executing double the proportion of value-destructive deals as the top quintile. Regression analyses yield similar results: A bank��s average returns are significantly linked to its past average returns, even when controlling for market share. When inter- acting past returns with market share, the interaction term is insignificant, but past returns alone remain significant, suggesting that the effect of past returns is similar for both frequent and infrequent advisors. The existence of persistence in M&A advice contrasts the lack of persistence in mutual funds. Berk and Green (2004) show that even if mutual funds possess skill, this does not translate into persistence if investor flows respond to past performance and there are diminishing returns to scale. We find that a bank��s market share is independent of its past CAR (also found by Rau 2000). This lack of performance chasing can potentially reconcile why persistence exists in M&A advice but not mutual fund performance (e.g., Carhart 1997). Instead, we find that mandate awards are highly correlated with past market share, even though market share negatively predicts future performance. 2289
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The selection of high-market-share rather than high-CAR advisors has two quite different interpretations. First, it may be efficient, if clients build up relationship-specific capital with particular banks and thus rationally retain the same advisor irrespective of past performance; market share has predictive power, as it measures a bank��s existing relationships. We show that retaining a past advisor is associated with worse future performance, particularly if the bank advised on negative-CAR transactions in the past, which does not support the view that relationship-specific capital improves future M&A performance. However, clients may derive other services from banking relationships, such as lending and underwriting, which rationally induces them to retain their ex- isting bank (Yasuda 2005; Ljungqvist, Marston, and Wilhelm 2006, 2009). Al- ternatively, it may be inefficient if clients are not locked into relationships but voluntarily choose advisors based on market share, under the misperception that it predicts superior returns. Such behavior is consistent with real-life prac- tices: Market-share league tables are widely publicized by both the media and the banks themselves, and so both academics and practitioners have come to use them as a measure of expertise.
Ertugurul and Krishnan (2010) also study the existence of ability in invest-
ment banking. They focus on individual bankers who switch advisors, rather than on banks themselves.3 Another difference is that in addition to identifying a fixed effect in the full sample, we also investigate persistence and thus the predictability of future outcomes using past performance. Jaffe, Pedersen, and
Voetmann (2009) demonstrate persistence in M&A performance at the client
level. Mikhail, Walther, and Willis (2004) and Hoberg (2007) document persis- tence in two other banking services, security analysis and equity underwriting. This article proceeds as follows. Section 1 discusses the potential sources of differential M&A returns across banks, and Section 2 describes the data. Section 3 documents significant advisor fixed effects to M&A returns and shows that average returns are predictable using past performance. Section 4 shows that mandate awards are correlated not with past performance but with market share. Section 5 concludes. 1. Motivation: Why Might Banks Be Associated with Differential M&A Returns? This section discusses a theoretical framework for why there may be a bank fixed effect to M&A returns, i.e., why certain banks may be systematically
3 We study banks rather than individual bankers for two reasons. First, a transaction typically leverages resources
across the entire bank (e.g., a debt-financed acquisition of a German chemicals target by a UK pharmaceuticals firm may involve the M&A and debt product groups and the pharmaceuticals, chemicals, UK, and Germany coverage teams). Second, it is difficult to know which particular banker worked on a certain deal. A bank��s chemicals team consists of several bankers, many of whom will not be involved in the deal. If a bank��s skill hinges on particular star bankers (who often move between firms) rather than the whole organization, we should find weak bank fixed effects and bank-level persistence.
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Do Investment Banks Matter for M&A Returns?
associated with high- or low-return deals. To understand the possible sources of correlation, we first outline the role that advisors play in M&A deals. Their actual level of involvement can vary significantly across transactions, and falls under three broad categories. In a bank-initiated deal, the advisor is involved in both selecting the deal and negotiating terms, and thus is responsible for the entire CAR. In a stan- dard client-initiated deal, the client proposes the transaction but lacks the ability to identify good deals and so relies on the bank to advise whether to pursue it. Since the bank can reject a bad deal, it is again responsible for deal selection in addition to negotiation, and thus the entire CAR. Not all banks will reject the deal, but this failure to reject is for reasons within their re- sponsibility. Some lack the ability to identify bad deals; others know that a deal will destroy value but undertake it to maximize their own fee income rather than pursuing the client��s interests. A bank cannot blame low CARs on having to work on bad deals, since it controls its deal flow—just as a lender cannot blame losses on poor credit quality, since it controls the loans it chooses to write. In a fixated client deal, the acquirer has already decided on the target and does not seek advice on its appropriateness; it uses the bank simply to exe- cute the transaction on the best terms possible. This may occur in two cases. First, the client may be skilled in identifying targets and does not need the bank��s input. At the other extreme, the client is empire building or hubristic and wishes to pursue a bad deal even if the bank cautions otherwise. By accepting the mandate, the bank may be adding value compared with the counterfactual of the client pursuing the acquisition with a rival advisor. The bank is not re- sponsible for the component of CAR that can be attributed to the acquirer��s skill or hubris, only the orthogonal component. Given banks�� varying levels of involvement, systematic differences in aver- age returns may stem from three sources. The skilled-advice hypothesis is that certain advisors possess ability, either in proposing targets (for bank-initiated deals) or in negotiating terms (for all deals). Alternatively, variation may stem from systematically turning away bad deals. This requires ability to identify such deals, plus trustworthiness to turn them down. We use the term ��skilled advice�� to include these three qualities of deal identification, transaction ne- gotiation, and trustworthiness. Thus, even a bank with high-quality employees may be classified as low-ability if its business model is to accept every man- date, regardless of whether it will create client value. The passive-execution hypothesis is that banks lack ability in target selec- tion or deal negotiation. Instead, variation in returns arises because the bank is systematically mandated by skilled (empire-building) clients.4 In reality, banks
4 Fernando, Gatchev, and Spindt (2005) construct an assignment model where high-quality banks are systemati-
cally mandated by high-quality clients.
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exert substantial effort in pitching deals to clients: They allocate the majority of bankers to client coverage (rather than deal execution) groups, which are pri- marily responsible for pitching. Therefore, it seems unlikely that fixated client deals are sufficiently prevalent to explain differences in average returns. How- ever, since it is impossible to observe which party initiates a transaction and provide direct statistics on this prevalence, to be conservative we also report results controlling for the component of returns attributable to acquirer char- acteristics. The passive-execution hypothesis would also be supported if bank fixed effects are not significant in the first place. The limited-capacity hypothesis posits that banks differ not in ability, but in their capacity to accept mandates. A bank may exhibit a high average CAR because it can work on only the highest value–creating deals, whereas low CAR may arise if a bank has the capacity to execute also mildly good deals. We evaluate this hypothesis by investigating whether a bank��s low average CAR stems from advising on deals with modest value, or value-destructive deals.5 2. Performance Metrics, Data, and Descriptive Statistics We use Thomson Financial��s Securities Data Company (SDC) data for merg- ers announced between January 1980 and December 2007. Since deals that involve a change of control are most likely to affect acquirer returns, we re- tain only transactions categorized as ��Merger,�� ��Acquisition,�� ��Acquisition of Assets,�� or ��Acquisition of Majority Interest�� and drop all deals for which the acquirer��s initial stake exceeded 50% or its final stake was below 50%. We also drop transactions for which the acquirer had no stock returns in the Center for Research in Security Prices (CRSP) database or the deal value was below $1 million (as in Rau 2000). Our final sample contains 15,344 deals. Our value-creation measure is the acquirer��s (−1, +1) CAR over the CRSP value-weighted index, which we winsorize at 1% and 99%.6 Stock returns are the relevant performance measure, as they represent the change in shareholder wealth, capitalizing all of the future effects of an acquisition; they are thus used in the vast majority of investment-banking studies (e.g., Bowers and Miller
1990; Servaes and Zenner 1996; Rau 2000). While CAR refers to one specific
deal, RET is the average CAR to all deals advised by a bank in a j-year period.
5 The limited-capacity hypothesis is less likely for investment banks than corporations or mutual funds. Even small
banks are able to work on very large transactions—for example, the boutique Gleacher employs 50 staff and ad- vised on Bank of Scotland��s $40 billion merger with Halifax, AT&T��s $22 billion sale to SBC Communications, and MFS Communications�� $14 billion merger with WorldCom.
6 We also obtain beta model returns from Eventus and find similar results. The correlation between beta model
returns and returns above the CRSP value-weighted (VW) index is 98.5%. Since the beta model cannot be calculated for several acquirers, we use returns above the CRSP VW index. In addition, Hackbarth and Morellec (2007) show that betas change substantially upon a merger, and so a beta calculated based on historical data is likely to be misleading. We use the CRSP VW index as a benchmark, because Rau and Vermaelen (1998) document biases when using size and book-to-market adjusted CARs.
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Do Investment Banks Matter for M&A Returns?
To be included in the analysis, a bank must have announced at least 2 j deals within the period.7 Some papers attribute the entire CAR to the bank, which constitutes an over- attribution in fixated client deals. Others remove the component explained by deal characteristics, but this leads to an underattribution, since deal charac- teristics may be chosen by the advisor, either directly by initiating the deal or indirectly by accepting a client-proposed mandate. We control for acquirer characteristics that proxy for client quality or empire building, since they are outside a bank��s control, taking its client base as given. Note that banks may have some control over their client base: If a bank advises a standard client to abandon a bad deal, it does not enter its client base. Therefore, controlling for acquirer characteristics is conservative. A number of our characteristics are related to governance. Masulis, Wang,
and Xie (2007) find that governance mechanisms are significantly related to
acquirer returns. Their primary measure is the Gompers, Ishii, and Metrick (2003) index. Unfortunately, it is not suitable here because it is available only from 1990 and we require a long time series to test for persistence; more- over, it is available for only a subset of firms in a given year. We therefore include other governance mechanisms studied by Masulis, Wang,
and Xie (2007): institutional ownership, leverage, and product market com-
petition (measured by the Herfindahl index and the industry��s median ratio of selling expense to sales). The second group of characteristics are proxies for acquirer quality, also from Masulis, Wang, and Xie (2007): Tobin��s Q, stock price runup, and operating performance. We also use the other bidder characteristics studied by Masulis, Wang, and Xie (2007): free cash flow (which may facilitate empire building) and size (which Moeller, Schlingemann, and
Stulz 2004 show is negatively correlated with returns). Since omitted acquirer
characteristics may overattribute CAR to the bank, we add additional controls over and above those featured in prior literature. We include inside ownership from Compact Disclosure, to measure management��s alignment with share- holders. Where missing, we impute it using firm sales and age.8 To proxy for empire-building intent, we include the number of acquirer Standard Industrial Classification (SIC) codes and a dummy for whether it made an acquisition in the previous five years. Finally, we include dummies for the bidder��s Fama-French industry.9 Full variable definitions are given in Table 1.
7 Where a deal has multiple advisors, the deal is credited to each advisor separately. This is consistent with how
SDC constructs market-share league tables.
8 Specifically, we winsorize sales at 1% and 99% and regress inside ownership on sales and age. We then use the
coefficients to predict inside ownership for the firms where it is missing. The R2 of the first-stage regression is 13%.
9 We use acquirer industry fixed effects rather than running the analysis for each industry separately (i.e., studying
the fixed effect of a particular bank-industry group) because very few banks undertake at least 2 j transactions within a given industry in j years, the minimum required to calculate an accurate RET measure.
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The Review of Financial Studies / v 24 n 7 2011 Table 1 Definition of variables used in the analyses Panel A: Used in the calculation of residuals for CAR Variable Definition RUNUP Log stock return for the acquirer from −210 to −11. Q Log of Tobin��s Q. Tobin��s Q = Market value of assets / Total assets (#6). Market value of common stock = Common shares outstanding (#25) * Price (#199). Market value of assets = Book value of assets (#6) + Market value of common stock – Book value of common stock (#60) – Balance sheet deferred taxes (#74). LEVERAGE LEVERAGE = Book debt / (Total assets (#6) – Book equity + Market equity). Book equity = Total assets (#6) – Total liabilities (#181) – Preferred stock (#10) + Deferred taxes (#35, if available) Substitute Redemption value of preferred stock (#56) if Preferred stock is missing. Book debt = Total assets (#6) – Book equity. Market equity = Common shares outstanding (#25) * Price (#199). FCF FCF = Free cash flow / Total assets (#6). Free cash flow = Operating income before depreciation (#13) – Interest expense (#15) – Income taxes (#16) + Deferred taxes and investment tax credit (#35 - #35 from previous year) – Preferred dividends (#19) – Common dividends (#21). SIZE Log of Total assets (#6). HERFINDAHL ��i ( firm salesi(#12)
industry sales )2
, where industries are defined by the Fama-French 49 industries. SELLEXP SELLEXP = median selling expenses (#189) over Sales (#12) for industry. INST Fraction of outstanding common shares owned by institutions from Thomson Financial 13f filings. OPPERF Firm operating performance minus the industry median in the past year. Operating performance = Operating income before depreciation (#13) / 0.5 (Total assets + last year��s total assets (#6)). INSIDER Insider ownership as a fraction of total shares outstanding, from Compact Disclo- sure. Where missing, we impute it using Sales (#12) and firm age (from CRSP). ACQSIC Log of number of acquirer SIC codes. REPEAT ACQUIRER Dummy variable that equals 1 if the acquirer announced an acquisition in the previous 5 years. Panel B: Constructed for direct use in quintile analysis and regressions Variable Definition RET Average CAR (3-day cumulative abnormal return) for deals advised by an invest- ment bank over a given number of years. RETRES Average residual from a regression of CAR on acquirer characteristics defined in Panel A. SHARE Market share by value of acquirer-advised deals for an investment bank over a calendar year. Where applicable, we include the Compustat item number in the description.
All variables are calculated for the fiscal year ending the year before deal announcement.10
10 Our regression of CAR on characteristics is run on the entire sample with year fixed effects. Using a rolling
window would cause data from the early period of the sample to be dropped and would also produce less precise estimates. Full-sample regressions are thus often used in asset pricing. We are not assuming that clients use past data to estimate the characteristics parameters for themselves when choosing banks. Instead, we posit that clients already have in mind a model of the effect of acquirer characteristics on returns, which they use to isolate the portion of CAR that is outside the bank��s control. As econometricians, we are attempting to estimate this model, for which we require the full sample.
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Do Investment Banks Matter for M&A Returns? Table 2 Regression of CAR on acquirer characteristics CAR RUNUP −0.0018 (1.07) Q −0.0055 (2.89)∗∗∗ LEVERAGE 0.0185 (4.05)∗∗∗ FCF −0.0569 (7.45)∗∗∗ SIZE −0.0033 (8.57)∗∗∗ HERFINDAHL 0.0008 (0.06) SELLEXP −0.0433 (2.22)∗∗ INST −0.0097 (4.06)∗∗∗ OPPERF 0.0475 (5.72)∗∗∗ INSIDER 0.0110 (2.08)∗∗ ACQSIC −0.0018 (1.92)∗ REPEAT ACQUIRER −0.0028 (1.89)∗ Year FE Yes Acquirer Industry FE Yes # obs 12,622 R2 (%) 4.17 * significant at 10%; ** significant at 5%; *** significant at 1%. CAR is the return in excess of the CRSP value-weighted index over a (–1, +1) window relative to the announce- ment date. The regressors are described in Table 1. The sample period is 1980–2007, and t-statistics are in parentheses.
We use CARRES to denote the residual CAR after controlling for acquirer characteristics, and define RETRES as the average CARRES over a j-year period. The regression results are shown in Table 2. Most coefficients are of the expected sign: Returns are increasing in leverage, operating performance, and inside ownership, and decreasing in free cash flow and the number of SIC codes. The R2 of 4% is commensurate with Masulis, Wang, and Xie��s (2007) R2 of 5%. Their R2 is marginally higher because they include deal characteristics, which are not appropriate here, since they are under the bank��s control. Since the bank is responsible for raw CAR in all but fixated client deals, it constitutes our core measure. As with any investment decision, an M&A transaction should be undertaken if the NPV, irrespective of project charac- teristics, exceeds zero. A bank cannot justify a negative-NPV transaction by arguing that other clients with, say, the same number of SIC codes undertook even worse deals if it had the option to turn away the deal in the first place. 2295
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3. Return Differences across Investment Banks 3.1 Full-sample fixed effects Most prior research on advisor ability attributes a deal��s CAR entirely to the bank and studies the association between average CAR and a variable hypoth- esized to proxy for bank quality, such as market share or reputation (measured by the prestige of the bank��s name). Such analyses will find significant results only if ability is correlated with their chosen measures of advisor quality. Thus, the absence of a link with market share or reputation need not imply that banks do not matter for M&A outcomes. We therefore take a broader approach. Rather than hypothesizing a measure of bank ability, we investigate whether banks exhibit differential CARs in the first place by estimating the bank fixed-effect component of a deal��s returns. We regress CAR on bank fixed effects while controlling for time fixed effects, since market enthusiasm for M&A may have varied over time. We then add acquirer characteristics to proxy for observable measures of quality or em- pire building. Finally, we add acquirer fixed effects to proxy for time-invariant unobservable measures of advisor quality. If two banks merge (e.g., Deutsche Bank buys Bankers Trust), we construct one fixed effect for the target (Bankers Trust) and a separate fixed effect for the acquirer (Deutsche Bank) both before and after the merger. The results are shown in Table 3 for the 143 banks that advised on at least ten deals over 1980–2007 or were acquired by a bank that advised on at least ten deals. Panel A finds that, in all specifications, the fixed effects are highly significantly different from each other (p < 0.01). Panel B demonstrates the economic significance of these differences. The difference between the 25th- and 75th-percentile banks is 0.9–1.3%,11 compared with the average CAR of 0.72% and the mean bidder size of $10 billion. To our knowledge, these results constitute the first large-scale evidence that certain banks are systematically as- sociated with superior M&A returns, and contradict prior findings that advisor quality has no positive effect on M&A outcomes. While supportive of the skilled-advice hypothesis, differential returns could also be consistent with the limited-capacity hypothesis, if the banks with the highest fixed effects advise on the fewest transactions. To investigate this, Figure 1 presents a scatter plot of a bank��s fixed effect against its market share (by number of deals). Inconsistent with the limited-capacity hypothesis, the most frequent advisors occupy the middle, rather than the lower end, of the distribution. There is a hump-shaped relationship between market share and fixed effects, and the correlation between the two variables is an insignificant −0.05 to 0.05.
11 Following Bertrand and Schoar (2003), we weight each fixed effect by the inverse of its standard error to account
for estimation error. Without this weighting, the interquartile range of fixed effects is 1.7–2.2%.
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Do Investment Banks Matter for M&A Returns? Table 3 Bank fixed effects to a deal��s CAR Panel A: Investment Bank Fixed Effects Controls Bank FE F-test Obs Adj R2 (%) (1) Time FE 1.65 (0.0000, 142) 15,344 0.97 (2) Acq chars, time FE 1.44 (0.0006, 141) 12,622 3.98 (3) Acq chars, acq FE, time FE 1.47 (0.0003, 138) 10,430 3.27 Panel B: Distribution of Bank Fixed Effects Std Dev 25th 75th Interquartile Range (1) 1.37% –0.91% 0.35% 1.26% (2) 1.39% –0.53% 0.41% 0.94% (3) 2.00% –0.65% 0.58% 1.23% Panel C: Fixed Effects for Active Banks Bank FE F-test Std Dev 25th 75th Interquartile Range (1) 1.69 (0.0038, 41) 0.64% –0.85% 0.21% 1.06% (2) 1.41 (0.0431, 41) 0.66% –0.45% 0.28% 0.74% (3) 1.60 (0.0092, 41) 0.86% –0.47% 0.27% 0.74% Panel D: Investment Bank Fixed Effects with Filters Bank FE F-test Banks FE F-test Controls (all banks) (active banks) Obs AdjR2(%) (1) Time FE 1.60 (0.0000, 142) 1.68 (0.0043, 41) 14,955 0.98 (2) Acq chars, time FE 1.45 (0.0004, 141) 1.45 (0.0305, 41) 12,347 3.62 (3) Acq chars, acq FE, time FE 1.48 (0.0002, 138) 1.70 (0.0036, 41) 10,235 3.37 Panel E: Distribution of Bank Fixed Effects with Filters Banks Std Dev 25th 75th Interquartile Range (1) All 1.36% –0.88% 0.34% 1.22% (1) Active banks 0.64% –0.80% 0.31% 1.11% (2) All 1.40% –0.74% 0.39% 1.12% (2) Active banks 0.67% –0.45% 0.17% 0.61% (3) All 2.00% –0.73% 0.51% 1.23% (3) Active banks 0.88% –0.53% 0.25% 0.78% Panel A reports F-tests for the equality of bank fixed effects from a regression of (–1, +1) CAR on bank fixed effects and listed controls. Acquirer characteristics are defined in Table 1. Panel B reports the distribution of bank fixed effects, weighted by the inverse of the standard errors of the fixed effects. Panel C reports analysis on fixed effects for banks that announced at least 84 deals over the full sample. Panels D and E repeat these analyses filtering out acquirers with stock prices below $1 and zero or missing trading volume on either the day after announcement or both two and three days before announcement. F-statistics, p-values, and numbers of constraints are listed. The number of constraints equals the number of bank fixed effects estimated in the regression minus one. In specification (3), the adjusted R2 for this specification is after removing acquirers that undertake only one transaction, and then demeaning by acquirer before regressing on acquirer characteristics, time dummies, and bank dummies. The sample period is 1980–2007.
While Figure 1 provides evidence against the limited-capacity hypothesis, it may raise a different concern: Perhaps the significant results of Table 3 are driven by infrequent advisors whose fixed effects are noisily measured, or who are relatively unimportant for the M&A market as a whole. To address this concern, the interquartile ranges of fixed effects that we reported above are after weighting each fixed effect by the inverse of its standard error. Moreover, we further investigate this hypothesis by testing for equality of fixed effects focusing only on the largest banks that advised on at least 84 transactions over 2297
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The Review of Financial Studies / v 24 n 7 2011 Figure 1 Plots of estimated bank fixed effects against number of deals announced in the full sample The fixed effects are estimated from regressions of the (–1, +1) CAR of deals on bank fixed effects and control variables. Plot (1) includes time fixed effects as controls, (2) includes acquirer characteristics and time fixed effects, and (3) includes acquirer characteristics, acquirer fixed effects, and time fixed effects.
the sample period, i.e., 3 per year. 42 banks meet this criterion. Panel C shows that the fixed effects remain jointly statistically significant at the 1% level, and the interquartile range is an economically meaningful 0.74%, even when continuing to control for acquirer characteristics and acquirer fixed effects. Another measurement concern is that while CAR measures the full value impact of a deal in an efficient market, it may understate it if part is incor- porated into prices before or after announcement. The former will occur if the deal leaks out early, bringing the measured returns of both good and bad 2298
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deals toward zero. The latter will occur if investors do not notice certain effects of the transaction until later and this failure is not unbiased (i.e., they fail to notice more good than bad effects, or vice versa). While long-run returns would capture a greater proportion of the transaction��s impact, they would also in- corporate many other corporate events (e.g., dividend and earnings changes not due to the acquisition) and hence suffer from a high noise-to-signal ratio. Moreover, errors resulting from failure to use the ��true�� benchmark model of stock returns are compounded over long horizons (Fama 1998). In the classical ��errors-in-variables�� problem, where measurement error is symmetric (i.e., the average error is zero) and similar across observations, mismeasurement simply attenuates the results. Our setting differs from the standard problem in two ways. First, mismeasurement arising from leakage or delayed reaction is asymmetric: Positive (negative) true returns are associated with negative (positive) errors. If the mean return were zero, positive and nega- tive true returns would be equally likely, and so the average error would also be zero—thus, the results would again be attenuated. By contrast, our mean CAR is positive, and so the average error is negative, biasing reported returns toward zero and thus below the mean. This would not be a problem if mismeasurement were similar across banks, since it would reduce measured RET evenly across the sample. However, our second difference is that mismeasurement may be more serious for certain banks. For example, small banks may advise small clients, whose deals are less closely followed by M&A arbitrageurs or the media, leading to less leakage and thus higher measured CAR. Thus, a low RET fixed effect may result from measurement error rather than underperfor- mance.12 We address the mismeasurement explanation in a number of ways. First, it is reasonable to assume that transactions with measured CARs exceed- ing 10% in absolute value did not suffer from attenuation. The remaining 87% of deals are the subset for which attenuation may be present. The mean CAR for this subset is −0.016%, very close to zero. Hence, any attenuation is in- deed toward the mean, as in the classic errors-in-variables setting, and leads to our results being understated.13 Second, in Panels D and E we filter out low- price and low-volume acquirers and show that the results are barely affected.14
12 For example, assume that banks A and B both execute deals with true value creation of 1.5%, 1.0%, and −1.0%
(i.e., a positive mean). Bank A��s CARs fully capture the value, and so its RET is 0.5%. Bank B��s CARs capture only half of the value due to leakage or underreaction, and so its CARs will be 0.75%, 0.5%, and −0.5%, yielding a RET of 0.25%.
13 Continuing the earlier example, assume that banks A and B both execute deals with true value creation of 12%,
1%, and −1%. Bank A��s CARs fully capture the value, and bank B��s CARs for the final two deals are halved. Since the deals that exhibit attenuation have a zero average return, both banks will have the same RET.
14 Specifically, we keep only cases where the acquirer��s price is at least $1 before the acquisition and where there
is trading volume either two or three days before the announcement and trading volume on the day after the announcement. These filters are similar in spirit to Diether, Lee, and Werner (2009), who use an initial $1 and zero volume screen. Our setting differs from the anomalies literature (e.g., Diether, Lee, and Werner 2009) in that we require only that CAR be a signal of value creation and do not require that it be tradable; thus, the M&A literature does not typically employ such filters. Further, CRSP reports the midpoint of bid and ask if a closing price is not available.
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For example, the interquartile range is 0.78%, even when focusing on the 42 largest banks and adding all controls. Third, we later show (in Panel D of Table 5) that the low average returns of low-RET banks stem from their undertaking a high proportion of value-destructive deals, rather than deals with small but positive measured value.15 To demonstrate the results on an individual bank level, Table 4 provides summary statistics on the fifteen largest banks by number of deals. There is significant variation in the average returns to each bank, which range from −0.12% (UBS) to 1.47% (Bank of America). Controlling for acquirer charac- teristics sometimes has a marked impact on banks�� performance measures: For example, Goldman Sachs has the second lowest fixed effect (−0.93%) when controlling only for year dummies, but it rises to an above-average 0.17% when adding acquirer characteristics and acquirer fixed effects. 3.2 Persistence in announcement returns While significant bank fixed effects suggest that advisors matter for M&A returns, the results of Table 3 are not actionable by clients in their selection
Table 4 Summary statistics for the top 15 investment banks by number of announced deals from 1980–2007 Number Market Share Bank FE Bank FE Bank FE Investment Bank of Deals by Value RET RETRES (1) (2) (3) Goldman Sachs 1,126 23.64% 0.07% 0.13% –0.93% 0.13% 0.17% Morgan Stanley 1,108 14.91% 0.09% 0.02% –0.91% –0.03% –0.31% Merrill Lynch 997 16.91% 0.18% –0.13% –0.85% –0.22% –0.47% CSFB 828 10.85% 0.38% –0.41% –0.60% –0.45% –0.70% SSB/Citigroup 821 16.74% 0.51% –0.00% –0.38% –0.08% 0.24% Lehman 803 8.08% 0.58% 0.15% –0.35% 0.07% 0.09% JP Morgan Chase 500 8.22% 0.28% 0.02% –0.48% 0.02% 0.27% Lazard 427 7.74% 0.46% 0.27% –0.47% 0.28% –0.24% DLJ 407 3.02% 0.68% –0.64% –0.45% –0.75% –1.12% Bear Stearns 374 4.43% 0.98% 0.23% –0.04% 0.14% 0.05% UBS Warburg 367 7.32% –0.12% –0.45% –0.98% –0.45% –0.41% Bank of America 299 4.21% 1.47% 0.67% 0.64% 0.65% 0.58% Salomon (pre-merger) 284 1.98% 0.66% –0.23% –0.56% –0.33% –0.62% JP Morgan (pre-merger) 267 3.41% 0.62% 0.43% –0.46% 0.41% 0.21% Deutsche Bank 244 4.34% 1.04% 0.82% 0.26% 0.88% 0.08% Avg over entire sample 15,344 0.72% 0.00% The averages provided in the last row include deals for all investment banks in the sample. RET is a bank��s average (–1, +1) CAR; RETRES is a bank��s average CARRES, the acquirer characteristic unexplained return (residuals from regressing CAR on acquirer characteristics). The three final columns display bank fixed effects as estimated in Table 3. Specification (1) includes time fixed effects; specification (2) includes time fixed effects and acquirer characteristics; specification (3) includes time and acquirer fixed effects, and acquirer characteristics.
15 A further hypothesis is that banks differ not in skill, but in the fees that they charge: Low RET banks may be
adding the same value as their rivals, but charging higher fees. We are unable to calculate ��pre-fee�� CARs, as fees are available for only 2,046 deals within our sample. However, among the deals that do have fee data, we find that the correlation between acquirer returns and fees (scaled by market cap) is a slightly positive 0.013, inconsistent with the hypothesis that low RET results from high fees.
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decisions, since they are based on the full 28-year sample. We therefore an- alyze whether clients can predict positive future returns based on historical data. The existence of a bank fixed effect implies a persistent component to a bank��s average CAR and thus provides a priori motivation for predicting future returns using an advisor��s past returns, rather than the market share and reputa- tion measures previously studied. This a priori motivation mitigates potential concerns of data-mining for advisor characteristics with predictive power. We calculate persistence in advisor performance in a similar manner to Jegadeesh
and Titman (1993) for stocks and Carhart (1997) for mutual funds. At the
start of each year, we sort banks into quintiles based on RET for the past j calendar years, where j = {1, 2, 3}. Next, for each quintile, we calculate RET for all banks within the quintile over the next k calendar years, where k = {1, 2, 3}. We report the difference in RET between the top (Q5) and bottom (Q1) quintiles.16,17 Table 5 illustrates the results. Panel A documents significant persistence in raw CAR in eight out of the nine time horizons. For example, when j = k = 2, the difference between Q1 and Q5 is 0.94%, significant at the 1% level. This result need not imply differential ability if fixated client deals comprise a sub- stantial proportion of all transactions, and so we next control for acquirer char- acteristics.18 Panel B illustrates persistence in the component attributable to advisors (RETRES). Thus, the persistence in RET does not arise because banks are systematically mandated by fixated acquirers, consistent with the skilled- advice hypothesis. The above panels calculate RET and RETRES using an equally weighted average of a bank��s CARs. Equal weighting is appropriate if one believes that each transaction is a separate measure of the bank��s ability and thus should be considered equally; it is also consistent with the main analysis on bank fixed effects, which treats each deal equally. Indeed, Mikhail, Walther, and Willis (2004) and Hoberg (2007), who study performance persistence in two other investment-banking services (analyst recommendations and equity underwrit- ing), also use equal weighting. However, weighting CARs by the size of the transaction may be appropriate if one believes that banks put greater effort into larger deals and so they are a more accurate measure of ability, or larger deals
16 As future returns are overlapping, we correct for mechanical autocovariance by using Newey-West standard
errors for panel data.
17 To illustrate our treatment of bank mergers, we continue with the example of the Deutsche Bank (DB) and
Bankers Trust (BT) merger in June 1999. Consider a regression of two-year RET on past two-year RET. For any observations where RET ends in 1998 or earlier, DB and BT enter separately, and both RET and RET are calculated on a stand alone basis. For any observations where RET ends in 1999 or later, we drop the two stand- alone observations and create one combined observation. Specifically, RET for 1998–1999 includes all deals advised by either DB, BT, or the merged entity during this period. To be consistent, the past RET measure will also include all deals advised by either bank or the merged entity in 1996–1997.
18 Since the persistence analysis uses short windows of one to three years, we do not include acquirer fixed effects,
because identifying purely on repeat acquirers who switch advisors would mean that very few deals are used to calculate RETRES and lead to substantial noise.
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The Review of Financial Studies / v 24 n 7 2011 Table 5 Persistence in a bank��s average returns and source of low returns Panel A: Persistence in Raw Returns Future RET Measured Over Quintiles Measured Over 1yr 2yrs 3yrs 1yr RET Q1 0.92% 0.66% 0.69% Q5 1.49% 1.48% 1.44% Q5-Q1 0.57% 0.82% 0.76% (1.32) (2.73)∗∗∗ (2.85)∗∗∗ 2yrs RET Q1 0.52% 0.61% 0.74% Q5 1.47% 1.55% 1.36% Q5-Q1 0.95% 0.94% 0.63% (2.58)∗∗ (3.07)∗∗∗ (1.99)∗∗ 3yrs RET Q1 0.59% 0.57% 0.60% Q5 1.68% 1.73% 1.54% Q5-Q1 1.09% 1.16% 0.94% (2.74)∗∗∗ (3.55)∗∗∗ (2.91)∗∗∗ Panel B: Persistence in Residual Returns Future RETRES Measured Over Quintiles Measured Over 1yr 2yrs 3yrs 1yr RETRES Q5-Q1 0.78% 0.80% 0.85% (1.71)∗ (2.51)∗∗ (3.32)∗∗∗ 2yrs RETRES Q5-Q1 1.35% 0.92% 0.55% (2.86)∗∗∗ (2.85)∗∗∗ (1.88)∗ 3yrs RETRES Q5-Q1 0.75% 0.55% 0.37% (1.76)∗ (1.82)∗ (1.29) Panel C: Persistence in Returns (Transaction value-weighted measures) Quintiles Future RET Measured Over Quintiles Future RETRES Measured Over Measured Over 1yr 2yrs 3yrs Measured Over 1yr 2yrs 3yrs 1yr RET 1yr RETRES Q5-Q1 0.62% 1.01% 1.00% Q5-Q1 0.95% 1.31% 0.82% (1.07) (2.01)∗∗ (2.22)∗∗ (1.63) (2.93)∗∗∗ (1.95)∗ 2yrs RET 2yrs RETRES Q5-Q1 0.98% 0.98% 0.81% Q5-Q1 0.38% 0.65% 0.23% (1.79)∗ (1.88)∗ (1.52) (0.69) (1.26) (0.43) 3yrs RET 3yrs RETRES Q5-Q1 0.45% 1.36% 1.07% Q5-Q1 0.62% 0.75% 0.50% (0.84) (2.47)∗∗ (2.01)∗∗ (1.08) (1.27) (0.80) Panel D: Percentage of Positive CAR deals, by RET Quintile RET Measured Over 1yr 2yrs 3yrs Q1 26.51% 34.90% 37.56% Q5 71.64% 67.41% 64.67% Q5-Q1 45.13% 32.51% 27.11% (25.12)∗∗∗ (22.07)∗∗∗ (19.26)∗∗∗ (continued)
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Do Investment Banks Matter for M&A Returns? Table 5 Continued Panel E: Persistence by Size of Advisor Large Advisors Small Advisors Terciles Future RET Measured Over Halves Future RET Measured Over Measured Over 1yr 2yrs 3yrs Measured Over 1yr 2yrs 3yrs 1yr RET 1yr RET T3-T1 0.40% 0.48% 0.41% H2-H1 1.07% 0.80% 1.24% (1.50) (2.53)∗∗ (2.29)∗∗ (1.62) (1.62) (2.79)∗∗∗ 2yrs RET 2yrs RET T3-T1 0.45% 0.43% 0.33% H2-H1 1.69% 1.04% 0.96% (1.82)∗ (2.23)∗∗ (1.58) (3.12)∗∗∗ (2.50)∗∗ (2.48)∗∗ 3yrs RET 3yrs RET T3-T1 0.48% 0.41% 0.37% H2-H1 1.17% 1.16% 1.02% (1.77)∗ (1.97)∗∗ (1.73)∗ (2.26)∗∗ (2.56)∗∗ (2.46)∗∗ * significant at 10%; ** significant at 5%; *** significant at 1%. Panel A sorts banks into quintiles based on their RET (average (–1, +1) CAR) over the past j calendar years, where j = {1,2,3}. To be included in the analysis, a bank must have announced at least 2j deals over the relevant period. Q1 represents the banks with the lowest past RET, Q5 the highest. For each quintile, we then calculate the average CAR to future acquisitions announced by the banks in that quintile over the next k calendar years, where k = {1,2,3}. Panel B repeats the analysis for RETRES (average of CARRES, the acquirer characteristic unexplained return). Panel C examines RET and RETRES calculated using transaction-value weighted measures. Panel D studies the correlation between RET and the contemporaneous percentage of positive CAR deals. Panel E subdivides the sample into the top 2/3 of banks by number of deals in each sorting period (Large Advisors) and the bottom 1/3 (Small Advisors); these banks are sorted each period into terciles and halves, respectively. The sample period is 1981–2007. Newey-West t-statistics are in parentheses.
are more complex and thus a truer test of advisor quality. Panel C repeats the analysis under size weighting. The results for our core performance measure, RET, are now stronger in terms of economic significance in eight of the nine cells and are at least 10% significant in six. The results for RETRES remain positive in all cells but are significant in only two. While suggestive of persistence, the above results admit other interpreta- tions. A notable feature of Panel A is that the average returns are positive even for Q1. Thus, it is consistent with both the limited capacity and mismeasurement hypotheses: Q1��s low RET arises not due to bad deals, but due to deals with modest measured returns—either because they actually generate small value, or because they actually generate large value that is undermeasured. We ad- dress both interpretations by calculating the ��success ratio�� of each bank: the percentage of deals that have a positive CAR. The correlation between one- year RET and bank success ratio is 0.68. Panel D illustrates that Q5 has ap- proximately double the success ratio of Q1, 65–72% compared with 27–38%. Therefore, inconsistent with these hypotheses, Q1��s low RET stems from a high proportion of value-destructive deals, rather than from deals with small but positive measured returns. Additional suggestive evidence against the limited-capacity hypothesis is in Figure 1, which shows no correlation between a bank��s RET and its market share. While Table 5 is a bank-level analysis, which considers each bank equally, another approach is a deal-level analysis. The Online Appendix shows that the 2303
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results remain robust—deals where the advisor was in the top quintile based on two-year prior performance outperform the bottom quintile by 0.91%, signifi- cant at the 1% level. This persistence continues to hold after controlling for ac- quirer characteristics. The Online Appendix also considers other performance measures than CAR. It documents persistence in a bank��s completion ratio and speed of completion and shows that these are weakly positively correlated with RET. As in Section 3.1, we also subdivide the analysis into the largest and smallest banks. Note that such a division significantly reduces power by lowering our sample size. Other persistence analyses in corporate finance typically study accounting variables (e.g., Bertrand and Schoar 2003; Lemmon, Roberts, and
Zender 2008), which are directly under the manager��s control. Moreover, many
accounting variables are naturally persistent: For example, Lemmon et al. study the level of leverage, which is naturally persistent, as any changes to leverage are with respect to last period��s leverage as a starting point. Thus, far fewer ob- servations are needed to achieve power. By contrast, M&A returns start from a ��clean slate�� each time—the starting point is zero, rather than the return of the last M&A deal. Moreover, M&A returns are notoriously noisy because they reflect the market��s reaction rather than a variable controlled by management; indeed, previous studies of M&A returns (e.g., Masulis, Wang, and Xie 2007) find low R2 values, implying that M&A returns are difficult to explain. Thus, stratifying the sample will markedly reduce the statistical power of our tests, as well as lower the cross-sectional variation within each subsample. Neverthe- less, we perform such a stratification to investigate the source of the significant results in Table 5. We define ��large banks�� as those in the top two-thirds by number of deals in the period over which past performance is measured, and ��small banks�� as those in the lowest third. We divide banks into terciles of past performance in the large-bank subsample and halves within the small-bank subsample, rather than quintiles, to obtain approximately the same number of banks in each quantile as before (since 2/3 * 5 is close to 3 and 1/3 * 5 is close to 2). While the number of banks remains similar within each quantile, moving from quintiles to terciles or halves reduces the variation between the top and bottom quantiles. The left side of Panel E of Table 5 gives the RET results for large banks. Seven of the nine cells are significant, which suggests that persistence does ex- ist among the large-bank subsample alone. The results are slightly weaker than in Table 5, consistent with reduced power. The right side of Panel E demon- strates the findings for small banks. The results remain significant in seven of the nine cells, even though the sample size has fallen by two-thirds. One of the motivations for removing small banks is that their RET may be measured with significant error given the small number of deals. Thus, the variations in bank fixed effects in Table 3 may reflect noise rather than true differences in performance. However, while noise may explain cross-sectional variation at a given point in time, it is unlikely to account for time-series persistence—noise 2304
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implies that returns might be measured to be very high in one period, and then very low in an adjacent period. By contrast, we find evidence of persistence for even infrequent advisors, suggesting that their return differences reflect true variation in performance rather than noise. 3.3 Regression analysis The analysis of Section 3.2 studies differences in average performance between the top and bottom quantiles. To ensure that the results are not driven purely by banks at the extremes (i.e., the best and worst banks alone), we conduct a similar analysis using regressions. In addition to using all banks in the sample, regressions also allow us to control for other potential predictors of past re- turns. In particular, Rau (2000) finds that RET is negatively related to a bank��s past market share; he does not examine the effect of past returns. We wish to examine whether the explanatory power of past RET remains when controlling for market share. We run the following pooled regression: RETt,t = ��t + ��RRETtj,t−1 + ��SSHAREtj,t−1, (1) where SHAREtj,t−1 is the market share over the past j calendar years, by dollar value of deals (using number of deals leads to similar results). Since we have shown that bank fixed effects are significant, and our regressors may not capture the full fixed effects, the residuals for deals advised by the same bank might be correlated. We therefore cluster standard errors by bank. The results are illustrated in Panel A of Table 6. The regressions replicate the posi- tive association between future RET and past RET documented in the quintile analysis, for all values of j. Market share is significantly negatively related to future returns in all three specifications in which it is included. The Online Appendix shows similar results in a deal-level analysis. We are again interested in whether our results apply to the M&A advisory market in general or are driven by small acquirers. Panel B adds an inter- action term between RET and SHARE. The interaction term is insignificant (t-statistics all below 0.5), suggesting that persistence is not weaker for larger banks. Moreover, the coefficients on RET and SHARE alone retain their signif- icant coefficients in all specifications. While stratifying the quintile analysis of Section 3.2 significantly reduces sample size, a regression analysis allows us to use an interaction term to study how the results vary across banks of different sizes without such a reduction. In addition, while the coefficient on RET gives the significance of the marginal effect if SHARE = 0, we also include the co- efficient if SHARE = 5%, i.e., for a frequent advisor with a 5% market share. From Table 4, a bank with a 5% market share would be among the top ten advisors. The table demonstrates that the marginal effect of past RET remains significant even for frequent advisors. Another interesting question is whether ability differences have decreased over time. Morrison and Wilhelm (2007, 2008) show that investment-banking 2305
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The Review of Financial Studies / v 24 n 7 2011 Table 6 Determinants of RET Panel A: Determinants of Bank-Level RET 1yr 1yr 2yrs 2yrs 3yrs 3yrs Past j years RET 0.0958 0.0908 0.1974 0.1876 0.1789 0.1619 (2.22)∗∗ (2.09)∗∗ (3.38)∗∗∗ (3.21)∗∗∗ (2.82)∗∗∗ (2.56)∗∗ SHARE −0.0346 −0.0351 −0.0416 (3.17)∗∗∗ (3.00)∗∗∗ (3.59)∗∗∗ # obs 993 993 947 947 897 897 R2 (%) 6.16 6.51 6.22 6.56 5.06 5.56 Panel B: Determinants of Bank-Level RET with RET-SHARE Interaction 1yr 1yr 2yrs 2yrs 3yrs 3yrs Past j years RET 0.0968 0.0919 0.2044 0.1920 0.1786 0.1566 (2.03)∗∗ (1.93)∗ (3.25)∗∗∗ (3.05)∗∗∗ (2.66)∗∗∗ (2.33)∗∗ SHARE −0.0346 −0.0347 −0.0421 (3.16)∗∗∗ (2.95)∗∗∗ (3.66)∗∗∗ RET X −0.0893 −0.0945 −0.6364 −0.3889 0.0280 0.4964 SHARE (0.11) (0.11) (0.46) (0.27) (0.02) (0.35) # obs 993 993 947 947 897 897 R2 (%) 6.16 6.52 6.24 6.57 5.06 5.57 RET (at 0.0924 0.0871 0.1726 0.1725 0.1800 0.1814 5% SHARE) (2.12)∗∗ (1.92)∗ (2.40)∗∗ (2.34)∗∗ (2.31)∗∗ (2.30)∗∗ Panel C: Determinants of Bank-Level RET with First-Half Interaction 1yr 1yr 2yrs 2yrs 3yrs 3yrs Past j years RET X 0.0406 0.0290 0.0998 0.0806 0.1222 0.0921 FIRST HALF (0.43) (0.31) (0.81) (0.64) (0.88) (0.64) RET 0.0826 0.0805 0.1666 0.1618 0.1404 0.1310 (1.49) (1.46) (2.19)∗∗ (2.14)∗∗ (1.87)∗ (1.76)∗ SHARE X −0.0256 −0.0219 −0.0326 FIRST HALF (1.05) (0.77) (1.16) SHARE −0.0244 −0.0265 −0.0304 (2.01)∗∗ (2.36)∗∗ (2.58)∗∗ # obs 993 993 947 947 897 897 R2 (%) 6.19 6.58 6.33 6.67 5.19 5.72 * significant at 10%; ** significant at 5%; *** significant at 1%. The dependent variable is RET, a bank��s average (–1, +1) CAR across all deals announced by the bank over a single calendar year. The explanatory variable RET is the average CAR over the past j calendar years, where j = {1,2,3}. SHARE is the bank��s market share, by value of deals, over the past j calendar years. To be included in the regression, a bank must have announced at least 2j deals over the relevant period. In Panel B, an interaction term between RET and SHARE is included. The marginal effect of RET at a SHARE of 5% is also included. In Panel C, FIRSTHALF equals 1 if the dependent variable is in 1994 or earlier, and 0 otherwise. The data are pooled across all banks, and regressions are estimated using year fixed effects, clustering standard errors by bank. The sample period is 1981–2007, and t-statistics are in parentheses.
skills have become commoditized over time, in part due to rapid increases in computer power and the rise in general skills resulting from MBA de- grees. If true, and if our results are driven primarily by the early part of our sample, our findings become less interesting because we are documenting a phenomenon that is no longer in existence. We investigate this hypothesis in 2306
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Panel C of Table 6 by adding an additional regressor, RET*FIRSTHALF, where FIRSTHALF is a dummy variable that equals 1 if the dependent variable is in the first half of our performance sample, 1981–1994. This is an appropriate cutoff, as Morrison and Wilhelm (2008, Figure 3) document a marked increase in computer power around 1993–1994. The interaction term is positive but insignificant in all six specifications. In the presence of the interaction term, the coefficient on RET now measures persistence using the second half of the dataset only. Importantly, even though we have less power, past RET remains significant in all specifications for j �� 2, suggesting that the ability differences we document remain significant today. 4. Do Clients Chase Performance? Even if a financial intermediary possesses superior ability (our fixed-effects re- sults of Table 3), Berk and Green (2004) show that this need not translate into performance persistence (our quantile and regression results, of Tables 5–6) if two necessary conditions hold: Clients chase past returns, and there are di- minishing returns to scale. Indeed, our persistence results for M&A advice contrast the lack of persistence in mutual funds (e.g., Carhart 1997). Since past RET positively predicts future RET, it would seem logical for clients to select banks based on past performance, in which case persistence will disappear if there are diminishing returns to scale. For ability to translate into persistence, it is sufficient for one of the necessary conditions present in Berk and Green (2004) to be violated. We investigate whether clients respond to past perfor- mance (the first condition) because it is much cleaner to measure than returns to scale. Table 7 studies the effect of past performance on a bank��s future market share. Since bank/client relationships take time to develop, large banks are likely to have persistently high market shares irrespective of past performance. We therefore include either a bank fixed effect or the bank��s past market share as a regressor. Standard errors are clustered at the bank level. Strikingly, even though RET is a positive predictor of future performance, the first six columns of Table 7 show that it is an insignificant determinant of market share (consistent with Rau 2000). By contrast, even though market share negatively predicts performance, it is strongly significantly related to future share. The last three columns of Table 7 repeat the analysis adding an interaction between RET and SHARE that is generally insignificant, suggesting that the results are similar across all sizes of banks.19 This lack of performance chasing is a sufficient condition for ability to translate into persistence and thus underpins the results of Section 3. The strong performance-chasing by
19 The coefficient on the interaction term in the one-year analysis is significant at the 10% level, but the marginal
effect of RET at a 5% market share is not significant.
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The Review of Financial Studies / v 24 n 7 2011 T able 7 Determinants of mark et share 1yr 2yrs 3yrs 1yr 2yr 3yrs 1yr 2yrs 3yrs Constant 0 .0099 0 .0068 0 .0059 0 .0098 0 .0068 0 .0059 (4 .55 )∗∗∗ (4 .08 )∗∗∗ (4 .64 )∗∗∗ (4 .59 )∗∗∗ (4 .07 )∗∗∗ (4 .65 )∗∗∗ RET 0 .0007 − 0 .0259 − 0 .0499 − 0 .0025 − 0 .0370 − 0 .0357 − 0 .0313 − 0 .0486 − 0 .0518 (0 .03 ) (0 .46 ) (0 .63 ) (0 .14 ) (1 .14 ) (1 .09 ) (1 .49 ) (1 .74 )∗ (1 .52 ) SHARE 0 .6805 0 .8118 0 .8489 0 .6830 0 .8114 0 .8480 (13 .09 )∗∗∗ (17 .67 )∗∗∗ (28 .08 )∗∗∗ (13 .26 )∗∗∗ (17 .70 )∗∗∗ (28 .82 )∗∗∗ RET X 3 .1504 1 .0993 1 .5095 SHARE (1 .80 )∗ (0 .30 ) (0 .42 ) Bank FE Y es Y es Y es No No No No No No # obs 1,168 1,079 1,017 1,168 1,079 1,017 1,168 1,079 1,017 R
2
(%) 59 .72 60 .69 61 .14 44 .53 54 .83 58 .23 44 .81 54 .85 58 .24 RET (at 0 .1262 0 .0063 0 .0237 5% SHARE) (1 .60 ) (0 .04 ) (0 .15 ) * significant at 10%; ** significant at 5%; *** significant at 1%. The dependent v ariable is a bank��s mark et share, by v alue of deals, in one particular year. RET is the bank��s av erage (–1, +1) CAR o v er the past j calendar years, where j ={1,2,3 }. SHARE is the bank��s mark et share, by v alue of deals, o v er the past j calendar years. The data are pooled across all banks, and regressions are estimated using bank fix ed effects and clustering standard errors at the bank lev el. T o be included in the results, a bank must hav e announced at least 2 j deals o v er the period used to estimate RET and SHARE . The sample period is 1981–2007, and t-statistics are in parentheses.
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mutual fund investors, and the absence of such behavior by acquirers, may thus reconcile the performance persistence in M&A with its absence in mutual funds. Our results echo those of Kaplan and Schoar (2005), who document persistence in private equity performance (also found by Axelson et al. 2010) and that top-performing funds grow proportionally more slowly than poorly performing funds, which explains why persistence can arise. The significance of past market share, even though it negatively predicts fu- ture performance, and irrelevance of past returns, even though it is a positive predictor, may appear at first glance to be inefficient. However, applying the
Berk and Green (2004) framework to our setting, responsiveness to perfor-
mance requires not only learning about ability from past returns but also competitive provision of M&A mandates by clients. Thus, lack of performance- chasing can occur if either assumption is violated, and therefore has different interpretations. Mandate awards may be noncompetitive if clients build up relationship-specific capital with a particular bank, which can be leveraged by continuing to use it for future deals. Thus, while past market share is a predictor of future mandates, it may not be a determinant (i.e., actively used by acquirers in their selection decisions)—instead, its significance arises as it proxies for the extent of existing relationships. Table 8 investigates the relationship-specific capital hypothesis by studying repeat acquirers, who have conducted at least one acquisition in the prior five years while being public. Using a previously mandated advisor is associated with a lower CAR of 0.24 percentage points (t-statistic of 1.74). If the advisor had generated a nega- tive average CAR for that particular client in question, the CAR is 0.79 per- centage points lower than using past advisors that generated positive CARs
Table 8 Relationship between a deal��s (–1, +1) CAR and the use of a past advisor Panel A (1) (2) (3) (4) Retained only Retained an old advisors with advisor with Did not retain Retained an positive past negative past and old advisor old advisor performance performance CAR 0.33% 0.09% 0.46% −0.33% t-statistic (3.28)∗∗∗ (0.89) (3.37)∗∗∗ (2.41)∗∗ # obs 3,793 4,291 2,263 2,028 Panel B (1)–(2) (1)–(3) (1)–(4) (3)–(4) CAR 0.24% −0.13% 0.66% 0.79% t-statistic (1.74)∗ (0.78) (3.89)∗∗∗ (4.09)∗∗∗ * significant at 10%; ** significant at 5%; *** significant at 1%. Deals in which the acquirer has made an acquisition in the last five years while public are considered. Panel A divides deals into groups according to whether the acquirer retained an advisor from a past transaction. Panel B examines the differences in average CAR between groups. The sample period is 1985–2007, and t-statistics are in parentheses.
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The Review of Financial Studies / v 24 n 7 2011
(t-statistic of 4.09). These findings are consistent with the RET persistence re- sults of Section 3. However, even if using a past advisor is correlated with poor future M&A performance, it is not inefficient if the bank provides many other non-M&A services. For example, Yasuda (2005) and Ljungqvist, Marston, and
Wilhelm (2006, 2009) find that clients select underwriters based on lending re-
lationships or analyst coverage. Even if banks are providing no other services, clients may be ��locked in�� to a past advisor because working on a previous deal gives the advisor an information monopoly: Ljungqvist, Marston, and Wilhelm (2006, 2009) find that past co- or lead underwriters are typically appointed as future lead underwriters. Asker and Ljungqvist (2010) highlight another source of lock-in: clients�� wish to avoid sharing banks with product market rivals. Learning about ability from past returns will not occur if clients are unaware of the persistence of RET and mistakenly believe that market share is a good measure of quality. Under this interpretation, market share is significant not because clients are locked in and it proxies for existing relationships, but be- cause clients actively select on it. Indeed, both the insignificance of RET and the significance of SHARE are fully consistent with real-life practices in the investment-banking industry, where league tables on market share are widely publicized and used as proxies for expertise. Therefore, industry participants have grown to take it for granted that market share equates with quality; sim- ilarly, many academic studies, such as those by Rau (2000), Kale, Kini, and
Ryan (2003), and Hunter and Jagtiani (2003), use market share as their mea-
sure of quality. However, we show that it is actually negatively correlated with performance. However, returns are less publicized, which hinders clients�� abil- ity to use returns as an additional metric to learn about quality even if they are not locked in. Given the lack of performance-chasing and the importance of prior relation- ships, it is logical for banks to accept even bad deals. Not only will the mandate boost fee income today, but it will also create new relationships and thus the ability to win future mandates. Even though accepting bad deals will depress RET, this does not reduce future mandates, due to either lock-in or failure to learn. Indeed, if certain banks are systematically nonselective and accept bad deals, this would lead to the negative correlation between market share and RET that we document. As a preliminary investigation into the importance of lock-in for M&A, we study the extent to which clients switch M&A advisors. During 1985–2007, we find that on 21.4% of M&A deals where the acquirer had made at least one acquisition in the past five years, the advisors also covered all of the acquirer��s M&A deals over the past five years. On 50.1% of deals, the advisors covered at least one deal over the past five years. Ljungqvist and
Wilhelm (2005) find that 64.1% of equity issuers used the same underwriter
for their initial public offering and first seasoned equity offering. Our fig- ures are somewhat lower, tentatively suggesting that lock-in may be slightly weaker in M&A. One potential explanation is that a significant amount of 2310
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M&A advice is target specific, and thus knowledge built up during one trans- action may be less applicable for future deals. Our findings are consistent with
Francis, Hasan, and Sun (2009), who find that fewer than 20% of deals are
conducted by exclusive advisors, which they interpret as evidence that exist- ing relationships are less influential for advisor choice in M&A than in other banking services. Given the importance of other banking services, we also gather SDC is- suance data to measure debt and equity underwriting relationships, and Dealscan data to measure lending relationships.20 Since Dealscan data are sparse until 1988, the lending numbers are calculated from 1993; for consis- tency, we use the same timeframe for the issuance data. We find that on 9.5% of M&A deals over 1993–2007, the advisors also covered all of the acquirer��s issuance over the past five years; on 45.4% of deals, banks covered at least one issue. For lending, the numbers drop to 6.7% and 15.8%, respectively. While we do not have a benchmark for a formal test, these relatively low num- bers suggest that clients use different advisors for their M&A business than for their underwriting and borrowing decisions. Indeed, while there are plau- sibly strong synergies between lending and bond underwriting (Yasuda 2005), the connection between lending and M&A advice appears to be weaker. We also attempt to study the selection decisions of clients who are not locked in and thus have freedom on advisor choice. We take a subsample of clients that engaged in no M&A deals, issuance, or borrowing with our 143 advisors in the past five years, and thus have no existing relationships. We note that this is not a perfect measure of freedom. First, it may include ��unfree�� clients: If some of these acquirers intend to engage in issuance or borrowing in the future, they may select an M&A advisor who provides these services. Second, it may exclude ��free�� clients: Some clients may have engaged in issuance or borrow- ing with an M&A advisor in the past but have no intention of doing so in the foreseeable future. If lock-in, rather than failure to learn, is the reason for the aggregate insignificance of RET for market share in Table 7, then RET should be positively correlated with future mandates for ��free�� acquirers. By contrast, Table 9 shows that RET is negatively correlated with future mandates in most specifications (though statistically insignificant) for this subsample. This sug- gests that failure to learn is a cause of the insignificance of past performance. However, we note that these results are only suggestive, due to the difficulty of identifying ��free�� clients, mentioned earlier, and our small sample size: We can identify ��free�� clients from only 1993 onward, and only 1,224 deals were conducted by such acquirers. If clients are indeed not fully locked in, our results suggest that they may be able to improve their advisor-selection decisions by using information on
20 We thank Michael Roberts for providing us with the table to link Dealscan to Compustat, used in Chava and
Roberts (2008).
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The Review of Financial Studies / v 24 n 7 2011 Table 9 Logit regressions of advisor choice by acquirers without a previous M&A, issuance, or lending relation- ship over the last five years with one of our 143 banks 1yr 1yr 2yrs 2yrs 3yrs 3yrs Past j years Constant −4.4597 −4.5908 −4.4199 −4.5797 −4.4063 −4.5746 (168.92)∗∗∗ (144.50)∗∗∗ (154.13)∗∗∗ (125.75)∗∗∗ (141.84)∗∗∗ (114.91)∗∗∗ RET −0.3508 0.0059 −2.2120 −1.3952 −2.5828 −1.1367 (0.37) (0.01) (1.56) (0.90) (1.47) (0.59) SHARE 4.0123 4.5560 4.5307 (8.26)∗∗∗ (8.40)∗∗∗ (8.07)∗∗∗ # obs 67,806 67,806 63,586 63,586 60,791 60,791 Pseudo R2 (%) 0.00 0.56 0.02 0.68 0.02 0.66 * significant at 10%; ** significant at 5%; *** significant at 1%. The analysis is at the acquirer-bank level, and the dependent variable equals 1 if the acquirer mandated that particular bank. Banks that do not advise on M&A in the year of the deal are treated as unavailable and excluded from the analysis. RET is the bank��s average (–1, +1) CAR over the past j calendar years. SHARE is the bank��s market share by value of deals over the past j calendar years. Standard errors are clustered by deal. The sample period is 1993–2007, and t-statistics are in parentheses.
past returns to supplement market-share information.21 An increased focus on returns may in turn dissuade banks from accepting value-destructive transac- tions. The findings also have implications for the nature of contracts between acquirers and advisors. McLaughlin (1990) finds that banks are paid for deal completion but not value creation. He suggests that reputational concerns may be sufficient to align banks with clients, since a bank that performs poorly will not win future mandates. However, the insignificance of RET implies that banks�� implicit incentives are also low—if clients are locked in or do not learn from RET, banks can be less concerned with returns—and so explicit incen- tives would be valuable. In a similar vein, clients frequently solicit fairness opinions to verify that the terms are ��fair�� (Kisgen, Qian, and Song 2009). As part of its mandate, an advisor should ensure that the client is undertaking only favorable deals in the first place, and there should be no need for a separate fairness opinion. The prevalence of such opinions is consistent with the view that incentives to act in clients�� interests are insufficient, due to either lock-in or failure to learn. 5. Conclusion This article finds a significant investment-bank fixed effect in the announce- ment returns to an acquisition. The positive association between certain banks and high returns can be predicted by clients using past performance—a bank��s
21 Since shareholders are diversified, they are not concerned with idiosyncratic risk. Thus, a value-maximizing
manager should maximize expected CAR rather than a measure adjusted for the variance. However, managers may care about the variance of CAR if they are concerned with their own undiversified human capital. We find that a bank��s RET is highly correlated with its ��Sharpe ratio�� of RET divided by the standard deviation of CAR. Over three-year non-overlapping windows beginning in 1981, we find the average Spearman rank correlation of RET and the Sharpe ratio to be 0.965.
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returns are persistent. While most prior research attributes the entire CAR to the advisor, we remove the component that can be explained by acquirer char- acteristics; the orthogonal component remains persistent. These results suggest that certain banks have ability in identifying acquisitions or negotiating terms, or trustworthiness in turning down bad deals. They contrast prior findings that bank quality, as measured by market share or reputation, has no positive ef- fect on M&A outcomes, thus suggesting that banks do not matter. The lack of performance-chasing by clients potentially explains why persistence exists in M&A advice even though it is absent in mutual fund performance. This article suggests a number of questions for future research. First, it may be interesting to study why clients do not chase performance and whether such behavior is efficient—i.e., disentangle whether it results from rational lock-in or inefficient failure to learn (e.g., due to an acquirer��s private benefits from working with a prestigious bank). Agency variables such as governance may explain advisor choice, just as they do for acquirer returns (Masulis, Wang,
and Xie 2007). Second, the low returns to ability appear puzzling. While su-
perior performance in equity underwriting is rewarded with higher future mar- ket share (Dunbar 2000; Hoberg 2007), and superior returns to private equity investing increases future fund flows (Chung et al. 2010), there seems to be little reward for good M&A advice. Third, we have focused on acquirer returns, because these are frequently negative, and so advisor selection is particularly important for bidders to ensure value creation; in addition, far more bidders are public than are targets. It would be fruitful to investigate whether target returns also exhibit a bank fixed effect, and whether the banks that consistently create value for bidders are also skilled at defense mandates.
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