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Journal of Accounting and Economics 45 (2008) 94–115

www.elsevier.com/locate/jae

What’s in a vote? The short- and long-run impact ofdual-class equity on IPO firm values$

Scott B. Smarta, Ramabhadran S. Thirumalaib, Chad J. Zutterc,�

aKelley School of Business, Indiana University, Bloomington, IN 47405, USAbIndian School of Business, Gachibowli, Hyderabad 500 032, India

cKatz Graduate School of Business, University of Pittsburgh, Pittsburgh, PA 15260, USA

Received 26 September 2006; received in revised form 12 June 2007; accepted 16 July 2007

Available online 22 July 2007

Abstract

We find that relative to fundamentals, dual-class firms trade at lower prices than do single-class firms, both at the IPO

and for at least the subsequent 5 years. The lower prices attached to duals do not foreshadow abnormally low stock or

accounting returns. Moreover, some types of CEO turnover are less frequent among duals, and in general CEO turnover is

sensitive to firm performance for singles but not for duals. Finally, when duals unify their share classes, statistically and

economically significant value gains occur. Collectively, our results suggest that the governance associated with dual-class

equity influences the pricing of duals.

r 2007 Elsevier B.V. All rights reserved.

JEL classification: G12; G14; G30; G32; G34

Keywords: Initial public offerings; Dual class; Ownership structure; Governance; Firm value; CEO turnover

1. Introduction

Prompted by the wave of corporate scandals around the turn of the century, Congress passed theSarbanes–Oxley (SOX) Act of 2002. This legislation, ostensibly designed to protect investors, containsprovisions which presuppose that firms’ governance practices affect shareholder value. Several research paperspublished in the wake of SOX look for possible connections between governance and value, but noprofessional consensus exists regarding whether or how governance and value are linked. In this paper, wecontribute to the debate by comparing several attributes of dual- and single-class firms, following them fromtheir IPO dates forward for 5 years. We focus on IPO firms because it is prior to the IPO that firms establish

e front matter r 2007 Elsevier B.V. All rights reserved.

cceco.2007.07.002

le comments, we thank Utpal Bhattacharya, Amy Dittmar, Robert Dittmar, Craig Holden, Sreenivas Kamma, William

derik Schlingemann, Shawn Thomas, Charles Trzcinka, Greg Udell, and presentation participants at the Haub School of

Joseph’s University, the Kelley School of Business Indiana University, and The Wharton School University of

e also thank Anil Shivdasani (the Referee) and S.P. Kothari (the Editor) for suggestions that greatly improved the paper.

ing author. Tel.: +1 412 648 2159; fax: +1 412 648 1693.

ess: czutter@pitt.edu (C.J. Zutter).

ARTICLE IN PRESSS.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115 95

the governance rules by which they will abide as public companies. Subsequent changes to governancepolicies, such as the adoption of poison pills, may be precipitated by imminent takeover threats, making itmore difficult to disentangle any relation between governance provisions and firm value. Also, by focusingexclusively on newly public dual-class firms, we avoid issues that arise when firms switch from single-class todual-class status through a recapitalization plan, which may be tied to other simultaneous value-relevantevents.1 Our research design emphasizes comparisons between dual-class and single-class firms because dual-class equity is a particularly effective method by which managers may entrench themselves. By separating cashflow rights from voting rights, dual-class equity enables corporate insiders to exert voting control over a firmin which they may hold a comparatively small economic interest. Hence, dual-class shares potentiallyexacerbate problems associated with the separation of ownership and control.

The importance of dual-class equity in U.S. markets appears to be rising as new firms adopt this particularownership structure. Smart and Zutter (2003) report that from January 1990 to May 1994 about 7 percent ofall U.S. IPOs adopt dual-class equity structures and that these firms account for about 11 percent of theaggregate market capitalization of IPO firms. However, from June 1994 to October 1998, these figures increasesignificantly. In the latter period, almost 12 percent of all IPOs were dual-class deals accounting for about 31percent of IPO market capitalization.

The Finish Line Inc. provides a prototypical example of a dual-class IPO. Holders of the Finish Line’s ClassA shares, originally offered to the public through a 1992 IPO, receive one vote per share, while the FinishLine’s Class B shareholders control ten votes per share. All other rights of the two share classes are identical.After its IPO, The Finish Line sold seasoned equity (Class A shares only) on three occasions. With eachsubsequent equity sale, insider ownership steadily decreased, but the Class B shares enabled insiders to retainvoting control. Recent Securities and Exchange Commission (SEC) filings reveal that The Finish Line’s threefounders collectively own just 205,934 of the firm’s 21,208,451 Class A shares, yet they own all 2,865,284 ClassB shares outstanding. Combining their holdings of both classes, the founders’ claim on the firm’s cash flowsbarely reaches 12.8 percent, but they still control almost 57.9 percent of the votes that can be cast on anymatter requiring shareholder approval. Does the wedge between insiders’ voting rights and cash-flow claimsthat dual-class equity facilitates influence the pricing of dual-class equity at the IPO date and beyond?2

Some of the largest and most vocal institutional investors subscribe to the view that dual-class stockarrangements entrench managers at shareholders’ expense. For example, in its proxy voting guidelines, theCalifornia Public Employees’ Retirement System (CalPERS) says that it votes against any proposal to createunequal voting rights across share classes. In response to Google’s announcement in 2004 that it would gopublic with dual-class equity, a high-ranking official at TIAA-CREF remarked that Google’s shares should bepriced at ‘‘a substantial discount’’ and that the dual-class stock ‘‘effectively disenfranchises outsideshareholders.’’3 More recently, the hedge fund Clinton Group, which owns just over 5 percent of the FinishLine’s Class A shares, wrote a public letter asking the firm to create shareholder value by eliminating its dual-class ownership structure, repurchasing shares, or going private. Determining whether investors apply adiscount to the shares of dual-class firms or more generally to the shares of firms that adopt governanceprovisions that strengthen the position of corporate insiders remains a challenging empirical question.

There are several channels through which dual-class governance structures could influence firm values. Forexample, Francis et al. (2005) find that the credibility of earnings information released by dual-class firms isinferior to that provided by single-class firms. Perhaps as a consequence, they also find that the salience ofdividends as a performance measure is greater in dual-class firms. In a study of firms in East Asian countries,Fan and Wong (2002) find that earnings informativeness declines as the gap between insiders’ cash flow rightsand voting rights widens.

1Current NYSE and NASDAQ listing requirements make it exceedingly difficult for existing public companies with a single share class

to recapitalize with multiple share classes. As a practical matter, therefore, it is only at the IPO stage when a firm can enact a dual-class

structure.2DeAngelo and DeAngelo (1985) report that insiders control a median of 56.9 percent of the votes while holding just 24.0 percent of the

cash flow rights in dual-class firms.3See London (2004). Some high-profile regulators also subscribe to this view. Charlie McCreevy, the internal market commissioner of

the European Union, said in a Financial Times interview, ‘‘It is my goal to get the one-share, one-vote principle accepted across the 25

member states.’’ See Buck (2005).

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Exploiting a unique feature of the banking industry which gives managers the authority to vote shares heldin trust, Adams and Santos (2006) explore the performance consequences of the wedge between managers’cash flow rights and voting rights. Using both Tobin’s Q and return on assets as operating performancemeasures, they find that performance is positively related to managerial voting control for small stakes, butcontrol has negative consequences for performance when managers control a large fraction of votes.

In a recent influential paper, Gompers et al. (2003) measure the protections afforded shareholders throughfirms’ corporate governance structures. They construct a corporate governance index (dubbed simply, G) andfind that firms with poor governance index scores underperform the market and have lower Tobin’s Q valuesthan firms with better governance scores.4 However, Core et al. (2006) find no evidence that poorly governedfirms as defined by Gompers et al. achieve unexpectedly poor operating results.5 Furthermore, they claim thatthe sub-par stock performance of badly-governed firms eventually reverses. Core et al. conclude that theirevidence does not support the hypothesis that bad governance leads to low stock returns and they propose thatthe anomalous returns reported in Gompers et al. may simply be a manifestation of the ‘‘new economy pricingpuzzle of the (late) 1990s.’’

In this paper, we conduct a variety of tests to build a case that dual-class equity is associated with lowershare prices. Our findings are consistent with Core et al. in the sense that although we find that dual-class firmstrade at relatively low price levels, returns on dual-class shares do not fall below standard benchmarks. Inother words, the market efficiently prices dual-class companies. In Section 2 we provide a literature review. InSection 3 we describe our data and provide descriptive statistics. In Section 4 we compare the market prices ofdual- and single-class companies, starting at the IPO date and continuing for the next 5 years. The data showthat firms choosing voting-right structures that favor management face a significant and persistent valuationdiscount in the market, even after controlling for the endogenous choice to go public with dual-class equity.This valuation gap persists while controlling for cross-sectional differences in industry valuation multiples aswell as firm-specific attributes including growth, profitability, and leverage. In this regard, our results alignwith those of Gompers et al. (2003) and other studies reporting a valuation discount for firms with governancemechanisms which favor insiders over shareholders.

In contrast to the results in Gompers et al. (2003), we find no connection between governance andmispricing. In Section 5 we evaluate the long-run stock returns realized by dual- and single-class IPOs and findno evidence that dual-class firms underperform. This result is obtained using Fama–French–Carhart four-factor pricing regressions. By this standard, the market’s valuation of dual-class IPOs is efficient.

If the difference in valuation multiples between duals and singles does not reflect a pricing error, what doesit reflect? One possibility is that poorly performing firms adopt governance systems to protect incumbents. Inthis case, we expect to see low operating performance from poorly governed firms. We also examine thispossibility in Section 5, reporting the results from tests for differences in operating performance betweensingle- and dual-class firms. As with stock returns, there is at best only scant evidence suggesting that dualsexhibit abnormally low operating performance. We acknowledge, however, that operating performancemeasures are notoriously noisy and can lead to tests with insufficient power to reject the null.

Although singles and duals generate similar accounting returns over time, a valuation gap between singlesand duals could still be tied to operating performance if dual-class firms are riskier than singles. That is,holding expected cash flows constant, dual-class firms would trade at lower multiples if they carry moresystematic risk than singles. The asset pricing literature does not give us a universally accepted method forevaluating risk differences between portfolios of stocks. Nevertheless, we can offer one observation suggesting

4Gompers et al. (2003) calculate G by simply adding up the number of provisions that reduce shareholders’ rights. Therefore, a higher G

corresponds to worse corporate governance. Whether each of the component provisions of G has a meaningful impact on either the degree

to which managers are entrenched or on firm value is an open question. Comment and Schwert (1995), for example, argue that poison pills

do very little to deter takeovers and have negligible wealth effects in most cases. It is interesting to note that Gompers et al. (2006) report

that dual-class firms on average have lower G index scores than do single-class firms. Presumably this is because dual-class equity is a

rather extreme form of entrenchment and firms adopting this structure do not need to adopt many other provisions to protect insiders’

interests.5Eldenburg and Krishnan (2003) examine the operating performance results of private versus public hospitals. They find that public

hospitals underpay chief executives and display weaker operating results compared to private hospitals. They conclude that weak

governance leads directly to lower performance.

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that the valuation discount applied to duals does not reflect a higher systematic risk. In Fama–French stylepricing regressions, we observe no systematic tendency for the factor betas for dual-class firms to exceed thoseof singles. This fact leads us to the conclusion that dual-class firms are no more risky than their single-classcounterparts.

Lacking evidence linking differences in operating performance to the valuation gap between singlesand duals, we turn to a governance explanation. Section 6 examines differences in CEO turnover events forsingles and duals. We first group all turnover events into two categories. External turnover refers to anevent when the CEO departs because the firm is acquired. All other cases of turnover we designate asinternal turnover. In the sample that includes both types of turnover events three interesting and suggestivefindings emerge. First, the incidence of CEO turnover is slightly lower for duals than for singles, consistentwith the hypothesis that dual-class shares entrench incumbents. Second, poor accounting performanceprecedes turnover events for single-class firms but not for duals. Third, stock returns surrounding turnoverevents vary between singles and duals. For single-class firms, negative abnormal returns precede internalturnover events, but not external turnover events. This suggests that internal governance mechanismsfor single-class firms work well enough to remove the CEO of an underperforming firm without interventionfrom the market for corporate control. However, for dual-class firms the pattern reverses. Negative abnormalreturns precede external turnover events, but no correlation exists between internal turnover and priorreturns.

We refine our turnover analysis by focusing exclusively on the set of internal events and classifying them aseither forced or unforced. As in the larger sample, negative accounting performance precedes turnover events,both forced and unforced, for single-class firms, but not for dual-class firms. Collectively, these results suggestthat the dual-class structure substantially weakens the link between an executive’s tenure on the job and theperformance of the company.

Our final test, presented in Section 7, focuses on firms that unwind their dual-class voting structures. Wetrack our sample of dual-class firms during the 5 years following their IPOs and find that 37 eventually unifytheir share classes. Conducting an event study around the effective date of these unifications, we find positiveabnormal returns on the order of 5–6 percent using either the market model or Fama–French model. Becauseunifications often occur gradually as insiders divest their Class B shares, and because firms typically make noformal announcement regarding their plans to unify shares, our event study estimates likely understate thetrue value increase associated with share unifications.

2. Literature on governance and firm value

Empirical research on the link between governance and firm value has a long history. Much of the researchin this area uses event-study methods to determine the short-term impact of changes in firms’ governancepractices on share prices. For example, dozens of studies use these methods to study the wealth effects of theadoption of anti-takeover amendments, the passage of state anti-takeover laws, and changes in thecomposition of corporate boards.6

Some of the early work on dual-class firms fills a niche in this literature. While our paper focuses on firmsthat adopt a dual-class structure at the IPO, some firms that currently have dual-class voting arrangementscreated them in a recapitalization transaction. Partch (1987), Jarrell and Poulsen (1988), and Millon-Cornettand Vetsuypens (1989) all offer event study evidence on dual-class recapitalizations, but they reach noconsensus on whether recapitalizations help, harm, or have no impact on shareholder wealth. One reasonableconclusion emerging from these papers is that a sample of firms that recapitalize from a single-class to a dual-class structure does not provide ideal conditions for an analysis of the effects of dual-class equity on firmvalue. For example, takeover activity concomitant with the recapitalization clouds the interpretation of theannouncement effect in the market. This ambiguity in the empirical evidence mirrors conflicting predictionsfrom theory. For example some authors conjecture that dual-class voting arrangements are less than optimal,

6For event-study evidence, see DeAngelo and Rice (1983), Linn and McConnell (1983), Jarrell and Poulsen (1987), McWilliams (1990),

Cotter et al. (1997), Shivdasani and Yermack (1999), Fich and Shivdasani (2005), Perry and Shivdasani (2005), Choi et al. (2007), Dahya

and McConnell (2007), Faleye (2007), Paul (2007), and Shivdasani (2006).

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while others argue that they can serve to maximize firm value by solving various types of underinvestmentproblems.7

Cox and Roden (2002) examine the relative prices of high-vote and low-vote shares in U.S. dual-class firms.They find that the voting premium on high-vote shares can be reduced when holders of low-vote shares receivehigher dividend payments. However, for the vast majority of dual-class firms in the U.S. the cash flow rightsare identical for the two (or more) share classes and only one share class trades in the open market. This isalways the case for firms that go public with dual-class shares. Furthermore, evidence of a voting premiumbetween the two publicly traded share classes speaks to the distribution of firm value between differentshareholder groups, but it does not imply that dual-class structures reduce firm value on the whole.

Though we supplement our primary results with event-study evidence, our main emphasis is on the long-runimpact of dual-class stock on valuation ratios such as the earnings-to-price (E/P) and EBITDA-to-price ratios.Our approach aligns closely with that of Yermack (1996), Gompers et al. (2003), Bebchuk and Cohen (2005),and Gompers et al. (2006). In a study designed to assess the relation between the size and effectiveness ofcorporate boards, Yermack reports that firms with smaller boards have higher Q values in the period1984–1991. Yermack also finds that smaller boards are more likely to dismiss CEOs following a period of poorperformance and that the sensitivity of CEO turnover to performance is greater among firms with smallboards. We report similar results when we compare circumstances surrounding CEO turnover among dual-class versus single-class firms.

Bebchuk and Cohen (2005) focus on firms with staggered boards, arguing that a staggered board providesincumbents with a high degree of insulation from outside monitoring. They find that firms with staggeredboards have significantly lower Tobin’s Q values. Bebchuk and Cohen argue that staggered boards represent,‘‘ythe key arrangement that protects incumbents from removal in U.S. publicly traded companies.’’However, they exclude dual-class firms from their analysis, commenting that, ‘‘yin such firms, the holding ofsuperior voting rights is likely to be the key for entrenching incumbents.’’

In a similar vein, Gompers et al. (2006) examine the relation between firm value, cash flow rights, and votingrights.8 They acknowledge that financial economists widely agree that insider stock ownership can have bothpositive and negative effects on firm value.9 Their most interesting result is that the value of dual-class firmsdepends positively on insiders’ cash flow rights and negatively on insiders’ voting rights.10

Our analysis differs from that in extant literature in several important ways. First, Gompers et al. (2003) aswell as Bebchuk and Cohen (2005) exclude dual-class firms from their analysis and focus instead on cross-sectional variation in governance provisions among single-class firms. Although dual-class firms constitute lessthan 10 percent of the universe of firms covered by the Investor Responsibility Research Center (IRRC), theprimary source of data on firm-specific governance characteristics, in the late 1990s the percentage of newfirms coming to market with dual-class equity was greater than 10 percent. Because they are typically largerthan single-class IPOs, dual-class companies account for a larger percentage of aggregate equity value thantheir numbers alone indicate.

Second, while our paper offers evidence on the relation between dual-class equity and firm value in the spiritof Gompers et al. (2006), our analysis departs from theirs in several interesting and important ways. Theirsample includes a broader cross-section of dual-class firms, but our analysis uses a longer time series.Moreover, their sample includes much older and larger dual-class firms than the IPO firms which we study.11

7See DeAngelo and DeAngelo (1985), Fischel (1987), Ruback (1988), Grossman and Hart (1988), Harris and Raviv (1988), and Denis

and Denis (1994) for very different theoretical perspectives on the merits of dual-class equity.8See Villalonga and Amit (2006) for a similar analysis of large, family-owned U.S. corporations.9See Stulz (1988) for theoretical work on the offsetting effects of managerial ownership. Morck et al. (1988) and McConnell and Servaes

(1990) offer empirical evidence that higher insider ownership at first increases, then decreases firm value.10Evidence that the separation of cash flow and voting rights leads to lower firm value is not restricted to U.S. markets. In a study of

emerging market firms, Lins (2003) finds lower firm value when managers’ voting ownership exceeds their cash flow ownership. Similarly,

in a study of firms in Asian countries, Claessens et al. (2002) report lower firm value when the largest shareholder’s voting ownership

exceeds their cash flow ownership. The legal environments and the protections they offer investors vary dramatically across countries, so it

is not clear whether the results of these studies can be extended to the United States.11They report a mean (median) age, as of 2000, for dual-class firms in their sample of 12.87 (7.21) years. All of the dual-class firms in our

sample went public between 1990 and 1998, so by the end of 2000, the maximum age of a dual-class firm in our study is just 11 years. The

definition of age here is years since the CRSP listing date.

ARTICLE IN PRESSS.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115 99

This difference in sampling choices allows us to address an interesting question—does the market apply adiscount to dual-class IPOs from their inception, or does the discount emerge gradually over time?Immediately after the IPO, it is common for insiders to own a large fraction of the outstanding shares in bothsingle-class and dual-class companies. However, as firms grow and return to the market to raise equitythrough seasoned offerings, the voting power of single-class insiders’ declines at the same rate as their cashflow rights, while dual-class insiders’ voting rights change at a much slower rate than their economicownership does. Megginson et al. (2007) report that more than 40 percent of IPO firms in the 1990s issuedseasoned equity within 5 years of the IPO date, and the probability of an IPO firm conducting a seasonedoffering is much higher if that firm is a dual-class company. Therefore, over time the gap between theeconomic incentives and voting power of dual-class firms widens. Our evidence suggests that investorsdiscount dual-class equity starting at the IPO date and that the discount is persistent for at least 5 yearsfollowing the IPO.

Third, we also provide two important previously undocumented pieces of evidence supporting thehypothesis that investors discount dual-class firms. Similar to the evidence on large versus small corporateboards presented by Yermack (1996), we find that CEO turnover in single-class firms follows a period of sub-par accounting performance, a pattern that dual-class turnover events do not mimic. Intuitively, one mightguess that the valuation discount we document exists because investors expect dual-class managers to be firmlyentrenched. Using a broad definition of CEO turnover, we find that turnover events are slightly less commonamong dual-class firms compared to single-class companies. We also show that the sign and magnitude ofabnormal returns surrounding turnover events depend on the type of turnover (e.g., acquisition related or not)and whether the firm is a single or a dual.

3. Data

Our primary data source is the Disclosure New Issues database from Disclosure Inc. The data set provides awealth of information about debt and equity offerings on an issue-by-issue basis for any original orsubsequent registration or prospectus filed with the SEC. Compared to another well-known IPO dataprovider, Securities Data Corp., Disclosure under-samples small IPOs, though it does not exclude them. Typesof firms excluded from our data set include closed-end funds, unit offers, investment companies, real-estateinvestment trusts, and limited partnerships.

We extract issues from Disclosure by selecting records for firm-commitment IPOs of common stock from1990 through 1998. We stop collecting data in 1998 because we want to track each IPO firm for several yearsafter the IPO date. The search yields 3,628 issues. We eliminate duplicate records, reducing our sample to2,787 issues. We further eliminate 165 issues because we cannot match data with Standard and Poors’COMPUSTAT. Our final sample covers 2,622 IPOs (including 253 dual-class issues) with offer prices rangingfrom $5 to $35.

Table 1 presents some descriptive statistics for our sample. Dual-class firms raise more money at the IPOand have a higher average market capitalization than single-class firms. Duals are less likely to use venturefinancing, but more likely to employ a high-reputation investment bank. Dual-class IPOs are significantlymore likely to be equity carveouts or firms that anticipate paying dividends than singles, but not reverse LBOsor firms with anti-takeover provisions. Finally, duals list less often on NASDAQ, list fewer uses for their IPOproceeds, and have higher institutional ownership following the IPO than do single-class IPO firms. In thenext section we control for these characteristics when examining the market’s pricing of duals and singles.

4. Market pricing of dual- and single-class IPOs

Our primary interest is comparing the market values of singles and duals. In this section we examine themarket’s pricing of dual- and single-class IPOs by testing for systematic differences in valuation ratios acrossthe different firm types. We offer two measures: the E/P ratio and the EBITDA-to-price ratio.12 If the market

12In both ratios a measure of market value appears in the denominator, so a higher market value, relative to fundamentals, results in a

lower ratio.

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Table 1

Descriptive statistics

Means

Dual class Single class P-value

Number 253 2,369

Offer value in millions $130.73 $63.06 0.0001***

Market capitalization in millions $739.35 $253.49 0.0001***

Venture-backed 0.162 0.398 0.0001***

High-reputation I-bank 0.712 0.496 0.0001***

Equity-carveout 0.127 0.083 0.0203**

Reverse-LBO 0.055 0.040 0.2487

Anti-takeover provisions 0.937 0.933 0.7952

No dividends anticipated 0.826 0.881 0.0127**

NASDAQ-listed 0.640 0.844 0.0001***

Number of uses of proceeds 2.810 3.149 0.0001***

Lagged market return 0.013 0.013 0.9693

Fraction of institutional ownership 0.282 0.198 0.0001***

Number of institutional owners 30.232 22.085 0.0001***

This table presents sample means by offer type. Offer value is the total number of shares offered times the CPI-adjusted final offer price.

Market capitalization is the total number of shares outstanding after the offering times the CPI-adjusted first-day closing price. Indicator

variables are set equal to one, respectively, for venture-backed, high-reputation I-bank, equity-carveout, reverse-LBO, anti-takeover

provisions, and NASDAQ-listed deals. Number of uses of proceeds is the number of uses of proceeds listed in the final prospectus. Lagged

market return is the compounded daily CRSP value-weighted return over the 22 trading days preceding the initial public offering.

Institutional ownership is end-of-quarter 13f institutional ownership of publicly traded shares for the quarter in which the IPO took place.

P-values refer to tests of equal means across offer types. Respectively, ***, **, and * denote significant difference at 1, 5, and 10 percent.

S.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115100

imposes a cost on companies that insulate managers through dual-class ownership, then we expect to find thatthese companies trade at lower prices, relative to earnings or EBITDA, than firms with the more typical one-share, one-vote structure and we expect this valuation discount to persist.

4.1. Univariate analysis

We begin by calculating the E/P and EBITDA-to-price ratios. In the ratio using earnings, we divide by pricebecause the more conventional approach of putting earnings in the denominator results in a highly skeweddistribution.13 For example, many IPO firms have very low earnings, so the distribution of price-to-earnings(P/E) ratios has a very long right tail. For consistency in reporting and interpretation, we divide EBITDA byprice. If the market believes that dual-class equity entrenches insiders to the detriment of outside shareholdersthen dual-class firms should have higher E/P and EBITDA-to-price ratios relative to singles.

In Table 2, we calculate mean and median values for each of these ratios at the time of the IPO and at 1-yearintervals for the first 5 years after the IPO. In each year, we collect data at the fiscal year end to calculateratios. For example, each year we take the earnings and market price of the stock at the end of the fiscal yearto calculate the E/P ratio. Time zero in our tests refers to the IPO year, not the IPO date. In other words, theE/P ratio at time zero is the first fiscal year end after the IPO.

In Table 2, the mean (median) E/P value for dual-class firms in the initial year is 0.053 (0.046), and forsingles the mean (median) value is 0.048 (0.040). If we invert the means to obtain P/E ratios we obtain 18.87for duals and 20.83 for singles, a difference of $1.96 per dollar of earnings. Correspondingly, the mean(median) EBITDA-to-price value for dual-class firms at the IPO date is 0.150 (0.119), and for singles the mean(median) value is 0.121 (0.095).

In addition to the mean and median values for each ratio for the 5 years following the IPO year, Table 2 alsoshows whether the difference between singles and duals is significant for each year. For the E/P ratio, thedifference in means is significant every year, and the difference in medians is significant in three out of 6 years.

13We also exclude firms with negative earnings, though including these firms does not change our fundamental conclusions.

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Table 2

Pricing multipliers

Event year Full sample Dual class Single class P-value

Earnings-to-price means

Year 0 0.048 0.053 0.048 0.0551*

Year 1 0.056 0.062 0.055 0.0424**

Year 2 0.059 0.074 0.057 0.0001***

Year 3 0.063 0.082 0.061 0.0004***

Year 4 0.060 0.074 0.058 0.0039***

Year 5 0.064 0.076 0.063 0.0606*

EBITDA-to-price means

Year 0 0.124 0.150 0.121 0.0034***

Year 1 0.148 0.175 0.145 0.0127**

Year 2 0.172 0.220 0.166 0.0014***

Year 3 0.192 0.315 0.179 0.0001***

Year 4 0.195 0.237 0.191 0.0607*

Year 5 0.205 0.265 0.199 0.0134**

Earnings-to-price medians

Year 0 0.041 0.046 0.040 0.2058

Year 1 0.048 0.051 0.048 0.3770

Year 2 0.051 0.060 0.051 0.0024***

Year 3 0.053 0.070 0.051 0.0001***

Year 4 0.052 0.060 0.051 0.0696*

Year 5 0.054 0.060 0.054 0.1586

EBITDA-to-price medians

Year 0 0.097 0.119 0.095 0.0000***

Year 1 0.114 0.136 0.111 0.0002***

Year 2 0.129 0.163 0.125 0.0000***

Year 3 0.139 0.184 0.135 0.0000***

Year 4 0.138 0.161 0.137 0.0626*

Year 5 0.138 0.172 0.136 0.0210**

This table presents mean and median ratio values by event year and offer type. E/P is earnings per share divided by price per share.

EBITDA-to-price is earnings before interest, tax, depreciation, and amortization per share divided by price per share. Year 0 refers to the

first fiscal year end following the IPO, Year 1 to the second fiscal year end following the IPO, and so on. P-values refer to tests of equal

means or medians across offer types. Respectively, ***, **, and * denote significant difference at 1, 5, and 10 percent.

S.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115 101

For the EBITDA-to-price ratio, the differences in means and medians are significant in every year, with thevalues for dual-class firms consistently exceeding those of singles. Thus, based on yearly average pricingmultiples, dual-class firms have lower relative valuations than do single-class companies and the valuation gappersists for at least 5 years after the IPO. These differences are economically significant as well. For example, ifwe invert the E/P ratio to the more familiar P/E ratio, the median dual-class P/E equals 25 for singles and 21.7for duals, a difference of 15.2 percent. The median time zero market capitalization for a dual-class firm equals$193 million, so the difference in P/E ratios implies an economically significant dollar difference of about$29.4 million. Pooling results across all years, the median E/P ratio equals 0.058 for duals and 0.049for singles, which translates into P/E ratios of 17.2 for duals and 20.4 for singles, a difference exceeding18 percent.

Though the dual-class firms in our sample come from a wide range of industries, it is possible that adisproportionate share of them compete in industries where we might expect low valuation ratios. Similarly,we know that dual-class firms tend to be larger than their single-class counterparts, so perhaps the valuationdifferences we are seeing merely represent differences in other characteristics and are not tied to the dual-classstructure per se. In the next section, we develop an empirical methodology that controls for the endogenousdecision to issue dual-class equity and controls for various firm-specific characteristics to address theseconcerns.

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4.2. Multivariate analysis

To establish a connection between the valuation differences between singles and duals and the dual-classvoting structure, we must account for the endogenous aspect of the decision that firms make when choosing togo public with dual- or single-class equity as well as control for other firm characteristics that can affectmarket values. In this section we describe a two-stage estimation process, which first models the decision toadopt dual-class shares and then estimates the impact of that decision on firm value at the IPO date andbeyond.

To address the endogeneity issue, we first estimate a probit model where the dependent variable equals onefor dual-class firms and zero for single-class firms. On the right-hand side are variables which may influencefirms’ decisions to adopt dual-class equity. Included in the list of control variables are the natural logarithm ofthe IPO offer size and dummy variables equal to one for IPO firms that receive venture backing, use a high-reputation investment bank, are equity carveouts, are reverse LBOs, have anti-takeover devices in theircharters, do not anticipate paying dividends, and list on NASDAQ. We also control for the number of uses ofproceeds listed in the IPO prospectus and the market return leading up to the IPO date. We use this model toestimate the probability of going public with dual-class shares for each firm in our sample. Next we use thefitted probability in our second stage regressions on valuation multiples.

Our reasons for selecting the variables included in the probit are as follows. Holding all else constant, firmsthat want to raise more money in the IPO have to sell a larger fraction to outside investors, which implies agreater loss of control. The dual-class structure may have more appeal to insiders who value control, butwhose firms have high financing requirements. It is well established in the literature that venture capitalists(VCs) not only provide financing to entrepreneurial firms, but also exercise some pre-IPO control throughboard seats and participation in the selection of top management. However, the IPO presents VCs with an exitopportunity, and presumably they have incentives to maximize the value of their claims at exit. Doing soenhances the returns that VCs provide to their clients and makes it easier to raise additional capital frominstitutional investors to fund new investments. Given the concerns expressed by institutions about investingin dual-class IPOs, we expect the presence of a VC investor to discourage the choice of this particularownership form.

To the extent that selling dual-class equity in an IPO is more difficult than selling single-class shares, firmsthat nevertheless want to go public with multiple share classes may need the additional certification of a highreputation investment banker to ensure the success of the offering. Consequently, we include in the probit ameasure of the underwriter’s reputation. An investment bank is considered to have a high reputation if itsmarket share of dollars raised is greater than the median bank’s market share for the sample. A positivecoefficient on this variable is consistent with our conjecture about the need for investment banker certificationto market dual-class equity to investors.

Zenner et al. (2005) report that tax laws create an incentive for parent companies conducting equitycarveouts to do so using dual-class equity. Accordingly, we include a dummy variable equal to one forequity carveouts in our sample. We also include a dummy for reverse LBOs. We do so because when a publicfirm goes private, control is typically concentrated in the hands of LBO investors and top management.Having concentrated control of the firm, we conjecture that LBO insiders may be more likely than others toattempt to maintain that control by going public as a dual-class entity. Of course, insiders have multipledevices at their disposal that help them maintain control, including anti-takeover provisions such as staggeredboards, poison pills, etc. Viewing these devices as substitutes for dual-class equity, we include a dummy equalto one when firms have at least one anti-takeover measure in their corporate charter other than dual-classshares.

If insiders place a high value on control, then why do they choose to go public at all? One reason is that overtime, a private firm simply becomes too large to manage with private sources of financing. Manager owners oflarge, mature firms may reach a point at which access to the public equity markets becomes necessary, and forthese firms dual-class equity offers a way to continue to exercise majority voting control. To control for thematurity of firms, we include a dummy variable equal to one for those firms who indicate in their prospectusthat they do not intend to pay dividends following the IPO. We also use a dummy variable to distinguishbetween NASDAQ and NYSE/AMEX listed firms.

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Finally, we include several variables designed to capture time series and cross-sectional variation in theincidence of dual-class IPOs. These include a lagged market return, IPO year dummies, and SIC codedummies.

Table 3, Panel A reports our probit estimates. As expected, larger firms and those going public with morereputable underwriters show a higher likelihood of using dual-class equity, while firms with venture backingare less likely to do so. None of the other variables in the model (with the exception of some of the SIC

Table 3

Two-stage regressions of pricing multipliers

Panel A. Dual-class IPO first-stage probit regression

Estimate Std. error Lower limit Upper limit w2 P-value

Intercept �2.045*** 0.493 �3.010 �1.079 17.230 0.0001

LN offer value in millions 0.163*** 0.057 0.051 0.274 8.200 0.0042

Venture-backed �0.420*** 0.101 �0.618 �0.222 17.310 0.0001

High-reputation I-bank 0.319*** 0.095 0.134 0.505 11.340 0.0008

Equity-carveout �0.221 0.142 �0.500 0.058 2.400 0.1210

Reverse-LBO 0.027 0.194 �0.354 0.407 0.020 0.8912

Anti-takeover provisions 0.009 0.163 �0.310 0.329 0.000 0.9537

No dividends anticipated 0.057 0.124 �0.186 0.299 0.210 0.6474

NASDAQ-listed �0.176 0.116 �0.403 0.052 2.290 0.1305

Number of uses of proceeds �0.013 0.034 �0.080 0.053 0.160 0.6934

Lagged market return �0.866 1.375 �3.560 1.828 0.400 0.5285

Log likelihood �641.275

Number of observations 2,555

Panel B. Earnings-to-price second-stage regressions

Year 0 Year 1 Year 2 Year 3 Year 4 Year 5

Intercept 0.079*** 0.087*** 0.096*** 0.104*** 0.096*** 0.095***

Dual-class fitted value 0.018** 0.027*** 0.039*** 0.053*** 0.070*** 0.063***

LN mkt. Cap. In millions �0.010*** �0.011*** �0.013*** �0.014*** �0.012*** �0.012***

Leverage 0.026*** 0.027*** 0.025*** 0.022*** 0.025*** 0.028***

R&D over total assets �0.062*** �0.068*** �0.074*** �0.077*** �0.030 �0.119***

EBITDA-to-assets 0.072*** 0.079*** 0.061*** 0.041*** 0.033** 0.059***

Two-year sales growth �0.019*** �0.012*** �0.004 0.000 �0.001 0.013*

CAPEX-to-sales �0.011*** 0.004* 0.000 0.006 0.000 �0.011

Industry earnings-to-price �0.001* 0.001* 0.000 0.000 0.000 �0.006

Dividend paying firm 0.027*** 0.021*** 0.024*** 0.020*** 0.010** 0.012**

S&P 500 firm 0.009 0.015 0.015 0.022 0.006 0.004

Adj. R2 (%) 26.80 25.31 17.73 16.07 13.92 14.85

Number of observations 1,502 1,436 1,313 1,096 928 792

Panel C. EBITDA-to-price second-stage regressions

Year 0 Year 1 Year 2 Year 3 Year 4 Year 5

Intercept 0.165*** 0.202*** 0.237*** 0.275*** 0.308*** 0.324***

Dual-class fitted value 0.073** 0.120*** 0.276*** 0.390*** 0.274*** 0.274***

LN mkt. cap. in millions �0.021*** �0.035*** �0.051*** �0.061*** �0.063*** �0.065***

Leverage 0.276*** 0.283*** 0.378*** 0.376*** 0.411*** 0.427***

R&D over total assets �0.139*** �0.119** �0.098 �0.068 �0.239** �0.300**

EBITDA-to-assets 0.077*** 0.218*** 0.286*** 0.288*** 0.262*** 0.307***

Two-year sales growth �0.054*** �0.045*** �0.003 0.011 0.030 0.022

CAPEX-to-sales 0.026*** 0.087*** 0.048*** 0.097*** 0.048 0.033

Industry EBITDA-to-price �0.003 0.000 0.000 �0.001 �0.003 �0.004

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Table 3 (continued )

Panel C. EBITDA-to-price second-stage regressions

Year 0 Year 1 Year 2 Year 3 Year 4 Year 5

Dividend paying firm 0.052*** 0.049*** 0.049*** 0.054*** 0.031* 0.027

S&P 500 firm 0.025 0.032 0.049 0.101* 0.099** 0.057

Adj. R2 (%) 36.16 37.49 36.05 35.33 36.69 40.39

Number of observations 1,499 1,427 1,306 1,088 923 785

This table presents two-stage regression analysis of pricing multipliers by event year on offer type. In Panel A the probability that an IPO

firm is a dual-class firm is estimated using a probit regression. The dependent variable is a dual-class (offer type) indicator variable equal to

one for dual-class IPOs and zero otherwise. Market capitalization is the total number of shares outstanding after the offering times the

S&P 500-adjusted first-day closing price. Indicator variables are set equal to one, respectively, for venture-backed, high-reputation I-bank,

equity-carveout, reverse-LBO, anti-takeover provisions, and NASDAQ-listed deals. Number of uses of proceeds is the number of uses of

proceeds listed in the final prospectus. Lagged market return is the compounded daily CRSP value-weighted return over the 22 trading

days preceding the initial public offering. The probit regression includes SIC and IPO year dummies. In Panels B and C, pricing multipliers

are estimated using OLS regression. The dependent variable in Panel B is earnings-to-price, which is earnings per share divided by price per

share. The dependent variable in Panel C is EBITDA-to-price, which is earnings before interest, tax, depreciation, and amortization per

share divided by price per share. Dual-class fitted value is the predicted probability that an IPO firm is a dual-class firm from the first-stage

probit regression. LN mkt. cap. in millions is the natural logarithm of the S&P 500 index-adjusted (to the beginning of 1990) fiscal year-

end market capitalization. Leverage is long-term debt plus short-term debt over total assets. R&D over total assets is research and

development expense divided by total assets. EBITDA-to-assets is earnings before interest, tax, depreciation, and amortization divided by

total assets. Two-year sales growth in year t is the compound annual growth rate in sales from year t�2 to year t. For years 0 and 1, the

2-year sales growth is the compound annual growth rate in sales from year 0 to year 2. CAPEX-to-sales is the total capital expenditures

divided by total sales. Industry multiplier is the respective multiplier for a group of at least five industry-comparable firms. Dividend

paying deal equals one for dividend paying firms. S&P 500 deal equals one if the firm is in the S&P 500 index. Respectively, ***, **, and *

denote significant difference from zero at 1, 5, and 10 percent.

S.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115104

dummies which we have suppressed in the table) have a significant effect on the dual-class decision, though theequity carveout and NASDAQ dummies fall just short of conventional significance levels.

Next, we use the fitted probability values from the probit model on the right hand side of our OLSregressions on pricing multipliers. That is, the variable ‘‘dual’’ in the regressions reported in Panels B and C ofTable 3 is not a dummy variable, but instead is the fitted probability of going public with dual-class equity foreach sample firm. This approach allows us to control for systematic differences between firms choosing single-class versus dual-class stock. We include additional control variables to account for other value-relevantdifferences across firms.

On average, duals are larger than singles, so the pricing differences we observe in the previous section mightbe attributable to differences in firm size rather than to the dual-class equity structure. In our second-stageregressions, we include the logged S&P 500 adjusted market capitalization as a control for firm size.

Firms that investors expect to grow rapidly can have lower valuation ratios. To control for this possibility,we include a sales growth variable in our regressions. Specifically, we include the prior 2-year sales growth ratein the regressions. Measuring sales growth prior to an IPO is sometimes impossible because data prior to theIPO is unavailable. Therefore, in event years zero, one, and two, we measure sales growth from year 0 to year2. This means that the sales growth control variable varies across firms but not across time until event year 3.14

From event year 3 and beyond, the sales growth rate is simply the compound annual growth rate over theprevious 2 event years. In addition, we control for growth opportunities by including in the regression theratio of R&D to assets and the ratio of capital expenditures to sales.

Table 1 indicates that dual-class firms use more leverage, so we include the ratio of long-term plus short-term debt divided by total assets as a control variable. To address the possibility that dual-class firms may gopublic at a later stage in their life cycles, we include a dummy variable equal to one for firms which pay

14Regressions that use pre-IPO sales growth figures when available produce similar results, as do regressions which exclude the sales

growth measure.

ARTICLE IN PRESSS.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115 105

dividends. We also add to the regression model a dummy variable equal to one for firms included in the S&P500. We control for differences in profitability across firms by including the ROA ratio, which equals earningsbefore interest, taxes, depreciation, and amortization divided by total assets. Finally, for each firm in oursample we identify a set of comparable firms that come from the same industry and we use these firms tocalculate an industry-level valuation ratio in each year. This approach controls for inter-industry valuationdifferences that vary over time.15

We run our regression models on each valuation ratio and on each year. In a regression that uses E/P orEBITDA-to-price as the dependent variable, a positive sign on a dual-class fitted value indicates lower relativevaluations for duals. Panel B of Table 3 shows the results when our dependent variable is the E/P ratio, andPanel C reports results when we use EBITDA-to-price as the dependent variable. For both valuation ratios,the regressions confirm the results in Table 2. Dual-class firms have higher E/P ratios in each year. In mostyears the point estimates indicate an E/P difference greater than 0.02, which is large relative to the differencein mean E/P ratios reported in Table 1. Similarly, the regressions in Panel C indicate that dual-class firms havelarger EBITDA-to-price ratios than do singles. The difference is significant in all 6 years, and the pointestimates suggest an economically significant difference in value between the two firm types.16

The coefficients on the control variables, though somewhat mixed in terms of significance, generally indicatethe following. First, larger firms have lower valuations ratios, a somewhat counterintuitive result. In most ofour sample years, larger IPOs (as measured by the IPO offer value) earn higher initial returns than do smallerdeals. This leads to the negative coefficient on size in our regressions. Second, firms with more leverage havehigher valuation ratios. Third, R&D intensive firms sell at lower multiples, consistent with greater growthopportunities. More profitable firms have higher valuation multiples, as do firms that pay dividends. Both ofthese estimates likely captures the tendency of more mature firms with higher cash flows and lower growthoptions to trade at lower prices relative to fundamentals. The signs and significance levels of other controlsvary across time.

With two dependent variables and 6 year-by-year regressions, we have 12 distinct estimates of the effect ofdual-class equity on firm value. All of the point estimates are positive and significant. These findings provideevidence that the market discounts the shares of companies with dual-class equity. The significance andmagnitude of this discount is striking and, indirectly, it provides some evidence of the value that managersplace on having control. It is hard to imagine that firms going through the IPO process fail to hear, either fromtheir investment bankers or from institutional investors on the road show, that issuers pay a price forinsulating managers through a dual-class equity offering.

5. Performance of dual-class firms

If the market discounts dual-class shares, then that discount may reflect investors’ concerns aboutmanagement’s ability to deliver acceptable financial performance in the future. In this section, we try to assesswhether the market’s discount of dual-class shares is consistent with the future stock and operatingperformance of these firms. That is, we ask whether dual- or single-class firms’ exhibit abnormal performanceafter the IPO.

5.1. Long-run stock returns

Our interpretation of the pricing gap between singles and duals is that the market imposes a penalty on firmsthat go public with dual-class equity. Examining post-IPO stock returns provides a way to assess whether themarket’s penalty is consistent with ex post performance. If dual-class shares exhibit abnormal positive returns,then we would conclude that over time investors decide that the initial penalty was too severe. Of course,entrenched managers might perform even worse than expected, in which case dual-class stocks would earnnegative abnormal returns.

15Our approach here is similar in spirit to that used to value IPOs in Kim and Ritter (1999).16We obtain similar results if we use alternative valuation ratios on the left hand side such as the inverse of Tobin’s Q, book-to-market,

or sales-to-price.

ARTICLE IN PRESSS.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115106

By the same token, the valuation gap between singles and duals may reflect excessive optimism about theperformance of single-class companies. Ritter (1991) was the first to document that going-public firms earnedabnormally low returns after their IPOs. Brav and Gompers (1997) argue that underperformance is not ageneral IPO phenomenon, but instead is concentrated among small firms without backing from VCs. In eithercase, because we observe a difference in the valuation ratios of singles and duals that persists for at least 5years beyond the IPO, it does not appear that any underperformance by single-class firms entirely eliminatesthe valuation gap.

We measure the stock returns of dual- and single-class firms for 3- and 5-years horizons following the IPOusing a Fama–French–Carhart four-factor pricing regression. In addition to the original Fama–French threefactors, we also employ Carhart’s momentum factor.17 Loughran and Ritter (2000) propose that whenrunning Fama–French style regressions on IPO firms the factors should be purged of firms that themselveshave recently issued new securities. The results reported in Table 4 use the conventional Fama–French factors;however our results do not change when the Loughran and Ritter purged factors are used. The test ofabnormal performance here is whether the regression intercept is different from zero. We estimate theseregressions in calendar time rather than event time to avoid the problem of overlapping returns for differentIPO firms. We estimate the pricing regressions using equal- and value-weighted portfolios of monthly stockreturns from CRSP.

Panel A of Table 4 reports results by combining firms into calendar-time portfolios using equal weights.Both 3- and 5-year horizons have regression intercepts that are statistically insignificant for duals, but singlesshow positive abnormal returns. Panel B of Table 4, shows results when portfolios are formed using valueweights. At both the 3- and 5-year horizons, duals and singles display no significant abnormal performance. InPanel C of Table 4, we pool all firms together and add a dummy variable for dual-class firms. The dual-classcoefficient is insignificant in both the equal- and value-weighted models at both 3- and 5-year horizons.18

The previous section established that investors discount the shares of dual-class IPO firms relative to sharesissued by new single-class firms. This discount does not reflect a pricing error and is rational in the sense thatdual-class firms show little or no sign of significant positive or negative abnormal stock returns after the IPO.Next, we examine the operating performance of firms in our sample to see if the lower valuations for dualssimply reflect inferior operating results.

5.2. Operating performance

In this section we estimate regressions with measures of operating performance as dependent variables. Inthe valuation-multiple regressions presented in Table 3, we used one measure of profitability, EBITDA overassets, as a control variable. In doing so, we attempted to limit the possibility that dual-class firms trade atlower prices relative to fundamentals simply because they are less profitable than single-class firms. Here, weconsider this possibility more deeply by asking whether dual-class firms underperform singles based onmeasures of operating performance.

To accomplish this, we regress annual operating performance measures on a dual-class dummy and controlvariables for firm size, growth, and industry profitability. The operating performance measures we include inour analysis are return on equity (EBITDA over book equity) and return on assets (EBITDA over totalassets).

Because we anticipate that firm size and profitability are related, we use the natural logarithm of marketcapitalization as a control variable. Market cap figures are adjusted using the S&P 500 index with a base yearof 1990. Likewise, we control for differences in growth rates by including in each yearly regression the rate ofgrowth in sales for the prior 2 years. Finally, for each sample firm, we find at least five matching firms with thesame SIC code and use those firms to calculate an industry-level profitability ratio.

Table 5 shows the results of our operating performance regressions. When we use ROE in Panel A as thedependent variable, the coefficient on the dual-class dummy is insignificant in all six regressions. The signs of

17For a complete description of the four factors, see Fama and French (1993) and Carhart (1997).18These results are robust to several alternative methods for calculating abnormal returns including 3- and 5-year buy-and-hold

abnormal returns (both market adjusted and style adjusted).

ARTICLE IN PRESS

Table 4

Three- and 5-year four-factor calendar-time regressions

Panel A. 4-Factor equal-weighted portfolios regressions

Dual-class IPOs Single-class IPOs

3-Year 5-Year 3-Year 5-Year

Intercept 0.001 0.003 0.008* 0.010***

Market factor 1.209*** 1.123*** 1.120*** 1.121***

SMB factor 0.828*** 0.928*** 1.172*** 1.040***

HML factor 0.103 0.433*** �0.233 �0.081

UMD factor �0.412*** �0.232*** �0.564*** �0.309***

Adj. R2 (%) 67.97 57.54 74.37 81.80

Number of observations 135 158 138 162

Panel B. 4-Factor value-weighted portfolios regressions

Dual-class IPOs Single-class IPOs

3-Year 5-Year 3-Year 5-Year

Intercept 0.002 �0.001 0.000 �0.003

Market factor 1.509*** 1.410*** 1.277*** 1.439***

SMB factor 0.394** 0.442*** 0.776*** 0.525***

HML factor �0.836*** �0.324** �0.525*** �0.392***

UMD factor �0.215* �0.226*** 0.069 �0.004

Adj. R2 (%) 69.65 73.40 79.60 83.72

Number of observations 135 158 138 162

Panel C. 4-Factor pooled regressions

Equal-weighted portfolios Value-weighted portfolios

3-Year 5-Year 3-Year 5-Year

Intercept 0.007 0.009** 0.001 �0.002

Dual-class �0.004 �0.005 0.000 0.000

Market factor 1.164*** 1.122*** 1.390*** 1.424***

SMB factor 1.001*** 0.985*** 0.586*** 0.483***

HML factor �0.068 0.175* �0.678*** �0.359***

UMD factor �0.490*** �0.271*** �0.070 �0.115**

Adj. R2 (%) 70.48 68.84 72.70 78.06

Number of observations 273 320 273 320

This table presents 3- and 5-year four-factor calendar-time regressions by offer type. Dual-class (offer type) indicator equals one for dual-

class IPOs. Market factor is the excess return on a value-weighted market index. SMB factor is the return on a zero investment portfolio

constructed by shorting a portfolio of large firms and investing in a portfolio of small firms. HML factor is the return on a zero investment

portfolio constructed by shorting low book-to-market stocks and buying high book-to-market stocks. UMD factor is the return on a zero

investment portfolio constructed by shorting a low prior return portfolio and investing in a high prior return portfolio. Respectively, ***,

**, and * denote significant difference from zero at 1, 5, and 10 percent.

S.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115 107

the coefficients also flip around with three positives and three negatives. Accordingly, our evidence suggeststhat singles and duals generate comparable EBITDA, relative to book equity.

Switching the dependent variable to ROA in Panel B, we find that the dual-class coefficient is insignificant in5 out of 6 years. The adjusted R2 values are much higher in these regressions, indicating that our modelexplains between 10 and 33 percent of the cross-sectional variation in ROA. In both sets of regressions, ourcontrol variables tend to have the expected signs. For example, the coefficients on the industry profitability

ARTICLE IN PRESS

Table 5

OLS regressions of operating performance

Year 0 Year 1 Year 2 Year 3 Year 4 Year 5

Panel A. EBITDA-to-equity regressions

Intercept �0.408 �0.114 �0.347*** �0.244** 0.216 �0.175

Dual-class 0.061 �0.130 0.152 �0.023 �0.056 0.180

LN mkt. cap. in millions 0.071 0.143 0.121*** 0.089*** 0.032 0.086

Two-year sales growth �0.147 �0.150 �0.081** �0.080 0.043 �0.254*

Industry EBITDA-to-Equity 0.001 0.057* 0.000 0.000 0.017 0.067

Adj. R2 (%) �0.25 0.13 2.05 0.55 �0.31 0.33

Number of observations 1,534 1,622 1,635 1,445 1,231 1,039

Panel B. EBITDA-to-assets regressions

Intercept �0.030 �0.139*** �0.155*** �0.141*** �0.142*** �0.135***

Dual-class �0.029 �0.025 �0.032 �0.012 �0.096*** �0.008

LN mkt. cap. in millions 0.034*** 0.053*** 0.055*** 0.050*** 0.052*** 0.048***

Two-year sales growth �0.065*** �0.034*** �0.022*** �0.025*** �0.043*** �0.027**

Industry EBITDA-to-Assets 1.102*** 0.882*** 0.819*** 0.855*** 0.789*** 0.099***

Adj. R2 (%) 32.42 33.21 31.58 29.86 19.87 9.98

Number of observations 1,545 1,622 1,635 1,445 1,232 1,039

This table presents ordinary least-squares regression analysis of operating performance on offer type. EBITDA-to-equity is earnings

before interest, tax, depreciation, and amortization divided by common equity. EBITDA-to-assets is earnings before interest, tax,

depreciation, and amortization divided by total assets. Dual-class (offer type) indicator equals one for dual-class IPOs. LN mkt. cap. in

millions is the natural logarithm of the S&P 500 index-adjusted (to the beginning of 1990) fiscal year-end market capitalization. Two-year

sales growth in year t is the compound annual growth rate in sales from year t�2 to year t. For years 0 and 1, the 2-year sales growth is the

compound annual growth rate in sales from year 0 to year 2. Industry profitability measure is the respective profitability measure for a

group of at least five industry-comparable firms. Respectively, ***, **, and * denote significant difference from zero at 1, 5, and 10 percent.

S.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115108

control are positive and significant in most cases, firm size is typically positive and significant, and sales growthis negative and significant.

Although not reported, we also run similar regressions to those in Table 5 using the net profit margin andgross profit margin as dependent variables with nearly identical results. The dual-class dummy is insignificant,leading us to conclude that duals exhibit neither better nor worse operating performance relative to singles.Thus, having ruled out the possibilities that the lower prices of dual-class shares reflect pricing errors orinferior operating performance, we shift our focus to search for evidence consistent with a governanceexplanation. If lower dual valuations reflect investors’ concerns that the voting structures adopted by thesefirms serve to entrench managers at the expense of shareholders, then we might observe significant differencesin turnover of senior management between singles and duals. It is to that evidence that we now turn.

6. CEO turnover

The Class B shares that dual-class insiders typically own give senior managers a voting majority even whenthey own a relatively small fraction of cash flow rights. Perhaps the valuation discount applied by the marketto duals reflects investors’ assessment of the difficulty of replacing underperforming dual-class managers. Inthis section we examine the incidence of CEO turnover in our sample to assess the relative job security ofexecutives in single- and dual-class firms. We also measure stock returns surrounding turnover events to seehow the market reacts to turnover news, and to characterize the differences in stock performance leading up toturnover events for singles and duals.

To identify turnover events, we check the identity of each firm’s CEO at the time of the IPO and 5 yearsafterwards. When a firm lists a different CEO in their fifth year, we search electronic news sources to determinewhen the original CEO departed. In constructing our sample of turnover events we also include CEO changesthat are the result of mergers. When an IPO firm in our sample is acquired we include this as a turnover event.

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Table 6

Logistic regression of all CEO turnovers within 5 years of the IPO

Parameter estimate Standard error P-value

Intercept �0.960*** 0.326 0.0032

Dual-class �0.312*** 0.115 0.0067

Venture-backed 0.078 0.069 0.2582

Nasdaq-listed �0.224** 0.093 0.0160

Fraction of institutional ownership 0.110 0.174 0.5265

LN of market capitalization �0.143*** 0.027 0.0001

Book-to-market 0.027 0.054 0.6138

EBITDA-to-assets �0.919*** 0.133 0.0001

Dual-class�EBITDA-to-assets 0.766*** 0.278 0.0059

Leverage 0.636*** 0.143 0.0001

CAPEX-to-sales 0.001 0.002 0.3924

Number of firm years 9,150

Number of single-class firm years 8,257

Number of dual-class firm years 893

Number of turnovers 1,451

Number of single-class turnovers 1,323

Number of dual-class turnovers 128

Correct predictions (%) 64.30

This table presents logistic regression analysis of all CEO turnovers within 5 years of the IPO. The dependent variable is an indicator

variable that is equal to zero for firm years without a turnover and equal to one for the firm year in which the CEO turnover occurs. Firm

years following a turnover are removed. Indicator variables are set equal to one, respectively, for dual-class, venture-backed, and

NASDAQ-listed deals. Fraction of institutional ownership is the percentage end-of-quarter institutional holdings for the quarter in which

the IPO took place. LN mkt. cap. in millions is the natural logarithm of the S&P 500 index-adjusted (to the beginning of 1990) fiscal year-

end market capitalization. Book-to-market is book value per share divided by price per share. EBITDA-to-assets is earnings before

interest, tax, depreciation, and amortization divided by total assets. Leverage is long-term debt plus short-term debt over total assets.

CAPEX-to-sales is the total capital expenditures divided by total sales. The regression includes SIC and year dummies. P-values refer to

tests of parameter estimates equal to zero. Respectively, ***, **, and * denote significant difference from zero at 1, 5, and 10 percent.

S.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115 109

We refer to acquisition-related turnover events, which account for slightly less than 40 percent of all turnoverevents, as ‘‘external’’ turnovers. We use the term ‘‘internal’’ to refer to all other turnover events.19

Though news accounts usually do not explicitly state which CEOs lose their jobs due to poor performance,we conjecture that the internal mechanisms for disciplining dual-class CEOs are weaker than those in single-class firms. When a dual-class firm underperforms, even if there is minimal internal pressure to replace theCEO, an acquirer may find it profitable to pay a large premium to persuade incumbent managers to relinquishtheir control. Smart and Zutter (2003) report higher takeover premiums for dual-class IPOs, and they interpretthis as evidence that dual-class equity protects incumbent managers’ private control benefits. Therefore, weanticipate that external turnover events in dual-class firms are preceded by negative performance.

Within 5 years of the IPO date, we identify turnover events for 58 percent of our dual-class firms. Almost 62percent of the single-class firms in our sample experience some form of CEO turnover; however this differencein turnover percentages is not significant. At horizons shorter than 5 years, the same pattern exists—turnoveroccurs less frequently among dual-class firms, but the difference in raw turnover percentages between singlesand duals is not significant.

Controlling for differences in singles and duals that may be correlated with turnover probabilities changesthe results slightly. From Table 1 we know that, compared to singles, dual-class firms are larger, are less likelyto have venture capital backing, are less likely to list on NASDAQ, and have higher institutional ownership.Table 6 shows estimates from a logistic regression in which the dependent variable equals one if a sample firm

19For those firms with more than one CEO turnover event in the first 5 years, we only include the first instance of CEO turnover. For

internal turnover events, the event date is the announcement date that the CEO will depart. For external turnover events, the event date is

the acquisition announcement date.

ARTICLE IN PRESS

Table 7

Logistic regression of internal CEO turnovers within 5 years of the IPO

Internal turnovers

Forced Unforced

Intercept �3.292*** �1.862***

Dual-class �0.129 �0.164

Venture-backed 0.087 0.102

Nasdaq-listed �0.174 �0.224*

Fraction of institutional ownership �0.008 0.022

LN of market capitalization �0.101 �0.094**

Book-to-market 0.129 �0.104

EBITDA-to-assets �1.238*** �1.007***

Dual-class�EBITDA-to-assets 0.040 0.927**

Leverage 0.494 0.203

CAPEX-to-sales �0.211 0.002

Number of firm years 5,830 7,005

Number of single-class firm years 5,234 6,299

Number of dual-class firm years 596 706

Number of turnovers 151 606

Number of single-class turnovers 137 549

Number of dual-class turnovers 14 57

Correct predictions (%) 64.47 60.55

This table presents logistic regression analysis of internal CEO turnovers within 5 years of the IPO. The dependent variable is an indicator

variable that is equal to zero for firm years without a forced (unforced) turnover and equal to one for the firm year in which a forced

(unforced) CEO turnover occurs. Firm years following a turnover are removed. Indicator variables are set equal to one, respectively, for

dual-class, venture-backed, and NASDAQ-listed deals. Fraction of institutional ownership is the percentage end-of-quarter institutional

holdings for the quarter in which the IPO took place. LN mkt. cap. in millions is the natural logarithm of the S&P 500 index-adjusted (to

the beginning of 1990) fiscal year-end market capitalization. Book-to-market is book value per share divided by price per share. EBITDA-

to-assets is earnings before interest, tax, depreciation, and amortization divided by total assets. Leverage is long-term debt plus short-term

debt over total assets. CAPEX-to-sales is the total capital expenditures divided by total sales. The regressions includes SIC and year

dummies. Respectively, ***, **, and * denote significant difference from zero at 1, 5, and 10 percent.

S.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115110

experiences a CEO turnover in a particular year and zero otherwise. Control variables in the regression includea dummy equal to one if the firm has VC backing when it goes public, a dummy equal to one if the firm lists onNASDAQ, the fraction of outstanding shares held by institutional investors, the logarithm of marketcapitalization, and the book-to-market ratio. Because we want to examine the relation between turnover andfirm performance, we include the prior year’s ratio of EBITDA-to-assets as a control. We also control for theinteraction of EBITDA-to-assets with the dual-class dummy. Finally, we also include measures to control forfirm leverage and growth opportunities.

Table 6 shows that the likelihood of a CEO change is significantly smaller for duals than for singles.However, the magnitude of this difference is relatively small. The probability derivative from this regressionindicates that the likelihood of CEO turnover is about 6.5 percent higher for singles than for duals. Theregression also indicates that poor accounting performance significantly increases the likelihood of CEOturnover, but that is only true for single-class firms. We fail to reject the null that the sum of the coefficients onthe ROA measure and the interaction term is zero, so turnover at dual-class firms does not appear to besensitive to the accounting returns being generated by management. In addition, the logit results indicate thatturnover is more likely at NYSE/AMEX firms, at smaller firms, and at firms that use more leverage.

Next we attempt to separate cases of forced turnover from those that are voluntary. To do so, we follow analgorithm used by Parrino (1997) and Jenter and Kanaan (2006), which relies on CEO age and otherinformation to identify forced turnover. Specifically, using only the sub-sample of internal turnover events(in other words, excluding acquisitions), we search Lexis/Nexis for news articles announcing details of theseevents to see if there is any indication of forced turnover. A turnover is classified as forced if the news articleclearly indicates the CEO left under duress. We also classify a turnover event as forced when the CEO age is

ARTICLE IN PRESSS.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115 111

less than 60 and none of the following are revealed in the news article: (1) the CEO remains as chair of theboard, (2) the CEO dies, (3) the CEO resigns due to health reasons, (4) the CEO retires with the retirementannouncement coming at least 3 months prior to the CEO’s actual departure, or (5) the CEO takes a topmanagement position at another firm within 2 months of the turnover announcement. Conversely, a turnoverevent is classified as unforced when the CEO retires for health reasons or dies, and when the CEO age is atleast 60 and there is no clear statement indicating that she was forced out. We omit a small number of eventsfor which we cannot find press coverage that allows us classify the event.

Table 7 reports estimates from two logit models, one in which the dependent variable equals one when aforced turnover event occurs and zero otherwise, and the other in which the dependent variable equals onewhen an unforced event takes place.20 In both models, the coefficient on the dual-class dummy is negative butinsignificant. However, both models display the same relation between turnover and performance as in Table6. That is, both forced and unforced turnover events are preceded by low accounting performance. However,in both regressions we fail to reject the null that the incremental effect of performance on turnover for duals iszero, as the sum of the coefficients on performance and the interaction term is not significant.21 Most of theother control variables in the regressions are insignificant.

When we examine stock returns surrounding turnover events, some interesting differences emerge betweensingles and duals, particularly when we separate internal and external turnover events. The distribution ofinternal versus external turnover is fairly similar for singles and duals. About 39 percent of all single-classturnover events are acquisition related, whereas, for duals that figure is roughly 42 percent. Panel A of Table 8reports cumulative abnormal returns (CARs) over several event windows for firms experiencing any type ofCEO turnover. We calculate a CAR simply by cumulating the daily difference in the CRSP value-weightedindex from the individual firm’s return. For both singles and duals, the turnover event follows a period ofbelow-average stock performance, with mean CARs of �15.3 percent for duals and �8.1 percent for singlesbetween the IPO date and the day before the turnover event. Though the negative pre-event CAR for duals isalmost twice as large in absolute value as the pre-event CAR for singles, the difference is not statisticallysignificant. During the three trading days surrounding the event, the market greets turnover news as positivefor both types of firms with abnormal returns just under 6 percent for singles and duals. Interestingly, over the126-day trading period following the event, single-class firms earn a small but significant positive CAR, whileduals earn neither negative nor positive abnormal returns.

Panel B of Table 8 examines abnormal returns only for internal (non-acquisition related) turnover events. Inthe pre-event period, single-class firms earn significant CARs of �16.8 percent, suggesting a disciplinarymotive for the event. In contrast, duals earn normal returns leading up to internal turnover events, consistentwith the hypothesis that internal CEO turnover events at dual-class firms are uncorrelated with performance.Furthermore, the market reaction immediately surrounding announcements of internal turnover events ismute, with no abnormal returns for singles or duals. Following internal turnover events, singles outperformduals by 10.4 percent.

In Panel C of Table 8 we examine CARs for external (acquisition-related) turnover events. Except for the 3days surrounding the event, the differences in performance between duals and singles is striking. Single-classfirms exhibit no abnormal pre-event performance. For dual-class firms the story is quite different. Duals thatare acquired earn negative CARs that are both highly significant and large in economic terms, butunderperformance reverses after the event.

In summary, Table 8 shows that for both singles and duals, poor performance precedes turnover events.However, while that statement is, on average, true for all turnover events, the pre-event abnormal returns

20We thank Dirk Jenter and Fadi Kanaan for providing their sample of turnover events which we cross-checked against our own

classification scheme. Though only about 100 firms appear in both their sample and ours, we found only two differences in how these

events were classified. Note that we exclude from these regressions all firm years involving firms that are subsequently acquired. Likewise,

the forced model excludes the data for firms that eventually have an unforced turnover event, and vice versa. This means that the

comparison group, when y ¼ 0, is the set of firms that experience no turnover events during our sample period.21It is interesting to note that the effect of performance on turnover for single-class firms is significant for both forced and unforced

events, and that the magnitude is also similar for both types. Perhaps this speaks to the difficulty of distinguishing forced from unforced

events. Alternatively, since our unforced events include a relatively high percentage of older CEOs, perhaps as they approach retirement

age, maybe they elect to leave on their own when performance is poor.

ARTICLE IN PRESS

Table 8

Cumulative abnormal returns for CEO turnover events

Event window Dual class Single class Difference

Panel A. All turnovers mean CARs

(IPO, �1) �0.153* �0.081*** �0.071

Number of observations 145 1,452

(�1, +1) 0.059*** 0.056*** 0.003

Number of observations 145 1,448

(+1, +126) �0.001 0.040*** �0.041

Number of observations 126 1,325

Panel B. Internal turnovers mean CARs

(IPO, �1) �0.046 �0.168*** 0.122

Number of observations 84 886

(�1, +1) �0.004 �0.005 0.002

Number of observations 84 882

(+1, +126) �0.078 0.026 �0.104**

Number of observations 84 881

Panel C. External turnovers mean CARs

(IPO, �1) �0.299* 0.056 �0.355**

Number of observations 61 565

(�1, +1) 0.146*** 0.152*** �0.006

Number of observations 61 565

(+1, +126) 0.153*** 0.069*** 0.083**

Number of observations 42 444

This table presents mean event-time cumulative abnormal returns (CARs) by offer type. Abnormal returns are calculated by subtracting

the CRSP value-weighted daily return from the corresponding IPO return. CARs are calculated by cumulating the daily abnormal returns

over the specified event window. Respectively, ***, **, and * denote significant difference from zero at 1, 5, and 10 percent.

S.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115112

depend on whether the firm is a single or dual and what type of turnover event we examine. For single-classfirms, poor stock returns occur prior to internal turnover, but not prior to external turnover. Duals exhibit justthe opposite pattern. Before, during, and after internal turnover events, the stock performance of duals isunremarkable. In contrast, dual-class external turnover events are preceded by severe underperformance andfollowed by significant positive abnormal returns. Although we are hesitant to make a definitive statementabout the role of internal and external turnover in disciplining managers, our results are consistent with thehypothesis that internal mechanisms discipline underperforming single-class managers, whereas the evidencefor duals is consistent with the hypothesis that external turnover events displace underperforming andentrenched managers.

7. Dual-class unifications

If dual-class voting structures entrench managers, and if shareholders discount dual-class shares as a result,then we should expect the market to react positively when firms eliminate the Class B shares owned byinsiders. Of course, a firm’s decision to unify its share classes is endogenous and could signal many things inaddition to a change in voting structure. Nevertheless, it is interesting to examine what happens when firmsunwind dual-class voting arrangements.

Of the 253 dual-class IPOs in our sample, we identify 37 firms that unify their share classes and continue asindependent firms.22 In most cases, share unifications occur over a period of time, often without a singleannouncement date. This of course limits our power to detect the market’s response to unifications. Even so,for the 37 firms unifying their share classes, we identify the effective dates of these transactions and conduct anevent study around that date. Table 9 reports our findings.

22That is, we are not including in this group firms that become takeover targets and eliminate Class B shares as part of an acquisition.

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Table 9

Unification CARs

Mean CAR

Event window Market model Fama–French model

(�2, +2) 0.027** 0.027**

(�5, +5) 0.052** 0.055*

Number of observations 37 37

This table presents mean cumulative abnormal returns (CARs) around unifications of dual-class firms to single-class firms. Day 0 is the

effective date of the unification. Market model uses the CRSP value-weighted index as the risk factor. Fama–French model uses excess

market returns, SMB, and HML as the risk factors. Both models are estimated over (�261, �6). Respectively, ***, **, and * denote

significant difference from zero at 1, 5, and 10 percent.

S.B. Smart et al. / Journal of Accounting and Economics 45 (2008) 94–115 113

Using 5- and 11-day windows around the effective date, and using both a market model and a three-factorFama–French model to calculate abnormal returns, we find a significant, positive reaction to the event.During the �5 to +5 event window, stock prices of firms unifying their share classes rise by more than 5percent. This represents a narrowing of between one-third to one-fifth of the valuation gap between singlesand duals discussed earlier. Given the uncertainty we face about clear announcement dates for unifications,the positive returns observed during the event window likely understate the true magnitude of shareunifications on stock prices.

8. Conclusion

This paper provides evidence that investors discount the shares of dual-class IPOs relative to newly publicsingle-class firms. Dual-class firms trade at lower prices relative to earnings and EBITDA than do single-classfirms. These valuation differences are not driven by systematic differences between singles and duals related tosize, industry, profitability, leverage, or growth opportunities. The discount assessed to dual-class equity at theIPO date appears to be rational in the sense that post-IPO dual-class abnormal returns are zero. Moreover, wefind no evidence that inferior operating results achieved by duals can explain the discount we observe.

The likelihood of CEO turnover is slightly higher for single-class firms than for duals, though that differenceis not large in absolute terms. However, turnover events at single-class firms appear to be linked to the firm’saccounting performance, but that is not true at dual-class firms. Abnormal returns leading up to turnoverevents vary depending on the type of firm and the type of turnover. For single-class firms, negative abnormalreturns precede instances of internal CEO turnover, but not external turnover. For dual-class companies, justthe opposite pattern exists. These patterns suggest that internal governance mechanisms and the externalmarket for corporate control play different roles in disciplining managers at single-class and dual-class firms.

Though only a handful of the dual-class firms in our sample eventually unify their share classes, those thatdo experience positive abnormal returns around the effective date of the unification. Collectively, these resultssuggest that investors discount dual-class shares because the superior voting rights held by insiders makes itdifficult for outsiders to replace incumbents.

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