Do private equity investors take firms private for

different reasons?∗

Jana P. Fidrmuc†

Warwick Business School

Peter Roosenboom‡

RSM Erasmus University

Dick van Dijk§

Econometric Institute

Erasmus University Rotterdam

April 2007

AbstractIn recent years, the going private market experienced a considerable boom insize and also became more interesting for private equity investors. This papershows that the higher involvement of private equity investors affects the goingprivate market as these investors approach firms with different characteristicsrelative to the traditional management buy-outs. Our results on a sample of212 UK going private transactions completed in the period 1997-2003 suggestthat private equity backed deals differ from management sponsored deals with-out any backing of private equity investors in four ways. First, even thoughboth types of deals suffer from market undervaluation, the mis-pricing is largerfor management sponsored deals. Second, it is only management sponsoreddeals that suffer low financial visibility as their stock’s analyst coverage andfrequency of trading are low. Third, Jensen’s free cash flow hypothesis seemsnot to apply to private equity backed deals as they have shortage of cash, lowdebt levels, and pay high dividends. Finally, the two types of deals differ inownership structure. Private equity backed deals seem to have higher owner-ship by financial institutions and their ownership is less concentrated.

Keywords: Going Private Transactions, Corporate Governance, Private Eq-uityJEL Classification: G32, G34

∗We would like to thank participants of the Conference on Mergers and Acquisitions in Exeterand research seminars at the Warwick Business School, Manchester Business School, XFi Centrefor Finance and Investment, University of Exeter, RSM Erasmus University and University ofAntwerp. Also, we are grateful to Ian Garrett, Abe de Jong and Ian Tonks for useful comments onan earlier draft of the paper and Shanaz Raja, Huiyan Xu and Hans van der Weijden for valuableresearch assistance.

†Warwick Business School, University of Warwick, Coventry CV4 7AL, United Kingdom, E-mail: Jana.Fidrmuc@wbs.ac.uk (corresponding author)

‡Department of Financial Management, RSM Erasmus University, P.O. Box 1738, NL-3000 DRRotterdam, The Netherlands, e-mail: proosenboom@rsm.nl

§Econometric Institute, Erasmus University Rotterdam, P.O. Box 1738, NL-3000 DR Rotter-dam, The Netherlands, e-mail: djvandijk@few.eur.nl

1 Introduction

We investigate whether private equity investors take firms private for different rea-

sons than the firm’s managers. The paper builds on the heterogeneity hypothesis of

Halpern et al. (1999) who argue that “the population of LBOs is heterogeneous and

understanding the differences among LBOs is important in understanding both why

each type of firm engaged in this transaction and what they did after the transac-

tion” (page 282). We use this idea of heterogeneity to highlight the involvement of

private equity funds. While undervaluation, low financial visibility, high managerial

ownership and free cash flow are key factors motivating management to take their

firm private, we examine whether private equity investors pursue similar goals or

may have other incentives to be involved in a public to private transaction.

Several empirical papers have examined the reasons for why publicly listed firms

decide to go private (see Maupin et al., 1984; Lehn and Poulsen, 1989; Denis, 1992;

Opler and Titman, 1993; Halpern et al., 1999 for US and Weir et al., 2004, 2005

for the UK). However, most of this empirical evidence investigates the going private

deals of the 1980s, focuses on testing of Jensen’s (1986) free cash flow hypothesis

and does not consider the effect of private equity involvement. Recently, private

equity houses have considerably increased their investment and at the same time

changed their strategy concerning the going private market. In the 1980s, private

equity investors often engaged in highly leveraged transactions, many of which were

seen as hostile by incumbent management. Nowadays, private equity investors are

often looking to partner with management. They are interested in sound strategic

reasons to justify any going private deal and low valuation on its own is not enough.

In fact, private equity investors are after fundamentally strong businesses. Their

typical target is a long-term player, has leading market share within their defined

niche, and has a strong customer base, good growth potential and good margins (The

Deal, 24 November 2003). Operational improvements have become more important

than they were before in the 1980s (Business Week, 27 February 2006).

In this paper, we conjecture that going private deals that have backing of a private

equity house have different characteristics and go private for different reasons than

1

management buy-outs that do not have the backing of a private equity house. In

other words, the population of going private firms is not homogeneous with respect to

the reasons for the deal and pooling all deals together may result in biased results.

The population of public to private (PtP) deals can be partitioned into separate

groups: (i) private equity backed going private transactions and (ii) management

buyouts where management initiates the deal without any direct involvement of

private equity funds (we refer to this group as ‘management sponsored deals’). We

show that these two groups have their own unique characteristics and their own

reasons for going private. To cover the whole population of PtP deals, our analysis

also includes a third type of ‘other’ going private transactions that are initiated by

different parties such as non-executive directors, wealthy private individual investors,

industrial firms or financial investors that are not private equity investors. This

group is included for completeness of our model. That is, we want to be sure that

any differences found between the private equity backed deals and management

sponsored deals are not arising due to omitting the remainder of the going private

firms.

We consider four different aspects that may motivate management sponsored

deals and argue that private equity sponsored deals do not share these incentives

as they target firms with shortage of cash that can be turned around and make

the investment to pay off (Gompers and Lerner, 2001). The first of our hypotheses

is that market undervaluation may motivate managers to take their firms private.

However, information asymmetry and undervaluation may be less important for

private equity backed deals because private equity investors are primarily interested

in a strategic justification for their transactions (Gompers and Lerner, 2001). The

second hypothesis concerns the recently recognized proposition that low financial

visibility plays an important role in going private decisions as it undermines the

main benefits of public ownership (see Boot et al., 2006 and Mehran and Peristiani,

2006). Again, we suggest that management sponsored deals are motivated by low

financial visibility of their stock because firm managers weight the costs and benefits

of the public listing. In contrast, private equity investors profit from the strategic

2

advantages of the deal and so financial visibility may be less important for them.

Third, we explore the free cash flow idea of Jensen (1986) separately for the two

groups of going private transactions. In particular, we propose that private equity

backed firms may have less agency problems associated with free cash flows as, in

fact suggested by Gompers and Lerner (2001), they are more likely to have low cash

balances. Finally, we highlight the differences in pre-buyout ownership structure

between the two types of firms. Elitzur at al. (1998) propose that managers who

have a large amount of their personal wealth invested in the company may be more

likely to decide in favor of a going private deal. In contrast, we argue that private

equity backed deals may have high institutional ownership as institutional investors

may be willing to support their deal. We note that several of the above hypotheses

have not been explored in this context so far, which constitutes another contribution

of this paper.

In our empirical analysis, we examine a sample of 212 UK going private trans-

actions that occurred during the period 1997-2003 and compare them to randomly

selected UK firms that remained listed on the London Stock Exchange.1 Using

multinomial logistic models, we find support for our conjecture that going private

firms are heterogeneous with respect to the private equity involvement in the deal

and, at the same time, the two groups of going private firms are different from

the firms that remained public. First, the results suggest that even though both

types of deals suffer market undervaluation, the mis-pricing is significantly larger

for the management sponsored deals. Second, we show that it is only management

sponsored deals that suffer from low financial visibility as they fail to attract suf-

ficient investor interest. The private equity backed deals are not distinguishable in

this respect from the control group of firms that remain public. Third, our results

suggest that the free cash flow hypothesis does not apply to private equity backed

deals. As expected, they seem to have lower cash balances and, at the same time,

pay higher dividends than management buyouts. Finally, the two types of deals are

1In the US, the total value of deals went up to $65 billion over the period 1997-2002 (Renneboogand Simons, 2005). The UK deals total £30 billion over the period 1997-2003. This makes the UKgoing private market second only to that in the US.

3

heterogeneous in pre-buyout ownership structure. Our results show that firms in

private equity backed deals have higher ownership by financial institutions and their

ownership is less concentrated relative to management sponsored deals.

The plan of the paper is as follows. In Section 2, we put forward our four

heterogeneity hypotheses. Section 3 describes our dataset, methodology and provides

basic descriptive statistics of our sample. Section 4 presents our empirical results.

Section 5 concludes.

2 Heterogeneity in going private transactions

In this paper, we propose that private equity involvement in going private transac-

tions is associated with different firm characteristics compared to transactions initi-

ated and led solely by managers. Thus, we argue that in order to get a better picture

of the reasons and characteristics of going private transactions, we should not pool

all going private firms together but rather separately evaluate the reasons for going

private in two different types of deals: (i) going private transactions backed by pri-

vate equity investors, and (ii) going private transactions led by the firms’ executive

directors without any backing of private equity investors (management sponsored

deals).

As active private equity involvement in going private transactions is a relatively

new phenomenon, the literature explaining the motives of private equity investors

to engage in these deals is very limited.2 Therefore, it is not easy to come up

with rigorous hypotheses backed up with theoretical models or previous empirical

findings. The literature, however, is quite useful in providing explanations for the

motives of managers (as opposed to private equity investors) engaging in the buy-

out deals. In fact, we form four relatively broad hypotheses explaining why firms

go private: (i) undervaluation of the going private firms relative to their peers in

the market; (ii) their low visibility in the form of illiquidity of their stock and low

analyst coverage; (iii) high free cash flows and tax shields; and (iv) high managerial

ownership. At the same time, we suggest that these four motivations may be less

2Gompers and Lerner (2001) is one of the few references.

4

applicable to private equity backed deals as the general perception of private equity

investors’ activities suggests that private equity houses are interested in turning

companies around within a three to five year horizon and making decent returns

on their investment (Business Week, 27 February 2006). Also Gompers and Lerner

(2001) support this view. The following subsections explain the four possible ways

in which the two types of going private transaction may differ.

2.1 Heterogeneity in undervaluation

Due to information asymmetry between management and outside shareholders, it

is possible that the market value of a company is above or below its fundamental

value (Merton, 1987). In going private transactions, the perceived undervaluation

may play an important role as it potentially limits management’s ability to use the

benefits available to public companies as, for example, the accessibility of funds

required to finance new investment projects or acquisitions (Allen and Gale, 1999;

Pagano et al., 1998). A survey among US managers who underwent a going private

transaction indicates that the perceived undervaluation is indeed one of the primary

reasons for managers to take their firm private (Maupin et al., 1984). Moreover,

undervalued firms are more likely to attract hostile takeover interest (Lehn and

Poulsen, 1989) that may lead to managers losing their jobs (Lowenstein, 1985).

Therefore, we conjecture that undervaluation of firms increases the likelihood of

managers taking their firms private as this allows the managers to keep control over

the firm at relatively low cost and avoid (hostile) takeovers.

Outside investors may also value a company more than the current market value

because they perceive unexploited growth opportunities. In particular, private equity

investors may target firms because of poor management that has not been taking full

advantage of the firms’ growth potential possibly due to a lack of cash (DeAngelo,

1990). Gompers and Lerner (2001) argue that private equity investors look for

companies that have the potential to evolve in ways that create value and at the

same time face problems in raising funds via regular bank debt or public equity.

This aspect may be more important in the UK versus the US since the UK venture

capital and buyout markets have traditionally been more closely linked (Toms and

5

Wright, 2005). In addition, private equity investors may also follow a ‘buy-and-

build’ strategy in which they focus on growth opportunities within an industry and

hire new managers with industry experience as needed. In such a buy-and-build

strategy the private equity investor acts as an industry consolidator taking over a

number of smaller rival firms. In this case, the value of the going private transaction

to the investor depends more on the investor’s previous portfolio investments rather

than the firm being undervalued (Smit and DeMaeseneire, 2005). Moreover, it seems

plausible that private equity investors have less superior information relative to the

management concerning the undervaluation of the firm.

In short, we predict that undervaluation plays a role in both types of going pri-

vate transactions but to a different degree. Undervaluation is expected to be more

important in management sponsored private equity transactions. In contrast, un-

dervaluation is less important in transactions backed by private equity because these

active investors could create additional value using their unique resources such as

their network and industry expertise and do not have private information advantage.

2.2 Heterogeneity in visibility

If a firm’s shares are thinly traded, being public may not be worth the cost (Bolton

and von Thadden, 1998). Boot et al. (2006) highlight liquidity and low cost of capital

as important benefits of public versus private ownership. Furthermore, thinly traded

stocks have lower analyst coverage in general and are at risk of being neglected by

investors when taking their investment decisions (Merton, 1987). Thus, firms that

are not able to attract adequate level of investor recognition have to bear the high

cost of stock exchange listing while not taking enough advantage of the benefits of

being a public company (Mehran and Peristiani, 2006). Therefore, we propose that

low financial visibility may motivate managers to take their firm private.

In contrast, the predominant and most important economic activity for private

equity investors is the restructuring of their targets. Often, it is argued that it is

easier to fix a private company rather than a public one. This is because public

shareholders are strongly focused on quarterly results, which often is in sharp con-

flict with long term strategic goals of multiyear restructuring efforts. Thus, going

6

private firms with private equity backing may actually want to hide from relatively

high visibility and scrutiny of the public market. At the same time, private equity

investors prefer projects where monitoring and selection costs are relatively low and

where the information asymmetry costs are less severe (Amit et al., 1998). So, rela-

tive to the management sponsored PtP firms, we expect private equity backed deals

to be associated with less problems in attracting financial visibility and investor

recognition.

2.3 Heterogeneity in free cash flows

Most of the empirical evidence concerning going private transactions so far is based

on Jensen’s free cash flow hypothesis. Jensen (1986) proposes that debt-financed

going private transactions may provide a solution to firms in cash-rich, slow growth

and declining industries. The main argument maintains that firms with large cash

balances but low growth prospects are vulnerable to conflicts of interest between

managers and shareholders over payout and investment policies. To mitigate this

problem, managers can increase dividends and thus provide payout of current cash

that would otherwise be invested in low-return projects or be wasted. However, a

promise of increased dividend is not credible as managers can easily reduce dividends

in the future at low cost. In contrast, issuing debt in exchange for stock is a credible

strategy to limit free cash flows because unpaid interests lead to bankruptcy.

The empirical studies testing the free cash flow hypothesis in the context of going

private transactions provide mixed results. Lehn and Poulsen (1989) document that

undistributed cash flow is a significant determinant of a firm’s decision to go private.

Also Opler and Titman (1993) show that high cash flow firms that also have a low

Tobin’s q (which they use to measure future growth prospects) are more likely to

undertake an LBO. In contrast, Kieschnick (1998) argues that after accounting for

choice based sampling, outliers in the data, and potentially misspecified variables,

prior growth rate and level of free cash flows are not significant determinants of the

odds of going private for the Lehn and Poulsen (1989) dataset. Furthermore, Weir

et al. (2004) also fail to find significance of free cash flows in determining the odds

of going private for UK firms. All the empirical papers, however, pool all going

7

private firms together and treat them as a single homogeneous group. Halpern et

al. (1999), in contrast, distinguish two groups of going private firms according to

the level of management ownership before the transaction but do not test the free

cash flow hypothesis on the two groups separately.3

We propose that private equity backed deals, as opposed to management spon-

sored deals, do not match the profile of Jensen’s cash-rich firms. Intuitively, firms

that search for private equity backing are more likely to suffer low cash balances and

high growth opportunities. (Gompers and Lerner, 2001). In contrast, managers of

cash-rich firms with low growth prospects may see the potential of leveraged transac-

tions as they may be ware of their ill-specified incentives and use the excess cash to

fund the going private transaction or to service new debt and gain control over their

firms (Fox and Marcus, 1992). Moreover, the leveraged transaction allows managers

to avoid the prospect of hostile takeover and/or shared control with an active private

equity investor who typically demands board representation and a say in the firm’s

long-term strategy (Cotter and Peck, 2001). Therefore, we predict that large cash

balances may motivate management sponsored going private transactions but not

private equity backed deals.

2.4 Heterogeneity in pre-buyout ownership structure

Halpern et al. (1999) and Elitzur at al. (1998) argue that managers who have a

large amount of their personal wealth invested in the company may prefer to diversify

their portfolio while keeping or increasing their control over the firm. Managers with

large equity stakes therefore have incentives to initiate a leveraged buyout and use

new debt to decrease their wealth invested in the firm. It is relatively inexpensive

for these managers to acquire a majority of the votes and force the leveraged buyout

given their large equity stakes and any potential information asymmetry (Elitzur et

al., 1998). Thus, we expect that managers with high ownership stakes in their own

firm are more likely to take their firm private and do not seek backing by private

3In particular, they propose that for the group of low management ownership firms, third-partytakeover pressures force management to consider a third party led buyout or face the prospectof hostile takeover. In the case of high prior management ownership group, managers hold largeundiversified portfolios and so have an incentive to take cash out of their firm.

8

equity investors.

Private equity investors typically look for support of a number of incumbent

blockholders before they take a firm private. This increases their chances of the deal

being successful. In fact, the private equity investor usually contacts the existing

blockholders in order to receive irrevocable undertakings wherein the existing block-

holders promise to accept the private equity investor’s offer (Wright et al., 2007).

After receiving the support of several blockholders, the private equity investor makes

a public offer for the remaining shares at the same price. These irrevocable under-

takings are easier to obtain when ownership is concentrated in the hands of a small

number of outside shareholders. Institutional investors may be of special importance

as most of them are passive and not interested in monitoring management closely

themselves (Faccio and Lasfer, 2000). They are likely to sell their shares in case they

are able to negotiate a premium price and earn a return on their otherwise illiquid

investment. Therefore, we expect that high institutional ownership increases the

likelihood of a going private transaction backed by private equity.

3 Data and methodology

3.1 Sample selection

Our original sample consists of 221 non financial firms that have gone private in the

United Kingdom during the years 1997-2003. We identify these public-to-private

(PtP) transactions from the database of the Centre for Management Buyout Re-

search (CMBOR). For all the PtP firms, we also obtain the offer documents ac-

companying the going private transaction from Thomson Research. We use these

documents to determine whether the deal is backed by private equity investors or not.

This is our primary classification criterion to decide upon whether a deal falls in the

private-equity backed group. In case the transaction is not backed by a private equity

house we further examine whether any of the firm’s executive directors are involved

in the deal. If this is the case, the transaction is coded as a management-sponsored

deal. In all other cases, the transaction is classified into the ‘other’ category. This

category includes deals backed by non-executive directors, wealthy families or insti-

9

tutional investors other than private equity houses and is included for completeness

of our model.4

We do not have data for 9 PtP firms and, therefore, our final sample of going

private transactions consists of 212 firms that are relatively evenly distributed be-

tween 90 private equity backed deals, 69 deals led by management and not backed

by private equity investors and 53 ‘other’ deals led by other parties.

In order to get a more detailed picture of the characteristics of the going private

transactions in general and the three individual types of going private transactions

in particular, we contrast the going private firms with firms that remained publicly

listed. We opt for a random sample of control firms that remained public: in each

of the years of the sample period of 1997-2003 we randomly select 200 control firms

from a population of around 1200 firms that continue to be publicly traded in a

given year. The sampling procedure allows for a control firm to be included in

the sample more than once. In total, we collect data on 1400 control firm-years

that cover 960 different control firms. For both the PtP firms and control firms,

we get market prices from Datastream, financial statement data from Worldscope,

and hand-collect ownership structure and board composition information from Price

Waterhouse Coopers’ Corporate Register (various issues).

In the going private literature it is common to use a matched control sample

of firms that remained public. Firm size and industry classification are the most

commonly used matching criteria. In general, the sampling method – whether we

use a random sample as in this paper or a matched sample – does not pose any

advantages or disadvantages except for the case where the characteristics used for

the matching process are important determinants of the going private process. We

prefer random to matched sampling because we believe that firm size is a relevant

factor influencing the decision to go private. Using a matched sample would not

allow us to test for this possibility.

4Note that the private equity backed deals may be both with and without management involve-ment. In fact, private equity funds often look for management involvement in their deal. We believeit is the presence of private equity involvement that matters for the deal characteristics. Privateequity investors are attracted to certain type of firms and managers behave differently conditionalon private equity interest.

10

3.2 Descriptive statistics

Table 1 lists our variable definitions and Table 2 shows summary statistics for all

PtP firms together, the three types of PtP deals separately, and the control firms.

All variables are trimmed at the 1st and 99th percentiles, except the ownership

and illiquidity variables. We test for differences in means and medians between the

control firms and PtP firms and among the three types of PtP firms. We use a t-test

for equal means allowing for unequal variances and a Mann-Whitney U-test for equal

medians. Below we first discuss the differences between the complete sample of PtP

firms versus the control firms and then we focus on the differences between the two

main types of going private firms. The latter comparison highlights the heterogeneity

idea proposed in this paper. We only discuss the statistically significant differences

for both the mean and the median.

- insert Tables 1 and 2 about here -

Firms that go private are valued less than control firms as indicated by their

lower market to book ratios. They also experience more takeover rumours. PtP

firms’ shares are traded less actively than the shares of the control firms and are

followed on average by fewer analysts. Moreover, the going private firms have higher

equity stakes held by executive directors and financial institutions than the control

firms. As a result, ownership concentration, as measured by the Herfindahl index,

is higher in PtP firms than in control firms. In addition, firms that go private have

less cash and marketable securities, lower sales growth, higher payout ratio and are

more profitable than control firms.

Turning to the differences between the different groups of going private firms, we

observe that the market to book ratio of the private equity backed deals (that we use

as a measure of stock market valuation) is higher indicating that firms that go private

with the help of private equity investors are less undervalued than the management

sponsored deals. They also have higher analyst following and their shares are more

actively traded. At the same time, the private equity backed deals are larger relative

to the deals sponsored by management. Moreover, the private equity backed PtP

11

deals have significantly lower executive ownership and higher ownership by financial

institutions relative to the management sponsored transactions. Also, their pre-

transaction ownership concentration is lower.

Finally, we compute the free cash flow proxy of Lehn and Poulsen (1989) used in

the going private literature so far. Table 2 shows that private equity backed deals

have higher free cash flow in the previous year. However, DeAngelo and DeAngelo

(2006) argue that the firm’s current cash flow is not a suitable measure of managerial

opportunism as it does not reflect the stock of resources at managers’ disposal. What

matters is managers’ access to the stock of liquid assets at all points in time which is

better reflected in the firm’s cash and marketable securities. In fact, free cash flow is

highly correlated with the firm’s profitability which indicates that it is restricted to

affect managerial incentives only at ’a distribution point’ (DeAngelo and DeAngelo,

2006) and, thus, it may not be enough to encompass all resources at managers’

disposal and therefore is not able to generate empirically valid predictions. This

suggests that the firm’s cash level is a more suitable proxy for company free cash

at the disposal and discretion of the managers. Table 2 reports that private equity

backed deals have less cash on their balance sheet than the management sponsored

deals.

3.3 Model

We employ multinomial logistic regression (MNLR) models to examine the hetero-

geneity hypotheses developed in the previous section. As mentioned before, we

divide our sample of UK firms into four different groups: (1) private equity backed

PtP deals, (2) management sponsored PtPs, (3) other PtPs, and (4) non-PtPs. We

denote the observed group for firm i by the variable yi, which can take the discrete

values 1, 2, . . . , M , where M = 4 in our case. In the MNLR model the probability

that firm i will belong to group m, conditional on the (k × 1) vector of explanatory

variables xi consisting of a constant and firm characteristics, is given by

P [yi = m|xi] =exp(β ′

mxi)∑M

l=1 exp(β ′lxi)

, for m = 1, . . . , M . (1)

12

For identification purposes, we set the coefficients for the non-PtP group of firms

equal to 0, that is β4 = 0.

Estimation of the coefficients in the MNLR model in (1) is straightforward by

means of maximum likelihood, except for the following caveat. Our data set is not

a random sample from the population of all firms. In particular, while we include

all known PtP deals during the period 1997-2003, each year we only sample 200 of

the firms that remain listed, which in total equal 1200, on average. This implies

that PtP firms are considerably overrepresented in our sample compared to the

underlying population of firms. Not accounting for this selective sampling would lead

to biased estimates of the intercepts and incorrect standard errors for all estimated

coefficients; see Kieschnick (1998) and Fok and Franses (2002) for detailed analysis

of selective sampling in the context of binary and ordered logit models, respectively.

The problem can be remedied by defining modified probabilities as

P̃ [yi = m|xi] =γmP [yi = m|xi]∑M

l=1 γlP [yi = l|xi], for m = 1, . . . , M, (2)

where γm is the fraction of firms in group m that is included in the sample. Hence,

in our case γ1 = γ2 = γ3 = 1 while γ4 = 1/6. The correct likelihood function, which

is used for parameter estimation then makes use of these corrected probabilities.

The effects of the firm characteristics xi on the probabilities that a firm engages

in the different types of PtP deals is a nonlinear function of the model parameters

βm, such that interpretation of these parameters is not straightforward. For inter-

pretation of the model, it is useful to consider the log-odds ratio of group m versus

group l, defined as

log

(

P [yi = m|xi]

P [yi = l|xi]

)

= x′i(βm − βl). (3)

This shows that firms with a larger value for xi,j more likely belongs to group m

than to group l if (βm,j − βl,j) > 0, where xi,j indicates the j-th element of xi, and

βm,j and βl,j are the corresponding coefficients. Note that this does not necessarily

imply that the probability that firm i belongs to group m increases with xi,j , as the

the odds ratios of group m versus the other categories also change. The net marginal

effect of a change in xi,j on the group probability follows from the partial derivative

13

of P [yi = m|xi] with respect to xi,j , which is given by

∂P [yi = m|xi]

∂xi,j

= P [yi = m|xi]

(

βm,j −

M∑

l=1

βl,jP [yi = l|xi]

)

. (4)

The sign of this derivative depends on the sign of the term between brackets, which

may be positive or negative depending on the value of xi. Hence, the sign of the

marginal effect of xi,j on P [yi = m|xi] will not always correspond with the sign of

the coefficient βm,j . Also note that the marginal effect depends on the values of the

other explanatory variables in xi, denoted as xi,−j . In order to obtain a clear view on

the effect of the variable of interest xi,j one should therefore consider∂P [yi=m|xi,j ]

∂xi,j=

∫

xi,−j

∂P [yi=m|xi]∂xi

dxi,−j, integrating out the effects of these other explanatory variables.

In practice this can be done by averaging (4) across all realizations of xi,−j in the

sample for each value of xi,j . In the empirical analysis presented in the next section

we follow a more direct approach by considering the group probabilities P [yi =

m|xi,j] themselves to gauge the effects of the different firm characteristics on the

probabilities of the different types of PtP deals. We do average out the effects of other

explanatory variables by computing P [yi = m|xi,j ] = 1N

∑N

n=1 P [yn = m|xi,j, xn,−j],

where N is the sample size.

An important assumption underlying the MNLR model in (1) is independence of

irrelevant alternatives, meaning to say that the odds ratio of, for example, PE-backed

and management-led PtP deals does not depend on the inclusion of the third group

of PtP transactions, which can also be seen from (3). We examine the validity of this

assumption by means of the specification test developed by Hausman and McFadden

(1984).

Our heterogeneity hypothesis of PE-backed and management-led PtP transac-

tions implies that certain firm characteristics such as free cash flow and management

ownership affect the relative probabilities of a firm belonging to the different groups.

Put differently, in the MNLR model the coefficients βm,j should differ across groups

m. For an individual variable, xi,j say, the null hypothesis of no heterogeneity across

groups m and l can easily be tested by means of a likelihood ratio test of the re-

striction βm,j = βl,j. The same holds for a given sub-set of the explanatory variables

included in the model. Testing whether there is no heterogeneity at all is slightly

14

more involved, and effectively boils down to testing whether two groups can be com-

bined into one. This is done by means of the likelihood ratio test developed by

Cramer and Ridder (1991). Finally, we should remark that the above likelihood

ratio test statistics also enable us to assess in which respects the third group of

PtP transactions is similar to the private equity backed and management sponsored

deals.

4 Results

The main contribution of this paper is to show that the population of the PtP firms

is heterogeneous and that the reasons for going private are different for private equity

backed deals versus management sponsored deals. To examine this hypothesis, we

estimate multinomial logistic regressions with private equity backed, management

sponsored and other PtP deals analyzed against the control firms that remained

public.

Table 3 reports the estimation results. To account for industry and time effects,

all regressors are taken in deviation from industry median and annual median values.

Our model treats the non-PtP control firms as the omitted category. Hence, Table

3 reports coefficients for private equity backed, management sponsored and other

deals separately in subsequent columns. These coefficients show how the explana-

tory variables affect the probability of going private through the particular type of

transaction relative to the probability of remaining public. The last three columns

in Table 3 show p-values for a likelihood ratio test of equal parameters among the

three types of PtP deals. Thus, these columns show significance of pair-wise coeffi-

cient differences among the going private types. Finally, the last two lines of Table

3 show p-values for the likelihood ratio test of Cramer and Ridder (1991) that all

parameters except the intercept are equal for the corresponding two groups and the

independence of irrelevant alternatives test, respectively.

- insert Table 3 about here -

Overall, the results suggest that going private transactions are indeed hetero-

geneous. The no heterogeneity test of Cramer and Ridder (1991) suggests that all

15

three going private groups have different deal characteristics from the non-PtP firms

as well as from each other. Moreover, the independence of irrelevant alternatives

test reported in the last row indicates that the choice of the particular type of going

private deal (whether the deal is indeed supported by a private equity house or is

fully led by the management or is sponsored through another party) is fully inde-

pendent. Thus, the decision to go private is made at the same time as the decision

about the type of the deal. In other words, it is not the case that the firm decides

about going private first and then looks for possible methods how to realize it. The

independence of irrelevant alternatives test also shows that the multinomial logistic

regression is the preferred estimation method and that this method fits the setting

of going private decision better than a nested logit model, for example.

4.1 Undervaluation

The results in Table 3 suggest that perceived undervaluation plays an important

role in management sponsored deals, but less so in private equity backed deals. The

coefficient for the market to book ratio is significantly negative for both types of

PtP deals showing that both the private equity backed and management sponsored

deals are on average undervalued relative to the firms that remain public. However,

the management sponsored deals are also significantly more undervalued (at the five

percent level) relative to the private equity backed deals. These results show that

low levels of market to book ratio increase the probability of going private but do

not necessarily support the link between market to book ratio and undervaluation.

Therefore, we perform several sensitivity checks that in our view provide additional

evidence that undervaluation is an important factor motivating (management spon-

sored) PtP deals.

First, we account for the possibility that the low market-to-book ratio reflects

low growth prospects or generally low performance of the PtP firms. Therefore,

we include average sales growth over the last three years and return on assets as

control variables in the regression in Table 3. The two variables are, however, not

statistically significant. It seems that neither growth prospects nor profitability

distinguish the going private firms from each other and from the firms that remained

16

public and suggests that market to book may indeed capture undervaluation of the

firms.

Second, we take advantage of the fact that takeover interest is often associated

with undervalued firms. The positive and significant (at the one percent level)

coefficients for rumours in Table 3 indicate that both PtP types experience relatively

high takeover interest in the period before the deal compared to the control group

of firms that remained publicly listed. Even though the two coefficients are not

statistically different from each other, a closer analysis of the relationship between

takeover interest and market to book ratio reveals interesting differences between the

management sponsored versus private equity backed deals. The literature suggests

that undervaluation of buyout firms increases chances of takeover bids from third

parties (Lehn and Poulsen, 1989) and then this jointly motivates management to

take their firms private. Our data confirm this conjecture for the management

sponsored deals: The frequency of rumours conditional on market to book being

below the sample median is 39%, 48%, and 23% for the management sponsored deals,

private equity backed deals and non-PtP firms, respectively, whereas the frequency

for firms with market to book above the sample median is 0%, 39%, and 18%,

respectively. So, the management sponsored PtP deals experience a very strong

negative correlation between market to book and takeover rumours. In other words,

the management sponsored deals are attractive for third parties only in case they

have relatively low market to book ratio and we believe this result supports the

conjecture that undervaluation together with takeover threat play a very important

role in motivating the management sponsored deals. This effect is not present for

the private equity backed deals which then indicates that undervaluation does not

play such a primary role for the latter type.

To confirm this conjecture in a regression setting, we augment our basic regression

model from Table 3 with an interaction term between market to book and rumours.

Results are reported in Table 4 and support a strong complementary effect between

market to book and rumours for the management sponsored deals: The coefficient

for the interaction term is, as expected, negative and significant and, moreover,

17

the rumour coefficient becomes statistically insignificant. The private equity backed

deals are not affected in the same way. In short, our data show that high takeover

interest strengthens the effect of low market to book ratio on the probability of

management sponsored PtP deals and supports our hypothesis on undervaluation.

- insert Table 4 about here -

Third, insider trading patterns in firms in our sample reported in Table 4 provide

further evidence supporting the view that our results on market to book ratio pick

up the effect of private information. If undervaluation is indeed one of the reasons

for going private, we could expect that managers’ trading in advance of the event

may partially reveal the importance of that information. In fact, Harlow and Howe

(1993) document significant increase in trading by insiders prior to the announcement

of US management-led buyouts over the period from 1980 to 1989. They show,

however, that this abnormal pattern arises not from increases in purchases but from

abnormally low levels of stock sales. Harlow and Howe (1993) argue that this passive

insider trading strategy is preferred by managers as it reduces their liability risk. In

line with this existing evidence, our management sponsored deals should experience

abnormally low insider sales relative to the private equity backed deals and non-PtP

firms.

In order to show this, Table 4 includes two dummy variables that reflect the

insider purchase and sale patterns of executive directors of firms in our sample. In

particular, the executive director purchase (sales) dummy is set to one in case an

executive director purchased (sold) some shares of his/her own firm in the calendar

year prior to the announcement and set to zero otherwise. Our results confirm that

managers of the management sponsored deals tend to sell their shares significantly

less often than their counterparts in non-PtP firms and private equity backed deals.

We do not see any significant differences for the purchase patterns. Thus, our results

are in line with Harlow and Howe (1993) and support our conclusion that firm

undervaluation is more important in motivating management sponsored deals versus

private equity backed deals.

18

Figures 1a to 1d show the probability that a firm belongs to a particular type of

going private transactions as a function of market-to-book, rumours, sales growth,

and ROA, respectively. These graphs provide additional intuition for the results

as the coefficient estimates in Tables 3 and 4 indicate only the sign but not the

functional form of the relation. It is important to note that the variable on the

x-axis is measured as deviation from the year and industry specific medians. In

addition, the effect of the remaining explanatory variables from the regression is

averaged out. Figure 1a supports our result from Table 3 that undervaluation tends

to be associated with the management sponsored deals and to a lesser extent with

the private equity backed deals. In fact, Figure 1a shows that the probability of

a management sponsored deal sharply increases as market to book falls whereas

this is not the case for the private equity backed deals. In Figure 1b we see that the

probability of both private equity backed and management sponsored deals increases

with the number of takeover rumours but the two functions are quite close to each

other.

In summary, our results suggest that market undervaluation plays an important

role for the management sponsored PtP transactions that have significantly lower

market-to-book ratio relative to the control firms that remained publicly listed.

In contrast, for private equity investors, undervaluation of their target is not so

important. On average, private equity targets are significantly less undervalued

than the management sponsored deals.

4.2 Visibility

The results in Table 3 also support our second conjecture that low financial visibility

may increase chances of management sponsored deals whereas it is not important

for private equity backed deals. The most widely accepted empirical measure of firm

visibility is the number of analysts following a firm (O’Brien and Bhushan, 1990).

Analyst reports are documented to be the primary source of information for most

buy-side investors (Baker et al., 2002). In Table 3, the coefficient for the number of

analysts following a firm is negative and significant at the ten percent level for the

management sponsored deals whereas it is positive but not significant for the private

19

equity backed deals. Importantly, the two coefficients are significantly different at

the ten percent level indicating that management sponsored deals suffer significantly

lower financial visibility relative to private equity backed deals.

We also employ an alternative measure of investor interest that is available for

all firms in our sample and is based on frequency of price changes. It is intuitive to

expect that stocks with low investor interest would be traded relatively infrequently.

Low trading frequency or days with no trading are then associated with no price

movements and zero daily returns. In contrast, stocks that attract investor inter-

est are traded relatively frequently and experience frequent (small) price changes

that reflect constant price discovery. So, our measure of thin trading measures the

frequency of zero daily returns in the previous calendar year and high levels indi-

cate low trading frequency, investor interest and financial visibility (Fidrmuc et al.,

2006). The results in Table 3 show that thin trading (high fraction of zero returns)

is associated with higher probability of a management sponsored deal relative to

both the non-PtP control firms as well as the private equity backed deals (both sta-

tistically significant at the one percent level), which confirms our financial visibility

hypothesis.

Firm size may also be closely associated with financial visibility of a firm (O’Brien

and Bhushan, 1990). The coefficient estimates for size (log of total book value of

assets), however, do not support this claim. In fact, the coefficient for management

sponsored deals is positive and significant indicating that management sponsored

deals on average are larger. Inspecting the correlation matrix, however, we find high

positive correlation between size and thin trading. This means that inclusion of the

thin trading variable in the regression strongly affects the coefficients for size. As

both variables are important and their correlation does not affect other coefficients,

we opt to include both total assets and thin trading in our model. However, the

coefficients for size should be interpreted with caution as excluding the thin trading

variable from the regression results in size being negative and statistically significant

(at the one percent level) for the management sponsored deals and not significant

for the private equity backed deals.

20

Another important issue is that our results for financial visibility may be driven

by size. Put differently, it may be the case that our PtP firms, especially the man-

agement sponsored deals, are relatively small and then of course, they are thinly

traded and not followed intensively by analysts. To make a stronger case for our

visibility hypothesis, we would like to see that the PtP firms are distributed rela-

tively evenly across the size spectrum of our sample and show lower levels of analyst

coverage and higher levels of thin trading across all sizes. In order to check this,

Table 5 shows mean values of our analyst coverage and thin trading measures as

well as frequencies of private equity backed and management sponsored deals across

size deciles (measured by total assets). Table 5 confirms that analyst coverage and

thin trading are indeed positively correlated with size. However, it also shows that

even though the going private firms are significantly underrepresented among the

smallest and largest firms, PtP firms are relatively evenly spread across the remain-

ing 8 middle size deciles. Thus, this indicates that the association between financial

visibility and probability of going private is not due to the size effect.

- insert Tables 5 and 6 about here -

To push this argument further, Table 6 reports the same statistics by size deciles

separately for the non-PtP firms, private equity backed firms and management spon-

sored firms, respectively. Panel A, reporting the means for the non-PtP firms, con-

firms the overall trend that analyst coverage is increasing and thin trading falling

with size. The same pattern is reflected in Panel B for the private equity backed

firms. Overall, the private equity backed deals seem to be slightly less frequently

traded but equally monitored by analysts relative to the non-PtP firms across all

size deciles. In contrast, Panel C shows sharply lower analyst coverage and thinner

trading for management sponsored deals relative to the non-PtP firms across all size

deciles. Thus, this shows that the management sponsored deals suffer lower market

visibility relative to both private equity backed deals as well as the non-PtP deals.

Moreover, this effect is clearly present across all size deciles and, thus, is not driven

by size.

21

Another closely related argument is that the going private firms might be more

likely to be listed on the Alternative Investment Market (AIM) with lower listing

requirements which in turn would drive the visibility result. The last column in

Tables 5 and 6 reports AIM listing frequency among our firms and does not detect

any significant trend. Also, in an unreported regression, we include an AIM dummy

as an additional regressor. As the coefficients are not significant and the other results

remain unaffected we conclude that AIM listing does not drive our results.

Figures 1e and 1f provide visual intuition and further support for our findings.

They show clearly that the probability of management sponsored deals is sharply

increasing with low analyst coverage and thin trading. In contrast, this is not the

case at all for the private equity backed firms. Overall, our results support the

hypothesis that the going private firms are heterogeneous with respect to financial

visibility. Management sponsored deals seem to suffer both low analyst coverage and

infrequent trading and therefore have less reasons to remain publicly listed whereas

this is not the case for the private equity backed deals.

4.3 Free cash flows

Our third hypothesis conjectures that more cash rich firms are more likely to go

private via a management sponsored deal whereas low cash levels increase the prob-

ability of a private equity backed deal. The coefficients for our cash variable in Table

3 support the hypothesis as they show that, relative to the non-PtP control firms, the

management sponsored deals and the private equity backed deals have significantly

higher and lower cash levels, respectively. The coefficients are significant at the ten

and one percent level, respectively and are statistically different from each other at

the one percent level. Still, it is important to control for sales growth as Jensen

(1986) suggests that the agency problems of free cash flow concern mature firms

with low growth prospects. The sales growth variable, however, is not significant

implying that growth opportunities do not affect the probability of going private.

Thus, controlling for growth opportunities, our results suggest a sharp difference in

the effect of cash levels: management sponsored deals seem to be cash rich while

private equity backed deals suffer very low cash levels.

22

An alternative interpretation of the low cash levels of private equity backed deals

might be firm insolvency and inability to pay interest payments. To account for

this possibility, we check average interest coverage across deciles of cash levels of the

private equity backed firms. This exercise, however, shows that interest coverage

is not related to cash level of the private equity backed deals. Moreover, including

interest coverage in the regression does not result in a significant coefficient for the

private equity backed deals nor does this affect the cash coefficient.5 Therefore, we

conclude that this alternative explanation is not plausible for our setting.

As a sensitivity check for our measure of managerial opportunism, we also es-

timate excess cash recently proposed in the literature on determinants of cash re-

serves level (Opler et al., 1999 and Dittmar and Mahrt-Smith, 2007). In line with

this methodology, we estimate a cash regression that should determine normal cash

levels used in companies to cover their liquidity needs.6 Residuals of this regres-

sion then measure cash reserves held in excess of those needed for operations and

investments. These resources are most at risk of being wasted at the managers’

discretion (Dittmar and Mahrt-Smith, 2007). Note also that excess cash satisfies

the stock requirement of DeAngelo and DeAngelo (2006) as discussed in section 3.2

above and therefore is a very good candidate for the proxy that should reflect man-

agerial opportunism in cash rich firms. In Table 4, we partition the cash variable

from Table 3 into two separate variables: normal cash and excess cash representing

the fitted values and residuals from the cash regression, repectively. The results are

equally strong now despite fewer observations due to data availability for the cash

regression. They show that high excess cash levels increase the probability of man-

agement sponsored deals whereas low excess cash levels are associated with private

equity backed deals. The regression in Table 4 includes also free cash flow and shows

5All results are available upon request.6Following Opler et al. (1999) and Dittmar and Mahrt-Smith (2007) we regress the natural

logarithm of cash over net assets (total assets minus cash and marketable securities) on the naturallogarithm of net assets, market to book adjusted for net assets, net working capital over net assets,capital expenditures over net assets, R&D over net assets, free cash flow over net assets, leverageand a dividend dummy. All variables are industry and time adjusted. Due to missing data, we areable to obtain only 1,437 observations for excess cash compared to 1,579 observations in our fullsample.

23

that free cash flow do not affect the decision to go private.

Finally, we include the payout ratio as an additional explanatory variable be-

cause firm dividend policy is closely related to the free cash flow hypothesis (Jensen,

1986): high dividend payout may mitigate the agency problems associated with free

cash flow. DeAngelo and DeAngelo (2006) provide a different argument. They com-

bine capital structure decisions with payout policy and constraints on liquid assets

(cash) and claim that firms can develop three potential sources of future financial

flexibility: cash accumulation, preservation of debt capacity and reputation for sta-

ble and substantial equity payouts. Cash accumulation as argued above increases

managerial opportunism. High leverage reduces the agency costs but it utilizes debt

capacity and therefore also financial flexibility. High ongoing equity payouts control

agency costs without high leverage and thus preserve the firm’s option to borrow,

and they also increase its future equity issuance capability. Therefore, it is impor-

tant to control for payout as well as leverage when analyzing liquidity position of

firms.

Table 3 shows that the two payout coefficients are positive but only the coef-

ficient for the private equity backed deals is significant (at the ten percent level).

The coefficients for leverage are of opposite signs – private equity backed deals are

less leveraged and management sponsored deals are more leveraged relative to the

non-PtP firms – but neither of the coefficients is statistically significant at a conven-

tional level. However, results in Table 4 are stronger: private equity backed firms

pay significantly more dividends relative to the non-PtP firms and at the same time

have lower leverage relative to both non-PtP firms and management sponsored deals

(all at the five percent level). This suggests that private equity deals pay high divi-

dends that reduce managerial opportunism of excess cash and keep high reputation

to equity holders while, at the same time, preserve firm future borrowing potential.

Clearly, private equity backed deals do not seem to suffer from agency problems and

managerial opportunism. If anything, private equity backed going private transac-

tions experience shortage of cash. Management sponsored deals, in contrast, enjoy

high excess cash levels while at the same time enjoying relatively common payout

24

and leverage. This indicates that management sponsored deals may have problems

with managerial opportunism. However, alternative and perhaps more plausible ex-

planation for the high excess cash levels is that managers build up financial slack to

finance the deal.

Finally, we turn to the functional form of the probability functions shown in

Figures 2a to 2c. Figure 2a shows that as the cash levels increase, the probability

of a private equity backed deal decreases whereas the probability of a management

sponsored deal increases, showing a contrasting characteristic of the private equity

versus management sponsored deals. At the same time, the payout ratio in Figure 2b

has a more dramatic effect on the probability of a private equity backed deal relative

to a management sponsored deal. Hence, the results suggest that the management

sponsored deals are more likely in firms with higher cash levels that may suffer the

agency problems associated with excess cash. Alternatively, availability of high cash

levels may simply make a leveraged buyout more feasible. In contrast, private equity

backed deals are associated with negative excess cash levels, high dividend payments

and low leverage.

4.4 Ownership structure

Our final hypothesis highlights the heterogeneity in pre-transaction ownership and

proposes that high executive ownership may increase chances of a management spon-

sored deal whereas high ownership by financial institutions increases chances of a

private equity backed deal. Table 3 shows that the coefficient for executive owner-

ship in the management sponsored deals is indeed positive and significant at the one

percent level indicating that high executive ownership increases the probability of a

management sponsored transaction relative to the firm staying public. However, ex-

ecutive ownership also increases the probability of a private equity backed deal. The

corresponding coefficient is positive and significant at the five percent level. This

supports the view that private equity investors look for support from management in

their deals and choose firms with relatively high managerial ownership. Even though

the MS coefficient is slightly larger, the difference between the two coefficients is not

statistically significant. So, higher executive ownership increases the chances of both

25

private equity backed as well as management sponsored deals.

In contrast, we find significant differences with respect to ownership by financial

institutions, confirming the second part of our hypothesis. High ownership by finan-

cial institutions increases the probability of a private equity deal relative to both

non-PtP firms as well as management sponsored deals (both statistically significant

at the one percent level). These results are visualized in Figures 2c and 2d. In par-

ticular, the probability of management sponsored as well as private equity backed

deals increase with executive ownership. However, only the probability of private

equity backed deals increases with ownership by financial institutions.

We also check for ownership concentration measured by the Herfindahl index with

the expectation that the management sponsored deals have higher concentration of

ownership by the management as other blockholders in these deals are relatively

small. The results suggest that ownership concentration matters: it is significantly

higher for the management sponsored transactions and lower for the private equity

backed deals.7

In summary, our heterogeneity in ownership hypothesis is partially supported.

Our results suggest that a management sponsored deal is more likely when the man-

agers have higher control and concentration of ownership. At the same time, how-

ever, also private equity backed deals are more likely in firms with high managerial

ownership suggesting cooperation between private equity investors and managers in

the deals. The heterogeneity in pre-transaction ownership structure stems mainly

from the differences in ownership by financial institutions. Willingness of the in-

stitutional investors to accept a deal proposed by a private equity party seems to

increase the success of the transaction and thus motivates private equity investors

to proceed with the deal in the first place. At the same time, very low ownership

by financial institutions in the management sponsored deals may also reflect low

investor interest for these deals that was discussed in section 4.2 above.

In addition to size, our model includes two other control variables: tax and

standard deviation of returns. The argument for taxes is that highly leveraged

7All results are available upon request.

26

buyout transactions are associated with positive interest tax shield that may further

increase incentives for going private. Our results in Table 3 suggest that taxes

increase the odds of going private for the management sponsored deals but this is

not supported in Table 4. The standard deviation of returns seems to be associated

with a higher probability of a private equity backed deal (significant at the five

percent level). This confirms the Gompers and Lerner (2001) suggestion that targets

of private equity deals are associated with higher uncertainty and riskiness.

5 Concluding remarks

This paper proposes that the recent increase of private equity investors’ involvement

in going private transactions may affect sources of value creation and reasons ex-

plaining why firms decide to go private. Our analysis shows that the probability

whether a deal is backed by private equity investment or is purely managed by in-

siders of the firm depends on different firm characteristics. Drawing upon several

theoretical arguments (including Jensen’s free cash flow hypothesis, ownership struc-

ture, and the framework of costs and benefits of being publicly listed versus privately

owned commonly applied in the IPO literature) we derive four testable hypotheses

explaining reasons for management sponsored PtP deals: (i) undervaluation, (ii)

financial visibility, (iii) free cash flow, and (iv) pre-transaction ownership structure.

At the same time, we propose that private equity backed deals do not share the same

characteristics as they are usually backed by strategic reasons and are not able to

raise funds via traditional market sources. Therefore, pooling all PtP firms together

may lead to weak estimation results and fail to detect the important determinants

motivating management sponsored as well as private equity backed deals.

In summary, our empirical results for the UK provide convincing evidence that

the population of going private firms is indeed heterogeneous. We show that private

equity backed and management sponsored deals have different reasons for their deci-

sion to take a firm from the stock market. Firms involved in management sponsored

deals are significantly undervalued, not followed by analysts, have relatively high

cash levels, high executive ownership, and ownership concentration. The private

27

equity backed deals, in contrast, have high executive ownership and high ownership

by financial institutions but have lower ownership concentration. These firms have

shortage of cash, low debt levels, and pay high dividends. Thus, our results suggest

that the management sponsored deals fit the characteristics of the Jensen’s cash rich

firms with strong management that are relatively thinly traded in the stock market,

not followed by analysts and are significantly undervalued. This suggests that the

benefits of remaining publicly listed fall short of the costs and the firms are more

likely to go private. In contrast, benefits of stock market listing seem to be larger

for the private equity backed deals as they are more actively traded, followed by

analysts and not so much undervalued. However, they seem to be short of cash.

Private equity backing may provide the necessary financing and extra know-how for

their strategic restructuring.

28

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Smit, H.T.J. and W. DeMaeseniere, 2005, “The Role of Investor Capabilities in Public-to-Private Transactions,” Working paper, Erasmus University Rotterdam.

Toms, S. and M. Wright, 2005, “Divergence and Convergence within Anglo-AmericanCorporate Governance Systems: Evidence from Divestment in the US and UK 1950-2000,” Business History, 47, 267–295.

Weir, C., D. Laing, M. Wright, 2005, “Incentive Effects, Monitoring Mechanisms andthe Market for Corporate Control: An Analysis of Factors Affecting Public to PrivateTransactions in the UK,” Journal of Business Finance and Accounting, 32, 909–943.

Weir, C., D. Laing, M. Wright, and A. Burrows, 2004, “Financial Distress Costs, IncentiveRealignment, Private Equity and the Decision to Go Private: Public to Private Activityin the UK,” working paper, Aberdeen Business School.

Wright, M., C. Weir, D. Laing, and A. Burrows, 2007, “Irrevocable Commitments andGoing Private”, European Financial Management, forthcoming.

30

Table 1: Variable Definitions

Total assets total assets (in millions) WorldscopeTotal debt total debt divided by total assets WorldscopeROA net income divided by total assets WorldscopeSt. dev. of stock returns standard deviation of stock returns over the period

from January to December of the calendar year be-fore PtP transaction

Datastream

Market to book market capitalization plus total debt divided by totalassets

Worldscope

Rumours number of takeover rumours during two calendaryears before PtP transaction

Lexis Nexis andSDC M&A

Sales growth sales growth during 3 financial years before PtPtransaction average

Worldscope

Director buying dummy variable that is set to one in case executivedirectors were buying shares of their own firm duringJanuary to December of the calendar year before PtPtransaction or in the previous year for the non-PtPfirms and zero otherwise

Hemmington Scott

Director selling dummy variable that is set to one in case executivedirectors were selling shares of their own firm duringJanuary to December of the calendar year before PtPtransaction or in the previous year for the non-PtPfirms and zero otherwise

Hemmington Scott

Analysts following number of analysts following the company in Decem-ber of the calender year before PtP transaction

IBES

Thin trading fraction of days with zero percent return during Jan-uary to December of the calendar year before PtPtransaction or in the previous year for the non-PtPfirms

Datastream

Ownership ofexecutives percentage of shares held by executive directors of

the companyCorporate Register

non-executives percentage of shares held by non-executive directorsof the company

Corporate Register

financial inst. percentage of shares held by financial institutions(e.g. pension funds, mutual funds, insurance com-panies, banks, venture capitalists)

Corporate Register

other firms percentage of shares held by industrial firms Corporate Registerindividuals percentage of shares held by persons that are not

directors of the companyCorporate Register

Herfindahl index sum of squared equity stakes held by the individualblockholders

Corporate Register

Cash cash and marketable securities divided by total assets WorldscopeFree cash flows (ebitda - taxes - interest - cash dividend - stock re-

purchases) divided by salesWorldscope

Investment capital expenditures divided by sales WorldscopePayout ratio cash dividend divided by the sum of net income and

depreciationWorldscope

Tax income taxes divided by sales WorldscopeAIM dummy variable that is set to one in case the firm

is listed on the alternative market with lower listingrequirements

Corporate Register

31

Table 2: Comparison of types of firms

t-test p-valuesMeans non-PtP PE PE MS

Variable non-PtP PtP PE MS Oth. PtP MS Oth. Oth.

Panel A

Total assets 692.4 202.8 223.3 100.8 300.8 0.000 0.008 0.211 0.013

Total debt 0.199 0.206 0.190 0.213 0.223 0.303 0.210 0.157 0.396

ROA −0.019 0.002 0.031 −0.019 −0.020 0.103 0.087 0.055 0.490

St.dev. of return 15.38 15.34 15.50 15.46 14.90 0.443 0.479 0.203 0.246

Market to book 1.528 0.879 1.050 0.702 0.818 0.000 0.000 0.010 0.052

Rumours 0.322 0.642 0.778 0.493 0.604 0.000 0.041 0.151 0.227

Sales growth 0.154 0.109 0.129 0.126 0.053 0.021 0.475 0.046 0.093

Director buying 0.341 0.316 0.267 0.377 0.321 0.231 0.073 0.250 0.261

Director selling 0.169 0.137 0.167 0.072 0.170 0.109 0.032 0.481 0.057

Analysts following 3.028 2.651 3.122 1.913 2.811 0.011 0.000 0.215 0.012

Thin trading 0.530 0.633 0.566 0.733 0.618 0.000 0.000 0.074 0.001

Ownership ofexecutives 0.088 0.121 0.098 0.198 0.060 0.006 0.001 0.080 0.000

non-executives 0.032 0.029 0.026 0.039 0.020 0.215 0.122 0.241 0.044

financial inst. 0.187 0.243 0.302 0.170 0.237 0.000 0.000 0.039 0.030

other firms 0.030 0.034 0.020 0.024 0.072 0.287 0.345 0.004 0.008

individuals 0.124 0.139 0.072 0.165 0.218 0.141 0.000 0.000 0.094

Herfindahl index 0.076 0.106 0.078 0.114 0.143 0.000 0.008 0.004 0.132

Cash 0.132 0.114 0.085 0.140 0.130 0.069 0.017 0.051 0.389

Free cash flows −0.084 −0.036 0.033 −0.130 −0.029 0.117 0.054 0.119 0.179

Investment 0.063 0.079 0.026 0.095 0.149 0.225 0.046 0.028 0.225

Payout ratio 0.188 0.195 0.286 0.190 0.045 0.417 0.060 0.005 0.066

Tax 0.020 0.022 0.022 0.019 0.026 0.214 0.293 0.235 0.144

Number of obs. 1400 212 90 69 53

continued on next page

32

continued from previous page

U -test p-valuesMedians non-PtP PE PE MS

Variable non-PtP PtP PE MS Oth. PtP MS Oth. Oth.

Panel B

Total assets 67.01 65.18 66.50 51.05 125.3 0.488 0.015 0.123 0.001

Total debt 0.173 0.177 0.172 0.170 0.187 0.493 0.399 0.293 0.385

ROA 0.044 0.038 0.057 0.038 0.020 0.349 0.002 0.000 0.054

St.dev. of return 14.54 14.42 14.62 14.64 13.90 0.369 0.288 0.084 0.255

Market to book 1.013 0.764 0.830 0.670 0.769 0.000 0.000 0.040 0.058

Rumours 0.000 0.000 0.000 0.000 0.000 0.000 0.107 0.430 0.160

Sales growth 0.071 0.048 0.052 0.028 0.041 0.013 0.238 0.049 0.155

Director buying 0.000 0.000 0.000 0.000 0.000 0.275 0.117 0.295 0.298

Director selling 0.000 0.000 0.000 0.000 0.000 0.228 0.155 0.487 0.179

Analysts following 2.000 2.000 3.000 1.000 2.000 0.036 0.000 0.146 0.005

Thin trading 0.571 0.663 0.598 0.750 0.663 0.000 0.000 0.013 0.002

Ownership ofexecutives 0.015 0.027 0.020 0.122 0.009 0.001 0.000 0.002 0.000

non-executives 0.002 0.001 0.001 0.004 0.002 0.491 0.085 0.406 0.105

financial inst. 0.151 0.221 0.306 0.112 0.189 0.000 0.000 0.024 0.054

other firms 0.000 0.000 0.000 0.000 0.000 0.423 0.304 0.006 0.023

individuals 0.055 0.064 0.031 0.111 0.110 0.227 0.000 0.001 0.349

Herfindahl index 0.047 0.065 0.050 0.092 0.088 0.000 0.000 0.001 0.493

Cash 0.070 0.050 0.045 0.076 0.039 0.030 0.060 0.407 0.164

Free cash flows 0.048 0.041 0.050 0.029 0.046 0.082 0.045 0.242 0.248

Investment 0.027 0.025 0.024 0.022 0.030 0.215 0.241 0.141 0.401

Payout ratio 0.180 0.207 0.235 0.208 0.167 0.091 0.083 0.018 0.193

Tax 0.015 0.013 0.018 0.013 0.008 0.236 0.151 0.218 0.478

Number of obs. 1400 212 90 69 53

Note: This table shows the means and medians across non-PtP, PtP firms as well as private equity backed (PE), management

sponsored (MS) and other (Oth.) deals. The last four columns show p-values for a t-test for equal means allowing for unequal

variances in Panel 1 and a Mann-Whitney U-test for equal medians in Panel 2. All variables are trimmed at the 1st and 99th

percentiles, except for the ownership and illiquidity variables. See Table 1 for variable definitions.

33

Table 3: Multinomial Logistic Regression Analysis of Factors Influencing the Likelihood of Going Private Transactions

Private equity Management p-value for LR testVariable backed deals sponsored deals Other deals of equal parameters

coeff. s.e. coeff. s.e. coeff. s.e. PE-MS PE-Oth MS-OthMarket to book −0.293 (0.170)∗ −0.887 (0.243)∗∗∗ −0.512 (0.233)∗∗ 0.038 0.431 0.251

Rumours 0.414 (0.099)∗∗∗ 0.440 (0.138)∗∗∗ 0.321 (0.135)∗∗ 0.865 0.537 0.513

Sales growth 0.010 (0.360) 0.282 (0.368) −1.085 (0.695) 0.586 0.128 0.054

ROA 1.530 (1.046) 0.419 (0.779) −0.216 (0.806) 0.375 0.176 0.564

Analysts following 0.009 (0.050) −0.134 (0.070)∗ 0.027 (0.062) 0.079 0.814 0.071

Thin trading 0.268 (0.800) 4.713 (0.951)∗∗∗ 3.075 (1.000)∗∗∗ 0.000 0.026 0.213

Cash −2.324 (1.150)∗∗ 1.589 (0.854)∗ 1.445 (0.958) 0.004 0.009 0.906

Payout ratio 0.565 (0.293)∗ 0.172 (0.340) −0.585 (0.256)∗∗ 0.372 0.003 0.058

Total debt −1.034 (0.825) 0.718 (0.796) 1.400 (0.868) 0.113 0.038 0.543

Executive ownership 2.044 (0.805)∗∗ 2.676 (0.692)∗∗∗ −1.608 (1.393) 0.528 0.013 0.001

Financial inst. own. 3.247 (0.618)∗∗∗ −0.170 (0.812) 0.094 (0.830) 0.000 0.001 0.814

Size −0.017 (0.122) 0.301 (0.140)∗∗ 0.278 (0.146)∗ 0.078 0.114 0.908

Tax 5.668 (4.600) 12.060 (5.030)∗∗ 15.715 (4.743)∗∗∗ 0.330 0.114 0.573

St.dev. of returns 0.079 (0.033)∗∗ 0.045 (0.034) −0.012 (0.041) 0.463 0.074 0.260

No heterogeneity test 0.000 0.000 0.000 0.000 0.000 0.000

IIA test 1.000 1.000 1.000

Note: The table reports estimation results for the multinomial logistic regression model given in (1), using the non-PtP firms asreference group. The model is estimated using 212 observations for UK PtP deals over the period 1997-2003. All regressors areindustry and time adjusted as they enter the multinomial logit model as deviations from the industry and year median. The numbersof observations in the PtP groups are 90 for PE-backed deals, 69 for MS-deals, and 53 for other deals. The non-PtP group consists of1400 observations. Standard errors are given in parentheses, with ∗∗∗, ∗∗, and ∗ indicating significance at the 1%, 5% and 10% level,respectively. The final three columns show p-values for the LR test of equal parameters across two sub-groups of PtP deals. The line“No heterogeneity test” reports p-values for the LR test of Cramer and Ridder (1991) that all parameters except the intercepts areequal for two groups. The first three numbers in this line compare the non-PtP group with one of the PtP groups. The line “IIA test”reports p-values for the Hausman and McFadden (1984) LR test for the validity of the independence of irrelevant alternatives (IIA)assumption, omitting the indicated group from the model. Variable definitions are provided in Table 1.

34

Table 4: Multinomial Logistic Regression Analysis of Factors Influencing the Likelihood of Going Private Transactions

Private equity Management p-value for LR testVariable backed deals sponsored deals Other deals of equal parameters

coeff. s.e. coeff. s.e. coeff. s.e. PE-MS PE-Oth MS-OthMarket to book −0.339 (0.234) −0.664 (0.287)∗∗ −0.409 (0.329) 0.367 0.860 0.560

Rumours 0.411 (0.112)∗∗∗ −0.013 (0.307) 0.276 (0.170) 0.144 0.452 0.387

M/B × Rumours 0.051 (0.121) −1.021 (0.527)∗ 0.035 (0.207) 0.024 0.946 0.047

Director buying −0.199 (0.298) 0.266 (0.334) −0.320 (0.407) 0.288 0.805 0.254

Director selling 0.048 (0.360) −1.424 (0.755)∗ 0.342 (0.462) 0.047 0.610 0.028

Sales growth 0.373 (0.346) 0.369 (0.402) −1.019 (0.832) 0.992 0.074 0.092

ROA −0.323 (1.118) 0.263 (0.803) −0.608 (1.541) 0.661 0.880 0.626

Analysts following −0.007 (0.056) −0.164 (0.089)∗ 0.042 (0.076) 0.120 0.596 0.070

Thin trading 0.306 (0.904) 3.807 (1.167)∗∗∗ 2.083 (1.203)∗ 0.015 0.232 0.292

Excess Cash −1.249 (0.659)∗ 0.588 (0.318)∗ 0.686 (0.356)∗ 0.006 0.006 0.830

Normal Cash −0.103 (0.079) −0.012 (0.093) 0.116 (0.117) 0.444 0.104 0.378

FCF 0.335 (0.740) 0.050 (0.220) 2.377 (1.476) 0.688 0.200 0.050

Payout ratio 0.720 (0.302)∗∗ 0.080 (0.381) −0.628 (0.283)∗∗ 0.190 0.002 0.119

Total debt −3.122 (1.556)∗∗ 0.988 (1.215) −0.670 (1.610) 0.031 0.267 0.399

Executive ownership 2.638 (0.938)∗∗∗ 2.793 (0.824)∗∗∗ −3.380 (2.220) 0.897 0.003 0.001

Financial inst. own. 3.605 (0.694)∗∗∗ −0.472 (1.038) 0.717 (0.975) 0.001 0.012 0.394

Size 0.105 (0.136) 0.123 (0.170) 0.130 (0.187) 0.931 0.912 0.978

Tax 7.688 (5.242) 8.296 (6.200) −7.381 (7.501) 0.939 0.091 0.100

St.dev. of returns 0.080 (0.038)∗∗ 0.052 (0.041) 0.039 (0.046) 0.610 0.492 0.834

No heterogeneity test 0.000 0.000 0.005 0.000 0.000 0.000

IIA test 1.000 1.000 1.000

Note: The table reports estimation results for the multinomial logistic regression model given in (1), using the non-PtP firms asreference group. The numbers of observations drop to 70 for PE-backed deals, 49 for MS-deals, and 35 for other deals. The non-PtPgroup consists of 1283 observations.Standard errors are given in parentheses, with ∗∗∗, ∗∗, and ∗ indicating significance at the 1%, 5%and 10% level, respectively. See Table 3 for further details.

35

Table 5: Summary Statistics of Firms Grouped by Size

Mean by deciles of sizeThin Analysts

Decile Size PtP(%) PE(%) MS(%) trading following AIM1 (small) 4.17 5.0 2.5 2.5 0.79 1.77 0.12

2 12.36 11.7 5.6 4.9 0.72 2.06 0.15

3 21.22 13.7 3.1 5.6 0.69 1.96 0.08

4 34.78 16.7 7.4 4.9 0.66 2.27 0.11

5 53.43 20.4 8.6 8.6 0.59 2.86 0.07

6 84.44 16.8 6.8 6.8 0.59 3.27 0.13

7 142.15 14.2 5.6 4.9 0.54 3.20 0.12

8 266.93 17.4 7.5 3.1 0.46 3.56 0.15

9 731.44 14.2 8.0 1.9 0.29 4.40 0.13

10 (large) 6912.01 2.5 0.6 0.0 0.10 4.36 0.14

Total 828.11 13.2 5.6 4.3 0.54 2.97 0.12

Note: The table reports mean values of several variables across size deciles where size is measured bytotal assets. Variable definitions are provided in Table 1.

36

Table 6: Summary Statistics of Firms Grouped by Size

Mean by deciles of sizeThin Analysts

Decile Size trading following AIM

Panel A: Non-PtP firms

1 (small) 4.09 0.79 1.79 0.15

2 12.32 0.72 2.10 0.15

3 21.49 0.68 2.03 0.09

4 34.97 0.66 2.25 0.11

5 53.42 0.57 2.95 0.06

6 84.80 0.58 3.41 0.13

7 141.75 0.53 3.36 0.13

8 268.83 0.45 3.62 0.11

9 734.29 0.27 4.39 0.12

10 (large) 7020.87 0.10 4.35 0.13

Panel B: Private equity backed PtP deals

1 (small) 5.72 0.71 2.00 0.25

2 12.46 0.72 2.33 0.11

3 19.19 0.70 2.40 0.00

4 36.41 0.60 2.42 0.08

5 54.64 0.52 2.43 0.07

6 82.19 0.60 3.09 0.09

7 136.51 0.60 3.00 0.22

8 270.71 0.45 4.25 0.17

9 769.57 0.46 4.69 0.15

10 (large) 3281.60 0.25 4.00 0.00

Panel C: Management sponsored PtP deals

1 (small) 5.37 0.91 0.75 0.25

2 12.44 0.76 1.25 0.12

3 19.33 0.78 0.78 0.00

4 31.73 0.79 2.00 0.12

5 52.89 0.76 2.21 0.07

6 80.27 0.74 2.27 0.09

7 148.03 0.64 1.88 0.00

8 228.58 0.66 2.40 0.00

9 825.51 0.43 4.33 0.33

10 (large) −− −− −− −−

Note: The table reports mean values of thin trading, analysts followingand frequency of AIM listing across size deciles for the non-PtP firms(Panel A), private equity backed (Panel B), and management sponsoreddeals (Panel C). Size is measured by total assets. Variable definitions areprovided in Table 1. 37

.00

.01

.02

.03

.04

.05

.06

-4 -2 0 2 4 6 8 10 12

PE-backed

Management-sponsored

Other

(a) Market to book

.00

.05

.10

.15

.20

.25

.30

0 1 2 3 4 5 6 7 8 9 10 11 12

PE-backed

Management-sponsored

Other

(b) Rumours

.000

.004

.008

.012

.016

.020

-0.8 -0.4 0.0 0.4 0.8 1.2 1.6 2.0 2.4

PE-backed

Management-sponsored

Other

(c) Sales growth

.000

.004

.008

.012

.016

.020

-2.0 -1.6 -1.2 -0.8 -0.4 0.0 0.4

PE-backed

Management-sponsored

Other

(d) ROA

.000

.004

.008

.012

.016

.020

.024

-8 -6 -4 -2 0 2 4 6 8 10

PE-backed

Management-sponsored

Other

(e) Analysts following

.00

.04

.08

.12

.16

.20

-0.8 -0.6 -0.4 -0.2 0.0 0.2 0.4 0.6 0.8 1.0

PE-backed

Management-sponsored

Other

(f) Thin trading

Figure 1: Multinomial Logit probabilities

38

.000

.005

.010

.015

.020

.025

.030

-.3 -.2 -.1 .0 .1 .2 .3 .4 .5 .6 .7 .8

PE-backed

Management-sponsored

Other

(a) Cash

.000

.005

.010

.015

.020

.025

.030

-3.0 -2.5 -2.0 -1.5 -1.0 -0.5 0.0 0.5 1.0 1.5 2.0

PE-backed

Management-sponsored

Other

(b) Payout ratio

.00

.01

.02

.03

.04

.05

-.1 .0 .1 .2 .3 .4 .5 .6 .7 .8 .9

PE-backed

Management-sponsored

Other

(c) Exec.own

.00

.01

.02

.03

.04

.05

.06

.07

.08

.09

-.3 -.2 -.1 .0 .1 .2 .3 .4 .5 .6 .7 .8

PE-backed

Management-sponsored

Other

(d) Fin.inst.own

Figure 2: Multinomial Logit probabilities

39