The productivity effects of bank mergers: Evidence from the UK building societies

Journal of Banking & Finance 23 (1999) 825±846

The productivity e€ects of bank mergers: Evidence from the UK building societies Michelle Haynes a, Steve Thompson a

b,*

Department of Economics, University of Nottingham, Nottingham NG7 2RD, UK b Department of Economics, University of Leicester, Leicester LE1 7RH, UK Received 19 May 1998; accepted 24 July 1998

Abstract This paper presents an empirical investigation of the impact of acquisition activity on ®nancial intermediary productivity. Speci®cally, it uses an augmented production function approach to investigate the impact of acquisition, after controls for input changes. The model is estimated on an unbalanced panel of 93 UK building societies over the period 1981±1993, using data on their core ®nancial intermediation activities which, it is suggested are particularly appropriate for our purposes. In contrast to much of the existing merger literature, which for the most part uses ®nancial performance data, our results DO indicate signi®cant and substantial productivity gains following acquisition. These are consistent with an acquisitions process in which less ecient ®rms are acquired and reorganized. The post-merger gains appear to increase substantially in the post-deregulation period, when pressures to minimize cost are widely considered to have increased. Ó 1999 Elsevier Science B.V. All rights reserved. JEL classi®cation: G21; G34 Keywords: Mergers; Productivity e€ects; Building societies

* Corresponding author. Tel.: +44 116 252 2897; fax: +44 116 252 2908; e-mail: [email protected]

0378-4266/99/$ ± see front matter Ó 1999 Elsevier Science B.V. All rights reserved. PII: S 0 3 7 8 - 4 2 6 6 ( 9 8 ) 0 0 1 1 7 - 4

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1. Introduction In a variety of di€erent countries and regulatory environments, banking systems continue to display very high rates of consolidation via acquisitions and mergers. While this activity is usually justi®ed by the participants on ef®ciency grounds, and frequently endorsed by the relevant regulatory authorities for the same reasons, the scienti®c evidence on the consequences of completed takeovers is generally disappointing. The merger process may be successful ex ante in identifying underperforming targets and situations where there is the potential for substantial performance improvements (e.g. Berger and Humphrey, 1991; Fixler and Zieschang, 1993; Sha€er, 1993). However, the studies of completed mergers have generally failed to reveal ex post evidence of realised performance bene®ts (e.g. Rhoades, 1993; Berger and Humphrey, 1992; Vander Vennet, 1996). Indeed Vander Vennet's recent conclusion on European bank mergers seems an apt summary of this literature to date (Vander Vennet, 1996, p. 1553): ``Target banks exhibit an inferior performance but the acquirers are unable to remedy this situation. This type of takeover clearly fails to purge the system of inecient banks.'' The purpose of this paper is to present some ®ndings from a direct investigation of the impact of acquisition activity on ®nancial intermediary productivity. Speci®cally, we use an augmented production function approach, a methodology which has been widely employed to investigate the productive impact of a variety of organizational changes, to examine the impact of acquisition on output after controls for input changes. The dataset consists of a panel of 93 UK building societies over the period 1981±1993. The UK building society sector, which is very similar to that of the US mutual savings and loans, is particularly appropriate for our purpose in that it experienced a high level of (almost) exclusively horizontal (i.e. intra-sector) mergers over the period of investigation. Furthermore, for regulatory purposes each ®rm's core ®nancial intermediation activity produced separate accounts, allowing us to collect input and output data for a homogeneous activity across the sample. This advantage becomes particularly important in the latter part of the period when some of the societies developed a diversi®ed mix of ®nancial activities. In contrast to much of the previous literature, our results do show evidence of productivity gains in ®nancial intermediation following acquisitions. This result is robust to changes in speci®cation and to the functional form employed, although the magnitude of the gains does vary with such changes. In general, the pattern of coecients is suggestive of a steadily increasing e€ect over six or more years subsequent to an acquisition. In the full model this grew from 3.4% in the year immediately following the merger to 5.6% ®ve years

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later. When ®xed assets were dropped from the model, to counter fears of bias resulting from revaluation e€ects, the pattern of coecients suggested a similar gain rising from 3.4% to 5.5%. A protracted generation of productivity gains was not unexpected, given that mergers between mutuals are frequently accompanied by employment assurances to sta€. Unlike most previous studies, which have tended to use ®nancial performance or management expense ratios, the augmented production function approach employed here has the advantage that it uses variables corresponding to real output and real inputs in the production process. Therefore it avoids contamination from extraordinary expenditures associated with post-acquisition changes, e.g. compensation payments to factor suppliers whose inputs are no longer required. However, our results do suggest that the productivity effects of mergers may depend critically on the regulatory and/or competitive environment in which the institutions operate. Whilst degrees of freedom problems prevented a thorough investigation of the longer lagged e€ects of more recent mergers, the results are indicative of much larger gains in the deregulated and more competitive environment which prevailed from 1987 onwards. The remainder of the paper is organized as follows: Section 2 reviews the arguments relating to the productivity e€ects of mergers and summarizes the existing literature. The augmented production function is outlined in Section 3. Section 4 describes the UK building society sector and examines the sample characteristics. The results are presented in Section 5 and a brief conclusion follows. 2. Bank mergers: background and issues It has been widely observed that banking systems operating in a variety of competitive and regulatory environments have been experiencing similarly high rates of consolidation via mergers and acquisitions. 1 However, a substantial volume of academic literature has failed to ®nd widespread empirical support for an eciency-enhancing role for bank mergers. In part this conclusion may be unsurprising. Many bank mergers are ostensibly justi®ed by their participants in terms of economies of scale, but a large volume of empirical evidence has produced somewhat ambiguous ®ndings on the extent and persistence of these ± see below. Furthermore, the results from the banking studies follow a longstanding trend in the wider industrial organization literature, in which the

1

For example, a very similar pattern of intense acquisition activity exists in US commercial and savings banking (Peristiani, 1996), co-operative banking in Germany (Bonus and Schmidt, 1990) and building societies in the UK (Thompson, 1997).

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performance e€ects of completed mergers have generally proved disappointing ± see Caves (1989), for a survey, and Dickerson et al. (1997) for a recent example. This ®nding has often been interpreted as support for the view that mergers are frequently motivated by managerial preferences for expansion. If managerial preferences are ignored and the pro®tability e€ects, if any, consequent upon increased market power are set aside, 2 horizontal mergers between non-diversi®ed banks have the potential to impact on bank performance in one of four ways: ®rst, via economies of scale; second, via the selective redeployment of assets; third, via the transfer of assets to better quality managers; and fourth, via the renegotiation of implicit labor contracts. These may be considered in turn. 2.1. Economies of scale As noted above, the existence and importance of economies of scale has proved a controversial empirical issue ± see Berger and Humphrey (1997) for a comprehensive recent survey ± with diverse results being reported for di€erent samples, time periods and functional speci®cations. At some risk of oversimpli®cation, the following ``stylized facts'' may be considered to have emerged from the US work: ®rst, that economies of scale are found at the level of smaller banks, with some recent studies suggesting that these may persist up to much larger sizes, perhaps even for banks with up to $10 billion assets (Peristiani, 1996); second, that such economies cease or even become negative at the scale of very large banks (e.g. Hunter and Timme, 1995); third, that even within the output range where scale economies do exist their measured extent is small, of the order of ®ve percent or less (Berger and Humphrey, 1992); and fourth, that samples of banks typically display high (and variable) levels of technical or X-ineciency (e.g. Berger and Humphrey, 1991; Miller and Noulas, 1996). This is generally suciently large as to dominate scale e€ects: for example, Berger and Humphrey (1992) suggest that it averages 20% or more across samples of banks irrespective of their size classes. Work on banking systems outside the United States has produced a broadly similar mix of ®ndings. For example, the empirical evidence (see Section 5.1) from the UK building society sector, the subject of this paper, presents an ambiguous picture on the existence of scale economies but supports the view of the existence of widespread technical ineciency. In short, the evidence from the bank production literature appears to suggest that while mergers may have some limited potential to raise performance through scale economies, the realisation of these will depend on controlling technical ineciency.

2 The available evidence suggests that market power e€ects are likely to be small and dominated by any signi®cant changes in X-eciency (see Berger and Humphrey (1992) for a survey).

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2.2. Selective redeployment of assets In addition to any e€ects of scale per se, the merging of banks, particularly those with overlapping coverage, may give rise to economies where underutilization allows selective retirement of assets. Dutz (1989) showed how horizontal mergers could generate savings in the presence of surplus capacity and capital of di€erent vintages as output is reassigned to the more productive capital. In banking, both operating systems and branch oces may have vintage or, in the case of the branches, location-speci®c characteristics which may be comparable to those described by Dutz, such that similar gains may follow selective development or closure. 2.3. Transfer of asset control to better quality managers If the acquisitions process results in assets passing from the stewardship of less able or less diligent managers to the control of their more successful counterparts, as the market for corporate control hypothesis contends, X-ef®ciency gains would be expected to result. While the regulatory regime may attenuate the competitiveness of the market for corporate control in banking, particularly insofar as it restricts the set of potential acquirers, it does not invalidate the argument provided that target banks tend to be drawn from the set of relative underperformers. A result which is generally supported in the literature. Similarly, while mutual ®rms cannot bid for one another's equity, a mutual acquisition process in which the underperformers become targets, as Thompson (1987) has described for the UK building society sector, should generate a qualitatively similar potential for gain. 2.4. Renegotiation of implicit labor contracts Shleifer and Summers (1988) have argued that the new management of an acquired ®rm have an opportunity to renegotiate the implicit terms of the labor contract which existed between that ®rm's owners and workers. 3 As they point out, this may allow a ``breach of trust'' with respect to employees long-term job expectations, perhaps with serious systemic consequences. However, judged in isolation, such a renegotiation, particularly if it involves changes in e€ort or work scheduling levels, may have implications for the eciency of the ®rm's operation. 3

This is most obviously the case for hostile acquisitions, where the incoming senior management has a clear credibility advantage going into any potential confrontation with its newly-acquired sta€. However, any new managerial team may possess greater freedom to impose changes by virtue of its lack of previous association with the organization. Bhagat et al. (1990) provide general evidence that job losses follow hostile takeovers.

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The existing empirical and simulation evidence on the performance impact of bank mergers is very largely indicative of a process in which potential gains of the foregoing types fail to be realised. Studies on completed US mergers, using cost and performance variables, (e.g. Berger and Humphrey, 1992; Rhoades, 1993) generally con®rm that targets have lower eciency levels than their acquirers, but usually fail to ®nd any performance improvement following the acquisition (see also Srinivasan and Wall, 1992). Simulation studies, such as Sha€er (1992) and Fixler and Zieschang (1993), provide additional con®rmation. Sha€er, for example, estimates a translog cost function for 210 large US banks and then computes hypothetical cost savings for random bank pairings. His results are indicative of an insigni®cant and extremely small mean gain, with rather larger improvements for inter-state mergers and substantial negative cost e€ects for larger bank acquisitions. Vander Vennet's analysis of 422 domestic and 70 cross-border bank acquisitions in Europe (Vander Vennet, 1996) produces a very similar conclusion to the US studies. The principal exception to the above literature is a study by Cornett and Tehranian (1992) of 30 large US bank mergers occurring between 1982 and 1987. The authors do report statistically signi®cant increases in the operating income to assets employed ratio and, especially, in the return on book equity. However, Berger and Humphrey (1992) point out that this latter result is strongly in¯uenced by capital write-o€s which reduced the denominator and gave a perverse boost to apparent performance. 3. Productivity e€ects: The augmented production function approach The basic approach adopted in this paper is the estimation of a production function augmented by variables intended to capture the impact of completed acquisitions. The augmented production function has been widely used in the literature on the economics of the ®rm to investigate the impact of organizational factors including the adoption of ESOPs and bonus plans (Jones and Kato, 1995), pro®t sharing (Cable and Wilson, 1989; Wadhwani and Wall, 1990; Kruse, 1992) and the introduction of worker participation (Defourney et al. 1985). The underlying rationale for such an approach ± see Nickell et al. (1996) for some discussion of these issues ± is that ®rm production e€ectively requires an additional and largely unobservable factor input in the form of managerial/organizational skills. Change in internal organizational arrangements, including improvements in incentives and control systems, and such changes as might be implemented following an acquisition may be considered to impact on the quality of this unobservable factor, manifesting themselves in a shift in the observed production function. This e€ect is typically investigated using dummy variables to isolate the contemporaneous and lagged e€ect of the change in question.

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Since the primary objective of the approach is to isolate the productive consequences of a class of events or characteristics, rather than provide a detailed analysis of a particular productive process, most studies use a simple generalized form of the production function. Cobb±Douglas and translogarithmic versions dominate. Many studies, including the present one, use single industry samples where it is realistic to assume reasonably high degrees of input and output homogeneity. However, others utilize much more heterogeneous samples: for example, Jones and Kato (1995) have a sample drawn from all Japanese manufacturing industries; whilst Nickell (1996) argues for the generality of a basic functional form. 4 Our paper uses a sample of ®rms which are unusually homogenous in that the societies were restricted, by regulation, to core ®nancial intermediation activities alone until 1987, and thereafter our data relate only to these activities and exclude any non-core diversi®cation which were separately recorded in the groupÕs accounts. This paper follows Murray and WhiteÕs approach to the bank production function (Murray and White, 1980), and uses a generalized Cobb±Douglas form with labor and capital inputs. Since, in accordance with the augmented production function literature, we wished to use a single output measure, rather than a vector of outputs as control variables as typically employed in cost function studies, we used the level of commercial (i.e. earning) assets. 5 Thus the basic model, prior to its augmentation with merger variables takes the form b2 b3 b4TIME ; Qit ˆ Ait Lb1 it K1it K2it e

…1†

where Qit is the output of ®rm i at time t, Lit is the labor input of ®rm i at time t, K1it is the level of ®xed assets of ®rm i at time t, K2it the level of liquid assets (cash and deposits) of ®rm i at time t, and eb4TIME an exponential time trend to capture Hicksian neutral technological progress.

4 Nickell has argued strongly for simplicity in the speci®cation of production functions in empirical work. Indeed, in Nickell (1996) he imposes constant returns Cobb±Douglas technology and has defended this (at a National Institute for Economic and Social Research seminar, May 1997) as a reasonable characterization of the population of ®rms. 5 Sealey and Lindley (1977) provide an eloquent justi®cation for a bankÕs output being proxied by a stock of earning assets rather than a ¯ow of services as the production function concept might be considered to imply. They point out that it is the bankerÕs function to maintain the existence of this stock. They use PesekÕs well-known analogy (Pesek, 1970, pp. 1254±1255) that the bankÕs stock of assets is not comparable to a stock of Rembrandts but rather with the river, constantly changing but with the banker in control of the sluice gates. The dependent variable in this paper is the societyÕs stock of commercial assets. Over the period of investigation, 1981±1993, this is overwhelmingly dominated by mortgages and other property/land-backed lending. After 1987 there was a strictly limited freedom to expand Class 3 commercial assets, including investments in new activities ± see Drake (1989) for details ± but substantial investments in such activities, estate agency, etc., were transferred to subsidiaries and do not ®gure in our data which are derived for the societiesÕ core accounts.

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In specifying appropriate merger terms it was necessary to address two di€erences between bank merger activity and other activities which have been examined using an augmented production function framework: First, there is no clear indication in the existing literature on the choice of an appropriate lag between acquisition activity and performance change, if any. For example, Rhoades (1993) uses a performance comparison which begins four years after the completion of the acquisition whilst the approach of Fixler and Zieschang (1993) uses immediate post-merger data. Second, completing an acquisition is not an o€/on experience as is, for example, the introduction of pro®t sharing or employee participation in the production function literature, nor is it necessarily an unusual experience for ®rms in the sample and it may be repeated on several occasions. In acknowledgement of these points two di€erent dummy variable approaches were employed: First, separate dummy variables were de®ned for the ®rm's ®rst year of operation subsequent to any intra-sector acquisition (MERGERit ) and then for each of the following ®ve years (MERGER1it , MERGER2it ,. . ., MERGER5it ) to capture lagged e€ects. This approach has the advantage of allowing for multiple acquisitions via overlapping sets of dummy variables. The second approach followed a recent paper on manufacturing mergers by Dickerson et al. (1997) in specifying a single post-merger binary variable (ANYMERGEit ) equal to one if the ®rm i had engaged in any merger activity over the period. While this approach does not distinguish the e€ects of multiple acquisitions it has the advantage of not imposing any prior lag structure on the anticipated productivity changes. After taking logs and augmenting with merger variables, ®rm speci®c ®xed e€ects (ai ) and a stochastic error term (uit ), Eq. (1) becomes Ln Qit ˆ ai ‡ b1 Ln Lit ‡ b2 Ln K1it ‡ b3 Ln K2it ‡ b4 TIME ‡ b5 MERGERit ‡ b6 MERGER1it ‡


‡ b10 MERGER5it ‡ uit : …2† Our expectation was that if productivity rose following an acquisition this would result largely from reduced X-ineciency, as assets were transferred to new managers and as under-utilized and/or ageing assets were retired, etc. Therefore it was expected that the gains, if any, would be captured in the lagged merger terms. Whilst the literature suggested that scale e€ects may not be substantial in this sector, it was anticipated that any bene®ts derived from increased scale would reinforce the X-eciency gains. However, the expectation for the initial e€ects of acquisition was ambiguous. If a more ecient society acquired a less ecient one the immediate (i.e. post-acquisition but prereorganization) result may be a reduction in the acquirer's eciency by a simple process of averaging.

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4. UK building societies: Sector and sample characteristics 4.1. Sector characteristics In their origins the UK building societies were very similar to the US savings and loans, with a traditional specialization in deposit collection and mortgage lending. They remained very largely restricted to these core activities until 1987, since when the larger societies have enjoyed a limited, but gradually increasing, freedom to diversify into other ®nancial product markets (Drake, 1989; Ingham and Thompson, 1995). Financial deregulation also permitted other ®nancial institutions, including commercial banks, to make a large-scale entry into the societies' formerly exclusive preserve of mortgage lending. Since 1987, the societies have also enjoyed the option of demutualizing their ownership structure and switching to the (generally less restrictive) regulatory regime of the commercial banks. This option has been widely exercised since 1995, but in the period of our study only one case occurred. In recent decades the building society sector has experienced substantial attrition, which until recently occurred exclusively through intra-sector merger activity. The number of independent societies has fallen by 75% in the past twenty years. For the most part these takeovers have been ``friendly'', in the sense of being endorsed by the target's board of directors (see Thompson, 1996) and not infrequently have been initiated by the target itself. A minority of acquisitions have been instigated by the sector's regulator, the Building Societies' Commissioner, as a means of securing savers' deposits when a society faced ®nancial instability. It is clear that the process of acquisition among mutually owned ®rms, with no marketable equity claims, is di€erent from that involving stock ®rms. However, the identi®able characteristics which distinguish targets from other societies, and these include smaller size, lower recent growth, ®nancial weakness and negative pro®tability (see Thompson, 1997), appear very similar to those reported in the empirical literature on stock acquisitions. 6 As with bank consolidations elsewhere, building society mergers are typically justi®ed by their architects in terms of economies of scale (Barnes, 1984). However, here as elsewhere in the banking world the evidence is more ambiguous. Some early studies (e.g. Gough, 1979) reported no evidence that the managerial expense ratio fell with size and concluded that mergers were principally motivated by managerial preferences for growth etc. More recent studies, using more sophisticated statistical designs, have found some evidence of scale economies, although there is considerable disagreement about their

6 Palepu (1986) provides a comprehensive review of the general literature on merger target prediction.

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extent and persistence. Hardwick (1989) in a single output translog cost function study reported economies up to £280m assets. In a later multi-output study, to a similar design, he (Hardwick, 1990) raised the threshold to £1500m. In contrast, Drake (1992, 1995) found, at best, economies of scale only for small societies. Work by Drake and Weyman-Jones (1992) and others has suggested that there may be substantial levels of technical or X-ineciency within the sector, probably sucient to dominate any scale eciency e€ects. Field (1990), for example, found 86% of his sample societies to exhibit technical ineciency and he concluded that eciency di€erences owed much more to managerial incompetence than to scale e€ects. What distinguishes mutual from stock mergers is the absence of a market in corporate control. Mutuals, including UK building societies may be ``owned'' by their members, at least in the sense that the latter have a legal claim on their net worth in the event of liquidation, but in most other respects the ownership claim is severely attenuated. In principle, new members may join on equal terms to existing ones, thus establishing new claims on net worth. Thus it is infeasible to establish a secondary market in equity claims through which a would-be acquirer could purchase control. Furthermore, in the building societies' case the opportunity for near costless exit on demand by dissatis®ed depositors, 7 together with a one-person-one-vote decision rule, might be expected to thwart e€ective attempts either to monitor management or to build up substantial voting blocks (Thompson, 1996). Moreover, the legislation covering society governance gives the incumbent management team considerable discretion in deciding whether or not to support a takeover approach. Since an unsupported takeover would not normally reach the stage of a membership ballot, nearly all sector acquisitions are ``friendly''. This carries the corollary that potential eciency gains may be bargained away in an ex ante bid to secure the backing of the incumbent management and its sta€. Barnes (1984) cites examples of employment guarantees being given to secure assent in cases where the merging parties had overlapping branch networks. It has been noted above that there was substantial deregulation of the sector in 1987, even if some of the new freedoms did not apply with immediate e€ect (see Drake, 1989). More generally, however, the sector also belonged to a ®nancial services industry which was itself experiencing substantial regulatory and structural change over the period of our analysis. There is a widespread perception that the overall e€ects of such change were to increase competitive pressures on the sector, certainly over the period of our investigation. If so, this is likely to have impacted on both the mean level of technical ineciency across our sample and its dispersion over our period of investigation. Some con®r-

7 Fama and Jensen (1983) have analyzed how low cost exit may dominate voice in the governance of mutuals.

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mation of this is provided by Drake and Weyman-Jones (1996), whose estimates using post-deregulation data suggest rather smaller levels of technical ineciency than previously found, smaller indeed than those normally encountered in studies of US banks. 4.2. Sample characteristics This paper uses a data set covering the years 1981±1993, inclusive, over which time the population of the building society sector fell from 200 to approximately 80. In every case but one this attrition was caused by an intrasector acquisition. To be included in the sample we required: ®rst, that at least ®ve years of continuous data were available; and second, that at least two years of observations lay in the post-deregulation period (i.e. after 1986). In the event we ®nished with an unbalanced panel (i.e. one with di€erent numbers of observations per cross-sectional element) of 93 societies. This number is very close to the population in the latter part of the period. The balance of the panel is shown in Table 1. The sample of 93 ®rms recorded 79 positive merger-year observations, 47 of which were in the pre-1987 sub-period and 31 in the subsequent post-deregulation sub-period. The actual number of acquisitions exceeded this ®gure since multiple acquisitions in any one year were treated as single mergers. Data was collected from the societies' annual reports and accounts using the Building Societies Yearbook (various years), accounts held at the libraries of London, Manchester and Warwick Business Schools and summaries produced by the Building Societies' Association. Comparability of data across societies was straightforward since published accounts are standardized for regulatory purposes. Until 1987, all societies were e€ectively restricted to the same core activities. Since then, separate society and group accounts have been required

Table 1 Balance of panel No. of years

No. of companies

5 6 7 8 9 10 11 12 13

2 3 4 5 3 10 14 18 34

Total

93

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to distinguish the core and peripheral activities. The data below relate solely to the former. The variables used in the analysis were as follows: 8 Output (Q) was the book value of the commercial assets (loans and investments) of society i at time t, as recorded in the annual accounts, expressed in 1985 prices. Labor (L) was the number of full-time equivalent employees of society i at time t. Each part-time employee is treated as half a full-timer. Fixed Assets (K1) was the book value of society i's ®xed assets at time t, as recorded in the annual accounts, expressed in 1985 prices. Liquid Assets (K2) was the total liquid assets (cash and deposits with other institutions) as given in the society's accounts, expressed in 1985 prices. MERGER was a binary variable equal to one if year t represented the ®rst year of society i's accounting data subsequent to an intra-sector acquisition and zero otherwise. MERGER1, MERGER2,. . ., MERGER5 represent binary variables equal to one for observations one, two,. . ., ®ve years subsequent to positive values for MERGER for society i, and zero otherwise. ANYMERGER was a binary variable equal to one for all observations subsequent to an intra-sector merger for society i, and zero otherwise. The summary statistics for the continuous variables are given in Table 2.

5. Productivity e€ects: Empirical results 5.1. Basic results Production functions, augmented with a time trend, ®rm-speci®c ®xed effects and merger variables, were estimated by OLS using our panel of 93 societies across the years 1981±1993, involving a total of 1034 observations. Estimations were carried out for both Cobb±Douglas and translog functional forms. In the event, each speci®cation yielded a similar pattern of results and comparable marginal productivities. However, since F-tests on the inclusion of the higher order and cross product terms of the translog [(Ln L)2 , (Ln K1)2 ,

8 Cost function studies of bank eciency typically use deposits as an input variable. However, this seemed inappropriate in this production function context: ®rst, the paper uses balance sheet data with assets as an output variable. To include deposits (i.e. the principal element of liabilities) as an input variable runs the clear danger of turning the production function into an identity. Second, deposits are not an homogeneous variable in this context: over the period of our investigation some societies begin to make extensive use of wholesale funds and high denomination postal deposits.

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Table 2 Full period characteristics for continuous variablesa

a

Variable

Mean

S.D.

n

Output Employment Fixed assets (£m) Liquid assets (£m)

1152.57 668.36 25.31 258.76

3160.40 1727.61 97.81 704.72

1034 1034 1034 1034

Number of ®rms ˆ 93; number of mergers ˆ 79; (pre-1987 ˆ 47, post-1987 ˆ 31).

etc.] did not generally support their incorporation, 9 the Cobb±Douglas results are reported for convenience. These results, when estimated for the unrestricted Cobb±Douglas form, appeared to be consistent with decreasing returns across the sector. Since panel estimation has been shown to produce a possible downward bias in production function coecients, 10 the Cobb±Douglas model was also re-estimated after the imposition of constant returns to scale. The unrestricted Cobb±Douglas estimates are presented in Table 3. It can be seen that the overall ®t of the production function is highly satisfactory, with adjusted R-squared exceeding 0.99. The coecients on labor (Ln L) and liquid assets (Ln K2) are well-determined and the positive coecient on the time trend (TIME) is consistent with an annual productivity growth of between 4% and 4.5%. The only unsatisfactory coecient on an input variable was that on ®xed assets (Ln K1). This was not entirely surprising as asset revaluations within the sector appear to have been somewhat arbitrary, particularly in the earlier part of the period. There was some concern that such revaluations may have been stimulated by merger completion, leading to some endogeneity in Ln K1. Having regard for this, and the low overall contribution of this variable, the unrestricted version of the Cobb±Douglas was estimated with and without this variable. Turning to the merger variables, it is clear that acquisition appears to be associated with subsequent productivity gains. In the full augmented model, in Table 3, there is a clear pattern of signi®cant positive coecients, of gradually increasing magnitude, across the set of binary variables MERGER, MERGER1, MERGER2,. . ., MERGER5, covering observations from one to six years after acquisitions. These suggest a productivity e€ect which rises from approximately 3% in the ®rst year after an acquisition to approximately 5.5% ®ve years later. When the merger e€ect is modeled as a once and for all change, ANYMERGE, a rather larger coecient equivalent to approximately 13% is 9 F-tests on the joint signi®cance of these higher order terms did not support their inclusion. However, when such models were estimated with full augmentation they provided con®rmation of the signi®cant productivity e€ects ± results are available with the authors. 10 This point is discussed in Jones and Kato (1995, p. 399).

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Table 3 Panel estimates of augmented Cobb±Douglas production function [Dependent variable ˆ Ln Q] Ln L Ln K1 Ln K2 TIME MERGER MERGER1 MERGER2 MERGER3 MERGER4 MERGER5 FIRM DUMMIES ANYMERGER RMERGER RMERGER1 RMERGER2 RMERGER3 RMERGER4 RMERGER5 DMERGER DMERGER1 DMERGER2 DMERGER3 DMERGER4 DMERGER5

0.429 (14.14) 0.011 (1.16) 0.348 (16.01) 0.045 (21.68) 0.034 (1.902) 0.033 (1.83) 0.036 (2.06) 0.045 (2.53) 0.048 (2.53) 0.056 (2.92) Yes

0.439 (15.02) 0.350 (16.17) 0.045 (22.15) 0.034 (1.91) 0.033 (1.87) 0.037 (2.08) 0.044 (2.49) 0.048 (2.53) 0.055 (2.89) Yes

0.431 (14.38) 0.008 (0.88) 0.337 (15.53) 0.043 (21.49)

0.438 (15.19) 0.339 (15.65) 0.043 (21.93)

Yes 0.137 (6.76)

Yes 0.138 (6.79)

0.418 (13.84) 0.002 (0.25) 0.325 (14.88) 0.045 (22.07)

Yes ÿ0.009 (0.40) 0.021 (0.98) 0.022 (1.05) 0.033 (1.58) 0.025 (1.14) 0.034 (1.54) 0.110 (4.15) 0.095 (3.28) 0.106 (3.27) 0.090 (2.63) 0.058 (1.45) 0.042 (0.97)

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839

Table 3 (Continued) Adjusted R2 F [d f] Observations

0.996 2785.6 [102,931]

0.996 2812.2 [101,932]

0.996 2985.0 [97,936]

0.996 3016.8 [96,937]

0.996 2716.2 [108,925]

1034

1034

1034

1034

1034

obtained. Dropping the ®xed assets input (Ln K1) from the estimation, in recognition of the limitations of this variable, had very little impact on the overall ®t of the equation or on the values of the post-acquisition dummy coecients. It is clear that the sector as a whole underwent substantial change to its regulatory and competitive environment over the period of our investigation, especially after partial deregulation from January 1987. Regulatory prohibitions on entry were removed for a whole range of ®nancial products; the societies were accorded the option of demutualizing to become commercial banks; whilst other ®nancial institutions made large-scale entries into the societies' core business of providing mortgage loans. It has been widely conjectured (e.g. by Drake, 1989) that such changes will have had the e€ect of increasing competitive pressures in the sector. This, in turn, might be expected to impact on the merger process and encourage acquirers to be more diligent in implementing cost savings. Furthermore, recent evidence on US bank mergers, presented by Berger (1998), does suggest changing eciency consequences over the last decade or so. To investigate possible changes in their productivity effects over the period we distinguished mergers before and after January 1987. We then generated two sets of dummy variables: the ®rst, RMERGER, RMERGER1,. . ., RMERGER5, represents the pre-1987 regulated period mergers and the second, DMERGER, DMERGER1,. . ., DMERGER5, represents those in the deregulated period. Of course, since our data ended in 1993, there were very few non-zero observations on the longer lags of the latter set of mergers. In the event the pattern of coecients on the two sets of variables gave clear support for our conjecture of a regime change coinciding with deregulation. The pre-1987 coecients, although invariably positive for RMERGER1 and subsequent observations, were small and generally insigni®cant. By contrast, the post-1987 binary variables DMERGER through DMERGER3 were signi®cant, with productivity gains between 11% and 9%. The ®nal two lags, DMERGER4 and DMERGER5, were insigni®cant, but as noted above there were very few positive cases here because of the proximity to the end of the dataset. The imposition of constant returns to scale produced a less satisfactory estimation, but provided broad con®rmation of the positive productivity e€ects of mergers. In Table 4 the full model is re-estimated after dividing the levels of

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Table 4 Panel estimates of augmented constant returns Cobb±Douglas production function [Dependent variable ˆ Ln (Q/L)] Ln(K1/L) Ln(K2/L) TIME MERGER MERGER1 MERGER2 MERGER3 MERGER4 MERGER5 ANYMERGER FIRM DUMMIES Adjusted R2 F [d f]

0.008 0.400 0.032 0.022 0.020 0.026 0.034 0.034 0.041

(0.85) (18.74) (25.29) (1.16) (1.12) (1.41) (1.87) (1.77) (2.06)

0.006 (0.64) 0.394 (18.53) 0.030 (23.39)

Yes 0.86 64.1 [101,932]

0.097 (4.76) Yes 0.86 68.35 [96,937]

Observations

1034

1034

output and input variables by labor, before taking logarithms. It can be seen that imposing constant returns worsens the overall ®t quite substantially. However, the positive coecients increase over time and reach signi®cant levels for MERGER3 and later dummies. Replacing these with a single post-merger shift variable again produces a large, signi®cant, positive e€ect, equivalent to a productivity gain of over 9%. Since our sample, which is an itself largely exhaustive of the relevant population, declines sharply over the period of investigation, there is a clear need to be alert to the possibility of ``survivor bias''. This is especially the case since all but one of the removals from the sample occur through mergers and acquisitions such that the merger process itself may be considered the industryÕs means of eliminating underperformers. However, while this may have the e€ect of raising the average productivity across the (surviving) sample there does not appear to be any reason why it should di€erentially impact upon the acquisition variables. Instead it might be anticipated that the rising average would be captured in the time trend as productivity growth across the (surviving) sample. The very strong performance of the TIME variables in our estimations would be consistent with this. 5.2. Additional experiments with the data At the outset it had been conjectured that the productivity gains associated with merger activity would be positively related to the degree of overlap between the two merging societies. In particular, it was considered that intraregional mergers would involve a greater consolidation of activities and head oces than would be feasible for more geographically separated mergers. To

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841

Table 5 Augmented Cobb±Douglas production function: Alternative speci®cations [Dependent variable ˆ Ln Q] Ln L Ln K1 Ln K2 TIME LARGETIM SMALLTIM REGTIME LMERGER LMERGER1 LMERGER2 LMERGER3 LMERGER4 LMERGER5 NMERGER NMERGER1 NMERGER2 NMERGER3 NMERGER4 NMERGER5 MERGER MERGER1 MERGER2 MERGER3 MERGER4 MERGER5 FIRM DUMMIES Adjusted R2 F [d f]

0.436 (15.30) 0.009 (1.06) 0.256 (11.63)

0.044 (2.57) 0.043 (2.66) 0.043 (2.62) 0.045 (2.71) 0.047 (2.66) 0.044 (2.44) Yes 0.997 3134.1 [103,930]

Yes 0.996 2638.8 [108,925]

0.035 (1.91) 0.033 (1.89) 0.037 (2.07) 0.045 (2.54) 0.048 (2.55) 0.056 (2.92) Yes 0.996 2755.9 [103,930]

Observations

1034

1034

1034

0.063 (24.91) 0.037 (18.18)

0.437 0.008 0.344 0.044

(14.34) (0.93) (15.78) (21.48)

0.013 0.005 0.015 0.023 0.047 0.041 0.055 0.048 0.050 0.064 0.044 0.068

(0.53) (0.22) (0.66) (1.03) (1.96) (1.63) (2.29) (2.12) (2.12) (2.65) (1.71) (2.59)

0.429 0.011 0.347 0.044

(14.13) (1.16) (15.98) (12.21)

0.001 (0.32)

investigate this we distinguished those mergers involving participants whose headquarters were less than 40 miles apart (LMERGER) from the remainder (NMERGER) and in each case a set of lagged variables (LMERGER1, LMERGER2 etc.) was generated. 11 The two sets of binary variables were substituted for the single set of MERGER dummies in Eq. (2). The results, in Table 5, did not accord with prior expectations. The local merger appeared to have no statistically signi®cant positive e€ect, at least until lags four and ®ve when a weak (10% level) impact was discernible. By contrast, the non-local merger e€ect was much larger and more rapidly observed. This result appeared

11

Several alternative de®nitions of ``regional'' mergers were tried using di€erent distances (20, 30, 50 miles etc.) , but these were not appreciably di€erent to the 40 mile de®nition illustrated in Table 5.

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surprising. However, it was noted earlier that mutual mergers e€ectively require the approval of both boards and, as such, are all ``friendly''. It is likely that obtaining such an agreement has often involved giving employment guarantees and promises of non-closure of branches, at least in the medium term (Barnes, 1984). By contrast, the set of non-local mergers is likely to include a disproportionate number of ``rescue'' cases, where the regulator has required one of the largest societies to acquire an ailing or illiquid rival to preclude any possibility of failure. Such acquisitions, which are as close to being ``hostile'' as the mutual sector experiences, appear more likely to be followed by major rationalization. As noted above, the augmented production function approach implicitly assumes that ®rms di€er in some unmeasurable dimensions of inputs (``management quality'' etc.) such that some observed organizational event, in this case engaging in an acquisition, moves observed output closer to its frontier. However, the consequences of such ®rm-level heterogeneities are unlikely to be restricted to merger e€ects. The research design adopted here makes some allowance for this by the inclusion of ®rm-level ®xed e€ects which will capture the constant inter-®rm di€erences in productivity levels. It is also possible that ®rm heterogeneities would lead to enduring di€erences in productivity growth rates. To investigate this, the sample societies were partitioned according to their 1987 asset levels into LARGE and SMALL societies and a separate time trend was estimated for each (LARGETIM and SMALLTIM). The results in Table 5 con®rms that the larger societies enjoyed a substantially greater growth rate in productivity across the period as a whole, an outcome consistent with enduring intra-sample heterogeneities. However, it can be seen that controlling for this e€ect leaves our basic merger results una€ected. In view of the importance of the deregulation e€ect revealed in Table 3, it appeared important to ensure that the MERGER e€ects were not swamped by any sector-wide post-deregulation productivity trends. Accordingly, a separate post-deregulation time trend variable (REGTIME) was introduced into Eq. (2) to test for any sector wide acceleration in productivity growth. In the event, the coecient was positive but insigni®cant (see Table 5) and the underlying MERGER variablesÕ performance was very largely una€ected. Our basic empirical design, following Eq. (2), assumed that the productivity e€ects following an acquisition occur through a shift in the production function. It is, of course, possible that acquisition directly impacts upon the production function changing its curvature or altering the scale parameter. To investigate this possibility we further augmented Eq. (2) with ANYMERGER*factor input interaction terms. This produced strictly limited success. An F-test on the joint inclusion of the three interaction terms was weakly signi®cant, but their inclusion inevitably injected an element of multicollinearity (inevitably given the correlations of Ln L and ANYMERGER* Ln L, ANYMERGER etc.) There was no change in the scale parameter and the ANY-

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843

Table 6 Panel estimates of augmented CES production function [Dependent variable ˆ Ln Q] Ln L Ln K2 (Ln L ± Ln K2)2 TIME MERGER MERGER1 MERGER2 MERGER3 MERGER4 MERGER5 ANYMERGER FIRM DUMMIES Adjusted R2 F [d f]

0.375 0.415 0.031 0.045 0.034 0.032 0.037 0.044 0.047 0.055

(5.263) (5.951) (0.985) (22.162) (1.883) (1.840) (2.087) (2.455) (2.492) (2.856)

0.373 0.404 0.031 0.043

(5.303) (5.854) (1.007) (21.951)

Yes 0.996 2784.52 [102, 931]

0.137 (6.723) Yes 0.996 2985.70 [97, 936]

Observations

1034

1034

MERGER shift e€ect remained large. However, there was some evidence of a small reduction in labor productivity and a corresponding small increase in capital productivity. 12 As a further check on the robustness of the results, a generalized CES production function, augmented with merger variables, was employed as an alternative to the Cobb±Douglas function. This was estimated, here without the insigni®cant K1 input, and the results are given in Table 6. These results overwhelmingly con®rm those of the Cobb±Douglas version in Table 3. Taken together, our results suggest that merger activity in the sector was associated with signi®cant productivity gains which were not the result of any detected economy of scale factor across the sample. However, since there does appear to have been considerable heterogeneity across the sample, the results are consistent with a merger process in which assets are transferred to the control of more productive managements. That is, the acquiring societies within the sample may possess some di€erence in intrinsic characteristics when compared to the set of non-acquirers. 13 However, such

12 Speci®cally, the interaction with Ln L produced a fall of ÿ0.09 (t ˆ 1.72), while that with Ln K2 was 0.10 (t ˆ 2.15). The impact on the other coecients was small. These results are available from the authors. 13 For example, if we compare the crude labor productivity of the set of acquirers with that of non-acquirers the former demonstrate signi®cantly higher levels, at least from 1988 onwards. The real output per full time equivalent employee, with numbers in round parentheses and t-ratios on the mean di€erence in square ones, with acquirers ®rst, are as follows: 1988 1.55 (29), 1.38 (62), [1.47]; 1989 1.59 (29), 1.35 (58), [2.20]; 1990 1.69 (29), 1.38 (56), [2.39]; 1991 1.74 (28), 1.39 (50), [2.57]; 1992 1.81 (28), 1.36 (50), [3.70]; and 1993 1.87 (26) 1.42 (45), [3.40].

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underlying explanations go beyond the scope of this paper whose design merely suggests a merger process which facilitates productivity improvements.

6. Conclusion This paper has used an augmented production function approach to explore the output e€ects of completed acquisitions by ®nancial intermediaries using a panel of UK building societies, over the period 1981±1993. It has been seen that the results are suggestive of signi®cant productivity gains, apparently rising over a period of six years or more subsequent to the acquisition itself. While the magnitude of the productivity e€ects is somewhat sensitive to the speci®cation, the overall results are robust to alternative functional form representations of bank production. The results also suggest that the gains from merger are not uniform across the period, but have increased under the deregulated and more competitive industry conditions that have prevailed since 1987. This conclusion is not unexpected because, since 1987, the UK building societies have been in direct competition with for-pro®t rivals and have had the option of demutualizing to become commercial banks. Such changes might be expected to shift a mutual's behavior in the direction of cost minimization. Furthermore, it parallels BergerÕs recent ®ndings in the US (Berger, 1998), where bank mergers appear to have become signi®cantly more productivityenhancing over the past decade. Unlike some previous studies, particularly those which have used ®nancial performance data, the production function approach presented here is not sensitive to any extraordinary expenditures associated with post-merger reorganization. Furthermore, to the extent that it directly addresses an aspect of the controversial issue of ®nancial services sector productivity, it would appear to o€er a clearer policy perspective than research which uses ®nancial performance alone.

Acknowledgements The authors are indebted to two referees and Sourafel Girma for many helpful comments on earlier versions of this paper. The assistance of the Building SocietiesÕ Association and Franey and Company in the collection of the data and ®nancial support from the Leverhulme Trust is acknowledged with gratitude.

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845

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