fficiency and market power in Latin American banking
onathan Williams ∗
angor Business School, Bangor University, Gwynedd, LL57 2DG, UK
r t i c l e i n f o
rticle history:vailable online 22 May 2012
EL classification:21342823
a b s t r a c t
I examine the relationship between bank efficiency and market power to test the quiet life hypothesisfor a sample of 419 Latin American commercial banks between 1985 and 2010. A two-stage least squaresmodel with instrumental variables controls for the simultaneous relationship between efficiency andmarket power. Citing developments in efficiency modelling, efficiencies are drawn from the randomparameters stochastic frontier function model that treats unobserved heterogeneity, whilst conventionalLerner indices and the efficiency adjusted Lerner index proxy market power. The quiet life hypothesisis firmly rejected after various robustness checks. To test if bank restructuring and governance changes
eywords:ank efficiencyarket power
ompetitiontochastic frontiers
affect efficiency and market power, I use a difference-in-differences approach to determine the impactof bank privatisation and foreign acquisition of local banks on efficiency and market power. Privatisationis preferred over foreign acquisition though its impact is concentrated on efficiency rather than marketpower. The evidence suggests that bank restructuring has promoted competition at the expense of marketpower and yielded efficiency gains at banks under conditions of monopolistic competition.
dlt2
mcttcmWiMclHt
overnance
. Introduction
The study explores the relationship between bank efficiencynd market power in Latin America. Market power measures aank’s ability to set prices above marginal cost with lower (higher)
evels of market power indicating a higher (lower) level of compe-ition in banking markets. Across the region, national governmentsave implemented liberal financial sector policies in expectationhat removing impediments to competition will yield efficiencyains (Carvalho et al., 2009). Yet, incentives for bank manage-ent to improve efficiency may be offset by the prospect of a
uiet life. According to the quiet life hypothesis, under monopolis-ic conditions bank management will forgo rents for inefficienciesn the allocation of resources (Hicks, 1935). These “inefficiencies”an manifest in the form of expense preference behaviour, these of lobbying to delay or prevent implementation of competi-ive policies, and choice of level of risk that is inconsistent withrofit maximisation (Berger and Hannan, 1998; Hughes et al.,
003). A quiet life can yield unfavourable outcomes: weak and
nefficient management may remain in post; banks may adopt non-ompetitive pricing leading to losses in social welfare (Maudos and
e Guevara, 2007);1 since market power is associated with higherevels of market concentration, it can limit financial deepening andhe development of more efficient banking sectors (Rojas-Suarez,007).
Financial sector developments influence bank efficiency andarket power. In Latin America, the mid 1990s banking sector
rises were resolved by government intervention and a consolida-ion process that accelerated market concentration to levels higherhan in other emerging markets (Domanski, 2005). Latin Ameri-an banking sectors are reported to operate under conditions ofonopolistic competition (IMF, 2001; Gelos and Roldós, 2004).hilst studies find that the level of monopolistic competition has
ncreased, this increase has not weakened competition (Yeyati andicco, 2007; Yildirim and Philippatos, 2007). Nevertheless, greater
oncentration can stifle competition because concentrated marketsack market discipline, which leads to lower efficiencies (Berger andannan, 1998). However, bank restructuring has changed not only
he market structure but also the governance structure of banks,nd governance changes should improve bank efficiency through
etter governing of banks and the risk taking behaviour of bankwners (Caprio et al., 2007; Laeven and Levine, 2009).
1 Estimates show the social welfare loss of market power was 0.54% of GDP in002 in the EU-15 (Maudos and de Guevara, 2007) and 0.15% of GDP in 2005 inexico (Solís and Maudos, 2008).
Governance changes resulted from the withdrawal of the Staterom the banking sector via privatisation, the consolidation of dis-ressed banks into healthier ones, and the removal of restrictionsn foreign banks that has allowed them to penetrate the markethrough acquisition of, often troubled, local banks. The reform pro-ess implicitly assumed that private ownership of banks yields aore efficient outcome. Public ownership of banks is a feature
f institutional and financial underdevelopment (La Porta et al.,002) and state-owned banks had tended to serve political andocial purposes in Latin America (Carvalho et al., 2009). In thearly 1990s, state-owned banks held 45 and 50% of banking sec-or assets in Argentina and Brazil (Carvalho et al., 2009), and 100%n Mexico following the 1982 bank nationalisation (Haber, 2005).gency problems can explain performance differences betweentate-owned and privately owned banks (Megginson, 2005) and theormer are characterised by low levels of profitability and capital-zation, poor credit control, and inefficient control of costs (Cornettt al., 2010). According to Ness (2000) public ownership createdoral hazards between the government’s economic and political
oals and bank’s business goals, and the relatively large size of pub-ic banks conferred a too-big-to-fail status that required frequentse of public funds to support ailing institutions, which justifiedhe bank privatisation process.
To facilitate competition, and to recapitalise distressed banks,overnments repealed restrictions on foreign bank entry.2 Theale of local banks to foreigners implicitly assumes that foreignanks possess superior management skills and technological capa-ilities due to competitive conditions in the host country thathould allow these firms to export their efficiency advantageso the host country (Berger et al., 2000). Foreign bank entry isxpected to boost banking sector efficiency because incumbentomestic banks must improve efficiency or face losing market shareClaessens et al., 2001). However, operational diseconomies asso-iated with distance from the home headquarters, and culturalifferences between the home and host countries, can raise costsnd lessen the efficiency of foreign banks (Berger et al., 2000; Mian,006). Evidence suggests that foreign bank penetration of Latinmerican banking sectors following restructuring did improveompetition, particularly when more efficient and less risky for-ign banks entered the market (Jeon et al., 2011). Foreign bankenetration can adversely affect the efficiency of incumbents if for-ign banks cherry pick the best customers and force local banks toervice higher risk customers (Dages et al., 2000; Paula and Alves,007; Jeon et al., 2011). Bank credit to the private sector could fallecause of information constraints facing foreign banks that are
ess effective at monitoring soft information (Detragiache et al.,008), and this feature could also result in certain sectors facingnancial exclusion (Mian, 2006).
This study will test the quiet life hypothesis on a sample of 419ommercial banks in the four most economically powerful Latinmerican countries between 1985 and 2010.3 There are grounds
o expect the quiet life hypothesis will hold: the bank restructur-ng and M&A processes have increased concentration and under
less competitive market structure banks might exploit marketower and behave less competitively (Boyd and De Nicolo, 2005).lternatively, the efficient structure hypothesis suggests that more
fficient banks are better equipped to survive competitive pres-ures and these banks gain market share at the expense of lessfficient firms (Demsetz, 1973). The quiet life hypothesis posits a
2 Between 1990 and 2004, foreign bank penetration of Argentina and Brazilncreased by around four-fold, by two-fold in Chile, and from 2 to over 80 per centn Mexico (Domanski, 2005).
3 The countries are Argentina, Brazil, Chile and Mexico.
iamsdctte
tability 8 (2012) 263– 276
egative relationship between bank efficiency and market powerhereas the expected relationship is positive should the efficient
tructure hypothesis hold. To date, the empirical evidence is incon-istent on the quiet life hypothesis. Rhoades and Rutz (1982) in atudy of US unit banks find market power is correlated with riskversion. Berger and Hannan (1998) report evidence of the quiet lifen concentrated regional banking markets in the US, and Delis andsionas (2009) accept the hypothesis for samples of US and Euro-ean banks. It is rejected for commercial banks in the US (Koettert al., 2012) and Europe (Maudos and de Guevara, 2007).
The dichotomy in the empirical evidence is attributed to theimultaneous relationship between efficiency and market power.n econometric models, simultaneity creates a contemporaneousorrelation between the regressors and the error term, which ifot controlled for leads to inconsistent estimates. The interdepen-ence between bank efficiency and market power implies thateverse causality is a possibility. The efficient structure hypoth-sis, for instance, posits that efficiency may be driving markettructure (Turk Ariss, 2010). The problem is illustrated further byhe empirical record that shows that the quiet life hypothesis isccepted in studies that do not control for simultaneity (Berger andannan, 1998) whilst it is rejected in studies that account for theroblem (Maudos and de Guevara, 2007; Koetter et al., 2012). Inhat follows I use two-stage least squares and instrumental vari-
bles (2SLSIV) methods to obtain consistent parameter estimatesf the relationship between bank efficiency and market power tonsure that the direction of causality runs from the latter to theormer.
This study draws on developments in the efficiency litera-ure that suggest ways to treat unobserved heterogeneity acrossanks. Standard panel data approaches confound any time invari-nt cross-firm heterogeneity with the inefficiency term, which canseriously distort” estimated efficiency (Greene, 2005a, p. 270;ester, 1997; Orea and Kumbhakar, 2004; Bos et al., 2009). In a
tudy of alternative frontier methods using US bank data, Greene2005a) finds estimated inefficiency and the standard deviationf inefficiency to be lower when drawn from random parame-ers stochastic cost function models compared to standard models.he findings emphasise the effectiveness of the former class ofodel in separating out heterogeneity from inefficiency, which
esults in more precise estimates of efficiency. Other studies reporthat the efficiency ranks of banks are sensitive to the treatmentf heterogeneity (Bos et al., 2009). Drawing on these findings,
use random parameters stochastic frontier models to estimateank efficiency. I use conventional Lerner indices to measure mar-et power under the assumption that banks are fully efficient. Toircumvent the limiting assumption, Koetter et al. (2012) haveeveloped an efficiency-adjusted Lerner index to allow for the pos-ibility that banks may forgo profits in return for a quiet life. Imploy both types of index for robustness purposes.
In what follows, I use two-stage least squares to estimate theelationship between bank efficiency and market power in a testf the quiet life hypothesis. The bulk of previous studies focusttention on the relationship between cost efficiency and marketower. In line with Koetter et al. (2012) I also employ the profit effi-iency concept to control for the fact that banks may forgo profitsn return for a quiet life. Following Maudos and de Guevara (2007) Issume that banks can exercise market power in loans and depositarkets and test the quiet life hypothesis in each market. I use
tochastic frontier methods and the translog functional form toerive marginal costs for loans, deposits, and assets to construct
onventional Lerner indices for loans and deposits markets, andhe efficiency adjusted Lerner index for assets markets, followed byests for the quiet life hypothesis in each market. As a test of param-ter stability and because of changes in accounting standards that
ncial S
ma
(aaiorvmaiiDiisr
dwahlteeftt
dlr
2
topcsarbbtebtihfafae
t
e
bIorpaeCiotft
aiCelcPsr(2la2stTnb2
mgaBoe2bfviWvAmSpdaec
p
J. Williams / Journal of Fina
ay affect the precision of reported data, I split the dataset into prend post-restructuring periods and re-estimate the relationships.
In a second set of regressions, I use a difference-in-differencesDiD) approach to tackle the question of how bank restructuringnd the resulting governance changes impact on bank efficiencynd market power. Stiroh and Strahan (2003) use this approach tonvestigate how deregulation affects bank performance in a studyf US banks. I select two treatments, privatisation and the repeal ofestrictions on foreign bank entry, and limit the sample to an inter-al of six years either side of the start of the treatments to gauge aedium term perspective. The DiD models show if privatised banks
nd foreign bank acquisitions significantly outperform the trendsn cost and profit efficiencies, and in the efficiency adjusted Lernerndex and marginal cost of assets once treatments are enacted.iD models offer an accurate test of public policy choices and it
s an empirical matter to resolve if there is a preferred restructur-ng strategy. The present study aims to fill a gap in the literatureince few studies investigate the effect on bank efficiency of bankestructuring.4
By way of preview, I estimate cost and profit efficiencies anderive market power indices for 419 banks across 1985–2010,hich yields a total of 4571 observations. Market power in loan
nd deposit markets has fallen over time. Cost efficiencies areigher and less volatile than profit efficiencies. Tests of the quiet
ife hypothesis firmly reject and the evidence offers support forhe efficient structure hypothesis. Privatisation has improved profitfficiencies at banks that were less profit efficient prior to the gov-rnance change. Whilst the efficiency adjusted Lerner index risesor the banking sector following privatisation it is driven by reduc-ions in marginal cost. The impact of foreign bank entry is mute inerms of efficiency and market power.
The rest of the paper is organised as follows. Section 2 reviewsevelopments in banking in Latin America and surveys relevant
iterature. Sections 3 and 4 present the methodologies and discussesults, respectively. Finally, Section 5 offers some conclusions.
. Banking sector developments in Latin America
Fundamental shifts in public policy have led to a reconfigura-ion of the industrial structure of Latin American banking sectorsver the past quarter of a century. In the late 1980s and early 1990s,olicies associated with financial repression, such as interest rateontrols and directed lending, gave way to liberal policies thatought to increase competition and efficiency. Changes in entrynd exit conditions, the privatisation of state-owned banks, andepeal of restrictions on foreign bank entry facilitated changes inank governance. The roots of the policy changes lay in nationalanking sector crises, which required government participationo reshape banking structures by reducing state ownership andncouraging greater private sector involvement. In the mid-1990s,ank resolution programmes revived troubled banks and preparedhe ground for a consolidation process. The restructuring processnvolved nationalisation of banks; the transfer of ownership toealthy institutions; liquidation of bankrupts; and use of public
unds to recapitalise and give liquidity to distressed banks (Gelosnd Roldós, 2004). Privatisation and repealing restrictive laws onoreign bank entry acted as catalysts for changes in bank ownershipnd lead to the reform of corporate governance at banks (Carvalho
t al., 2009).
Across Latin America, consolidation produced a sharp decline inhe number of banks. Between 1994 and 2000, bank numbers fell
4 One exception is a study of the impact of the IMF bank restructurings on thefficiency of banks following the Asian crisis in 1997–1998 (Ariff and Can, 2009).
hgfsofb
tability 8 (2012) 263– 276 265
y 45% in Argentina; 21% in Brazil; 22% in Chile; and 36% in Mexico.n general, the rate of decrease in bank numbers exceeded the ratef increase in market concentration. The three-firm concentrationatio increased for Brazil and Mexico by around 5 and 8 percentageoints to 55.2% and 56.3% between 1994 and 2000. For Argentinand Chile, concentration remained constant at around 40%. Thextent of foreign bank penetration varies across countries. In 1990,hile had the highest level (19% of banking sector assets) and it
ncreased to 42% in 2004. By comparison, foreign banks in Mexicowned 2% of assets in 1990 before growing to 82% in 2004. Overhe same period, foreign bank market share in Argentina increasedrom 10% to 48% whilst it grew from 6% to 27% in Brazil, which hashe lowest level of foreign bank penetration.
Conventional indicators of concentration – concentration ratiosnd Herfindahl indices – show a general increase in concentrationn the latter half of the 1990s (IMF, 2001; Gelos and Roldós, 2004).onsistent with evidence from the industrial countries and othermerging markets, Latin American banks operate under monopo-istic competition; in general, greater concentration did not weakenompetitive conditions (Yeyati and Micco, 2007; Yildirim andhilippatos, 2007; Gelos and Roldós, 2004). Yet there are incon-istencies: studies confirm competition increased in Argentina,emained constant in Mexico, and changed little in Brazil and ChileGelos and Roldós, 2004) or weakened (Yildirim and Philippatos,007). Other evidence suggests national level findings of monopo-
istic competition do not generalise across bank size in Argentinand Chile (Yildirim and Philippatos, 2007) and Brazil (Belaisch,003). Competitiveness differs across bank ownership: in Brazil,mall banks and public banks operate under monopolistic compe-ition whilst large banks and foreign banks behave competitively.he evidence implies competitive conditions differ across local andational markets (Belaisch, 2003): in local markets, private-ownedanks are more pro-competitive than public banks (Coelho et al.,007).
Bank privatisation and foreign bank penetration altered thearket structure of national banking sectors and transformed the
overnance structure of banks as new, private owners (domesticnd foreign) assumed control of banks. Formerly, Argentina andrazil had extensive state-owned banking sectors but privatisationffloaded banking sector assets onto the private sector that wasxpected to manage the assets more efficiently (Carvalho et al.,009). State-owned banks had served political and social purposesut their characteristics included weak loan quality, underper-ormance, and poor cost control. Yet, privatisation outcomes areariable. In Argentina and Brazil, privatised bank performancemproved post-privatisation (Berger et al., 2005; Nakane and
eintraub, 2005). In contrast, the failed 1991 Mexican bank pri-atisation programme cost an estimated $65 billion (Haber, 2005).cross the region, foreign banks have acquired large, local banks,any under temporary government control for restructuring.
ome evidence finds a positive association between foreign bankenetration and bank efficiency. There are caveats: the need toistinguish between the performance of existing foreign banksnd local banks acquired by foreign banks; and to disentangle theffects of foreign bank entry from other liberalisation effects thatould impact bank efficiency.
Studies report differences in performance between local,rivate-owned and foreign-owned banks. Foreign banks achievedigher average loan growth (in Argentina and Chile) with loanrowth stronger at existing foreign banks compared to acquiredoreign banks. This suggests management at foreign bank acqui-
itions focused on restructuring their acquisitions and integratingperations with the parent (foreign) bank. The cautious nature oforeign bank strategies explains why foreign banks, and foreignank acquisitions in particular, achieved better loan quality than
2 ncial S
lailetbr2
ftcfeogidilVedtts2
oeMlgbaerclatb–sl–2
ctmoobmtteboitoa
3
3
ogfacseamcbsr
f(fitmafdiaw
wyifa(v�
r
icj�iarac
66 J. Williams / Journal of Fina
ocal banks (Clarke et al., 2005), although stronger provisioningnd higher loan recovery rates translated into weaker profitabil-ty at foreign banks. Foreign banks are relatively more liquid, relyess on deposit financing, and produce stronger loan growth duringpisodes of financial distress than domestic banks. It is suggestedhat the greater intermediation efficiency of foreign banks aroseecause they were more able to evaluate credit risks and allocatedesources at a faster pace than their local competitors (Crystal et al.,002).
Evidence from Argentina shows state-owned banks underper-ormed against private-owned and foreign-owned banks due partlyo poor loan quality associated with direct lending and subsidisedredit. Bank privatisation produced efficiency gains because ofalling non-performing loans and higher profit efficiencies. How-ver, local M&A activity and foreign bank entry exerted little effectn bank performance (Berger et al., 2005). These findings do noteneralise to Brazil where foreign banks faced difficulties in adapt-ng to the peculiarities of the Brazilian banking sector, which isominated by local, private-owned banks (Paula, 2002). The empir-
cal record offers no support to suggest foreign banks are more oress efficient than domestic banks (Guimarães, 2002; Paula, 2002;asconcelos and Fucidji, 2002). This is unsurprising in the light ofvidence that the operational characteristics and balance sheets ofomestic and foreign banks are similar (Carvalho, 2002). Hence,he expected benefits of foreign bank penetration have been slowo emerge because foreign banks follow operational characteristicsimilar to large domestic, private-owned banks (Paula and Alves,007).
Although foreign bank penetration and foreign banks’ sharef bank lending are positively related, evidence suggests for-ign banks engage in cherry-picking behaviour. In Argentina andexico, foreign banks concentrated lending in the commercial
oans market and limited exposure to the household and mort-age sectors (Dages et al., 2000; Paula and Alves, 2007). Foreignank acquisitions in Argentina used growth in lending to diversifyway from manufacturing and target consumer markets (Bergert al., 2005). In addition, foreign banks aggressively penetratedegional markets that eliminated concerns over geographic con-entration, and increased regional lending to offset changes inocal banks’ lending (Clarke et al., 2005). Lastly, foreign banks aren important source of finance. Their loan growth is higher (bet-er quality and less volatile) than local (especially state-owned)anks (Dages et al., 2000). Foreign banks – and private local banks
responded to market signals with pro-cyclical lending that isensitive to movements in GDP and interest rates. Foreign banks’oan growth and lower volatility – even during crisis periods
suggests they can help to stabilise bank credit (Dages et al.,000).
Whereas policymakers expect consolidation to lead to greaterompetition and efficiency improvements, there is the possibilityhat competitive gains would not materialize, and instead bank
arket power would increase. The latter implies that the evolutionf highly concentrated market structures could limit the deepeningf financial intermediation and the development of more efficientanking sectors (Rojas-Suarez, 2007). Since a non-competitivearket structure often produces oligopolistic behaviour by banks,
he suggestion is that more consolidation could incentivise bankso exploit market power rather than become more efficient. In gen-ral, the literature rejects the notion of collusion between banks,ut evidence from Brazil suggests that banks possess some degreef market power (Nakane et al., 2006). Other Brazilian evidence
llustrates the complexities associated with identifying competi-ion effects. Whereas the banking sector operates under conditionsf monopolistic competition, this finding cannot be generalisedcross bank ownership and size.
dwTi
tability 8 (2012) 263– 276
. Preferred econometric framework
This section introduces the methods we shall use in the analysis.
.1. Stochastic frontier model
Despite methodological advances in efficiency modelling, anutstanding anomaly is how to account for unobserved hetero-eneity. Unobserved heterogeneity generally enters the stochasticrontier through the form of either fixed or random effects. Thispproach can confound cross firm heterogeneity with the ineffi-iency term that will bias estimated inefficiency. Greene (2005a,b)olves this problem by extending both fixed effects and randomffects models to account for unobserved heterogeneity. The liter-ture refers to them as “true” effects models. Applications of theseodels in banking are rare. Greene (2005a) applies several specifi-
ations of a stochastic frontier cost function to a sample of 500 USanks in a seminal paper demonstrating statistical advances. In atudy of German savings banks, Bos et al. (2009) find the efficiencyanks of banks are sensitive to the treatment of heterogeneity.
The estimations reported in this study are based on the generalramework developed by Greene (2005a). In the most general costand profit) function specification that is reported below, the coef-cients on the linear terms in the output variables, input prices, andhe constant term, are assumed to be random with heterogeneous
eans. The heterogeneous means of these random coefficientsre linear in average asset size. The coefficients on all other costunction covariates are assumed to be constant. The log standardeviation of the half-normal distribution that is used to define the
nefficiency term in the cost function is assumed to be linear insset size. The coefficient on asset size is assumed to be randomith a constant mean. The most general specification is as follows:
cit = ˛i + �′ixit + �′yit + vit + uit
vit∼N(0, �2v ), where �2
v is constant
uit = |Uit |, where Uit∼N(0, �2ui
) and �ui = �u exp(�i)
(˛i �′i)
′ = ( ¯ ¯ ′)′ + �˛,� si + �˛,� (w˛iw�′
i)′
�i = � + ı�si + ��w�i
(1)
here xit is a (3 × 1) vector of output variables and input prices;it is a (11 × 1) vector of other cost function covariates; and sis the average asset size of bank i. The coefficient vectors are asollows: (�i ˇ′
i)′ is a (4 × 1) vector of random coefficients; ( ¯ ¯ ′)′
nd ��,� are (4 × 1) vectors of (fixed) coefficients; ��,� is a free4 × 4) lower-triangular matrix of (fixed) coefficients; � is a (17 × 1)ector of (fixed) coefficients; �i is a random coefficient; � and
� are (fixed) coefficients. (w˛iwˇ′
i)′ is a (4 × 1) vector of NIID
andom disturbances, where wˇ′i= {wˇji
} for j = 1, . . ., 3; and w�i
s a NIID random disturbance. The individual elements of theoefficient vectors are denoted as follows: ˇ′
i = {ˇji}, ¯ ′ = { ¯j} for
= 1, . . ., 3; �˛,�′ = {ıj} for j = 0, . . ., 3; �′ = {�j} for j = 1, . . ., 11; and
�,� = {� jk} for j = 0, . . ., 3 and k = 0, . . ., j. The specification of ��,�mplies the variances of the random coefficients conditional on si
re var(˛i|si) = �200, var(ˇji|si) = ∑j
k=0�2jk
, var(�i|si) = �2�
. The cor-esponding conditional standard deviations are denoted �(˛i|si),nd so on. The specification of ��,� allows for non-zero conditionalovariances between the elements of (˛i ˇ′
i)′.
I report estimations of a restricted version of Eq. (1). For ran-
om effects and random coefficients on the four output variablesith heterogeneous means, the restrictions are � = ı� = �� = 0.
he heterogeneous means of the random coefficients are linearn average asset size. An advantage of the random parameters
ncial S
sfd
ipatt(dfe
dt1tfesaoctterohps
wedrv((ioceptt
sf
3
m
iefb
L
ti
rlImapapt
ittb(dpoe
L
TfpmIpataefc
3
mf
J. Williams / Journal of Fina
tochastic frontier over standard panel data approaches is that theormer model relaxes the restrictive assumption of a common pro-uction technology across firms (Tsionas, 2002).
The random parameters stochastic frontier cost (profit) functions estimated by maximum simulated likelihood. In the estimationrocedure I use 500 Halton draws to speed up estimation andchieve a satisfactory approximation to the true likelihood func-ion. uit has a half-normal distribution truncated at zero to signifyhat each bank’s cost lies either on or above (on or below) the costprofit) frontier. Deviations from the frontier are interpreted as evi-ence of the quality of bank management. The choice of distributionor the inefficiency term is arbitrary and other distributions aremployed elsewhere in the literature (see Greene, 2008).
To estimate Eq. (1) I model bank production using the interme-iation approach that assumes banks are financial intermediarieshat purchase input to generate earning assets (Sealey and Lindley,977). The quiet life hypothesis refers to bank management’s con-rol of operating cost. In this study I estimate an operating costunction that excludes the price of financial capital and interestxpense of banks. Since the study aims to estimate the relation-hip between market power (competition) and efficiency in loannd deposit markets, the cost function specifies both variables asutput. The cost efficiency concept is limited because it does notonsider the fact that banks may assume additional cost in ordero raise profit (Berger and Mester, 1997). I estimate the alterna-ive profit function (Humphrey and Pulley, 1997) to derive profitfficiencies under the assumption that increases in bank cost canealise improvements in profit. The frontier model contains twoutputs and two inputs; I apply standard restrictions of linearomogeneity in input prices and symmetry of the second orderarameters. To model the underlying cost structure of the bankingector I use the translog functional form – see Eq. (2):
here ln OC is the log of operating cost (the sum of personnelxpense and non-interest expense); in the profit function, theependent variable (ln PBT) is the log of profit before tax. For bankseporting a loss, the dependent variable is adjusted to take thealue of 1 and the absolute value of loss is specified as a covariateunity for banks achieving profit) termed the net profit indicatorBos and Koetter, 2011). In the profit function the sign before �is negative; ln Loansi is the log of gross loans; ln Depi is the logf customer deposits; ln wh is the log of input prices (physicalapital (non-interest expense/fixed assets) and labour (personnelxpenses/total assets)); T is a time trend; �i are identical and inde-endently distributed random variables, which are independent ofhe �i, which are non-negative random variables that are assumedo account for inefficiency.
Table 1 shows descriptive statistics of the variables used in thetochastic frontier models by country. I estimate Eq. (2) separatelyor each country.
.2. Calculation of the Lerner index
Market power reflects a bank’s ability to set price (p) abovearginal cost (mc) (Lerner, 1934). It is measured by the Lerner
odTb
tability 8 (2012) 263– 276 267
ndex to proxy the competitive behaviour of banks. Eq. (3)xpresses the Lerner index (Lerner) of market power as the dif-erence between price and marginal cost scaled by the price forank i at time t:
ernerit = pit − mcit
pit(3)
Following Solís and Maudos (2008) I estimate marginal costs byaking the derivatives of loans and deposits from Eq. (2) as shownn Eq. (4):
mcLit = ∂cit
∂Loansit=
[�L + �LL ln Loansit +
∑�hL ln whit
+ �LD ln Depit + �LTrendt
]cit
Loansit
mcDit = ∂cit
∂Depit=
[�D + �DD ln Depit +
∑�hD ln whit
+ �LD ln Loansit + �DTrendt
]cit
Depit
(4)
In the conventional Lerner index for loans, pit is expressed as theatio of interest received-to-earning assets and the marginal cost ofoans (mcLit) is derived from Eq. (4). In the Lerner index for deposits,
define pit as the ratio of interest paid-to-purchased funds with thearginal cost of deposits (mcDit) derived from Eq. (4). I use earning
ssets and purchased funds as denominators in the expressions forrice because of difficulties in separating interest flows for loansnd deposits from interest received and paid on other items. Thisroblem is particular not only to this study and other studies usehe aforementioned solution for price (Turk Ariss, 2010).
Standard Lerner indices are biased if price or marginal cost isncorrectly estimated. Koetter et al. (2012) note that the conven-ional approach assumes banks are fully efficient. Unless it holds,his assumption biases the conventional Lerner index becauseanks may exploit pricing opportunities arising from market powerTurk Ariss, 2010). To circumvent this anomaly, Koetter et al. (2012)evelop an efficiency-adjusted Lerner index that incorporates theossibility that banks may forgo profits – in terms of inefficientutput pricing – in exchange for a quiet life. Eq. (5) shows thefficiency-adjusted Lerner index:
erner =
(PBT/TA + TVC/TA
)− MC
PBT/TA + TVC/TA(5)
o obtain Eq. (5) I estimate variable cost and alternative profitunctions. The predicted values of total variable cost (TVC) androfit before tax (PBT) are scaled by total assets (TA) and estimatedarginal cost is subtracted from the numerator. For this purpose,
re-specify Eq. (4) on the right-hand side to include a single out-ut (assets) and include the price of financial capital (measureds the ratio of interest expense-to-customer deposits) in additiono the two input prices. In the cost function the dependent vari-ble, variable cost, includes interest expense. In what follows, Imploy efficiency-adjusted indicators of market power to controlor the assumption that banks are fully efficient in addition to theonventional measures.
.3. The relationship between market power and bank efficiency
The relationship between bank efficiency and market poweray be subject to reverse causality. Indeed, the empirical support
or the quiet life hypothesis reported in early US studies is contested
n the grounds that the results may be driven by the interdepen-ence between efficiency and market power (Koetter et al., 2012).o ensure that the direction of causality runs from market power toank efficiency I use an instrumental variables approach to control
268 J. Williams / Journal of Financial Stability 8 (2012) 263– 276
Table 1Descriptive statistics for stochastic frontier models.
Argentina, N = 117; Obs = 1288 Brazil, N = 193; Obs = 2083 Chile, N = 40; Obs = 570 Mexico, N = 61; Obs = 631
Mean Std Dev Mean Std Dev Mean Std Dev Mean Std Dev
Notes: The data are denominated in millions of US dollars at 2000 prices, except for the price data that are expressed as ratios.O s loan( expep
fo
mtwtAtpm2rripcrwnTodEthrHtiXrh
lfiicboLc2olbT
eciorowto(gsb
3
caoemda
y
wcgtobattptMrao
t
COST, operating cost; VCOST, variable cost; PBT, profit before tax; LOANS, grosinterest expense/purchased funds); PPCAP, price of physical capital (non-interestrofit indicator.
or the endogenous relationship between the two main variablesf interest.
In what follows, the dependent variable is bank efficiency esti-ated from Eq. (2). The endogenous covariate is the Lerner index
hat I instrument using its one period lag. A negative coefficientill indicate support for the quiet life hypothesis whilst a posi-
ive coefficient is suggestive of the efficient structure hypothesis. vector of exogenous covariates controls for bank specific charac-
eristics. The logarithm of total assets is proxy for bank size. Oneotential outcome of the consolidation process is that larger banksight eventually behave less competitively (Boyd and De Nicolo,
005). Indeed, the empirical evidence is mixed on the sign of theelationship between size and efficiency. Stiroh and Strahan (2003)eport that successful banks survive and increase market share. Include a market share variable (for loans, deposits, assets) andosit that a positive relationship between market share and effi-iency adds support for the efficient structure hypothesis. If growthesults from mergers and acquisitions, the expected relationshipith efficiency is less clear cut because of lags in implementingew operating structures and systems etcetera (Jeon et al., 2011).he ratio of loans-to-total assets measures the loan intensivenessf bank balance sheets. It is reasonable to assume that loan pro-uction is relatively more costly than say purchasing securities.ffective screening and monitoring is required to reduce informa-ion asymmetries but it raises cost. Banks in emerging marketsave become more diversified in terms of revenue and this shouldeduce risk and enhance profit (Sanya and Wolfe, 2011). I use aerfindahl–Hirschman index measure of revenue diversification
hat is calculated as∑n
i=1(Xi/Q )2 where the X variables are netnterest revenue and net non-interest income and Q is the sum of
(Acharya et al., 2006). The ratio of capital-to-assets is proxy forisk with higher levels of prudence expected to be correlated withigher efficiencies (Berger and Mester, 1997).
Four other exogenous covariates control for time and countryevel effects. The ratio of banking sector credit-to-GDP indicatesnancial deepening, which Levine (2005) suggests is important
n exerting corporate governance on bank borrowers. Incrementalredit provision requires further screening and monitoring costs foranks that could reduce cost efficiencies. An unintended outcomef the former specialised financial sector models that characterisedatin America until recently is the relatively low levels of finan-ial depth and private sector supply of bank credit (Rojas-Suarez,007). I use an alternative measure of financial depth, the ratio
f M2-to-GDP to capture depth in deposit markets. The typicallyow depth ratios reported in Latin America reveal the potential foranks to expand credit and take deposits (Carvalho et al., 2009).he growth rates of M2 and real GDP are proxy for financial sector
eofa
s; DEP, customer deposits; ASSETS, total assets; PFCAP, price of financial capitalnse/fixed assets); PLAB, price of labour (personnel expense/total assets); NPI, net
xpansion over the real economy (potential bubbles) and businessycle effects. Lastly, two dummy variables account for differencesn bank ownership. The dummies identify periods (years) of state-wnership and foreign-ownership of banks. State-ownership hasesulted in less efficient banks (Megginson, 2005) and poorly devel-ped banks (Barth et al., 2001). This underperformance is correlatedith the level of government involvement and the perverse incen-
ives of political bureaucrats (Cornett et al., 2010). The evidencen the relative efficiency of foreign and local banks is subjectiveBerger et al., 2000) with no clear conclusion. Recent evidence sug-ests that the mode of foreign bank entry is important in generatingpill over effects for incumbent banks with de novo entry s moreeneficial than entry by M&A (Jeon et al., 2011).
.4. Difference-in-differences analysis of governance change
To explicitly consider the impact of governance changes asso-iated with the bank restructuring process in Latin America, I use
difference-in-differences (DiD) approach to identify the effectsf two treatments, privatisation and foreign acquisition, on bankfficiency and market power. The DiD approach requires a treat-ent group of banks that undergo a change that does not impact
irectly upon a control group. The analysis considers before andfter treatment periods. The DiD model is presented in Eq. (6):
here yit is measured efficiency; dTi is the treatment dummy thatontrols for differences in outcomes of the treatment and controlroups prior to the treatment; d2t is a time dummy that controls forhe aggregate factor, which causes efficiency to change even with-ut the treatment; ı0 identifies how the two groups were impactedy time or aggregate factors that are affected only with time; ˇ1ccounts for differences in the means of the treatment and con-rol groups; and ı1 is the estimate of the effect of the treatment onhe treated banks. The vector Xit contains covariates lagged by oneeriod to control for endogeneity and � is the vector of coefficientso be estimated. The covariates are the ratios of credit-to-GDP and
2-to-GDP, the rates of growth of real GDP and M2; the loga-ithm of the Herfindahl–Hirschman index of banking sector assetss proxy for market concentration; and country controls in the formf dummy variables.
I estimate Eq. (6) twice. The first treatment is privatisation. Thereatment group contains privatised banks only. To obtain clean
stimates of the impact of privatisation I identify the start datef the privatisation processes by country. The start date is 1991or Argentina and Mexico and it is 1997 for Brazil. I construct
period of six years either side of the start date to identify a
ncial S
mppss1b
3
dfmDUbcrTc
4
4v
aapcopa
LcitpptWfiLibtdtmiwcpcyw
4
aas
TW
Nf
J. Williams / Journal of Fina
edium-term impact. Chile is omitted. The censoring of the sam-les reduces the number of observations to 2001. I follow the samerocedure for foreign bank acquisitions. 1995 is selected as thetart date for Argentina, Brazil and Chile since this year saw thetart of a wave of foreign bank entry. The start year for Mexico is998 when the (1995) repeal of restrictions on foreign bank entryecame effective. There are 2604 observations in this sub-sample.
.5. Data
A unique feature of this study is the construction of a panelataset covering over a quarter of a century from 1985 to 2010or banks from Argentina, Brazil, Chile and Mexico. Financial state-
ent data are sourced from the IBCA and BankScope databases.ata are deflated by national GDP deflators and converted intoS$ millions at 2000 prices. The dimension of the dataset is 419anks and 4571 observations over 26 years. Bank governance andhanges therein have been identified using BankScope, central bankeports, academic papers, newswire services, and bank web sites.he macroeconomic data are from the World Bank Financial Indi-ators and World Economic Outlook databases.
. Empirical results
.1. Summary findings for the efficiency and market powerariables
Table 2 shows weighted averages of the Lerner indices for loansnd deposits markets, the efficiency adjusted Lerner index forssets, and estimated operating cost, variable cost, and alternativerofit efficiencies for the 419 banks by year from 1985 to 2010. The
onventional Lerner indices exhibit a mild reduction in magnitudever time suggesting that product markets in the post-resolutioneriod became more competitive. The data show deposit marketsre more competitive than lending markets. A comparison with
tpme
able 2eighted efficiencies and Lerner indices of market power; by year – all banks.
Year Lerner index
Loans Deposits Eff-adj
1985 0.9381 0.9513 0.1440
1986 0.9107 0.9072 0.0969
1987 0.9103 0.9616 0.0668
1988 0.9411 0.9795 0.0102
1989 0.9469 0.9829 −0.0299
1990 0.9000 0.9697 0.0440
1991 0.8723 0.9552 0.0222
1992 0.8018 0.8943 0.0750
1993 0.7772 0.8395 0.1127
1994 0.8356 0.8543 0.1162
1995 0.9050 0.9159 0.0887
1996 0.8881 0.8921 0.1026
1997 0.8839 0.8750 0.1138
1998 0.8901 0.8584 0.1079
1999 0.8839 0.8568 0.1221
2000 0.8803 0.8638 0.1131
2001 0.8920 0.8775 0.1417
2002 0.8999 0.8997 0.1315
2003 0.8575 0.8242 0.1133
2004 0.8578 0.7670 0.2057
2005 0.8764 0.7778 0.2218
2006 0.8669 0.7812 0.2630
2007 0.8755 0.8128 0.2464
2008 0.9068 0.8766 0.1870
2009 0.8781 0.7769 0.2797
2010 0.8846 0.7917 0.2362
otes: The data are weighted by bank asset share by year. The Lerner indices for loans arom Eqs. (2)–(4). The efficiency-adjusted Lerner index, variable cost efficiency and profit
tability 8 (2012) 263– 276 269
erner indices reported elsewhere in the literature is made diffi-ult by the fact that the estimates presented in this study do notnclude the money market rate in the calculation of the numera-or because interest rates were not fully deregulated for the wholeeriod in each country. The efficiency adjusted Lerner index is com-arable. Koetter et al. (2012) report efficiency adjusted indices inhe region of 50% for US commercial banks between 2002 and 2007.
hilst this figure is roughly twice as large as the comparative dataor Latin American banks, over time the trend shows an increasen the market power of US and Latin banks. Closer analysis of theatin American data shows very high levels of mean marginal costn the late 1980s that began to fall in the first half of the 1990sefore plateauing out to the current level in the early 2000s. Whilsthis suggests better cost management at banks one should note theownward trend in interest rates and convergence of rates acrosshe region as a possible conditioning factor. Nevertheless, for the
ost recent period the efficiency adjusted Lerner index reportedn this study is comparable in magnitude to the US results. The
eighted efficiencies exhibit a pattern that is familiar in the effi-iency literature. Cost efficiencies are higher and less volatile thanrofit efficiencies. The data suggest that the average Latin Ameri-an bank loses over 50% of potential profit to inefficiencies in mostears. I observe that cost efficiencies are weaker in the years inhich market power is greatest.
.2. The relationship between efficiency and market power
The estimation strategy is to run OLS regressions that do not takeccount of the endogenous relationship between bank efficiencynd market power before running the 2SLSIV model to control forimultaneity and endogeneity. First, I shall present the results for
he relationship between cost efficiency and market power for theeriod from 1985 to 2010 each for the loans, deposits, and assetsarkets. In the second set of results, profit efficiency replaces cost
fficiency as the dependent variable. A third and fourth set of results
nd deposits, operating cost efficiency, and profit efficiency Model 1 are estimated efficiency Model 2 are estimated from the single output version of Eqs. (2) and (5).
270 J. Williams / Journal of Financial Stability 8 (2012) 263– 276
Table 3Estimation of the relationship between cost efficiency and market power from Eq. (6); full period 1985–2010.
Robust standard errors clustered at the firm level are reported.
sfdc
ctsLentmfieatsishftasSLebb
iT
Aanmitaaebmmrlh
trrattT
* Statistical significance at the 10 percentage level.** Statistical significance at the 5 percentage level.
*** Statistical significance at the 1 percentage level.
how two-stage least squares estimates of the above relationshipsor the sub-periods 1985–1997 and 1998–2010. In each model stan-ard errors are calculated using the robust cluster method withlustering on bank level.
Table 3 shows the coefficients from the regression in whichost efficiency is the dependent variable. The OLS results appearo support the quiet life hypothesis in the deposit markets andtrongly refute it in the assets market when the efficiency adjustederner index is the market power indicator. The next step is tostimate the 2SLSIV model. In the first-stage regressions, all exoge-ous variables including the instrument, the one period lag ofhe Lerner index, are regressed against the Lerner indicators of
arket power and are reported in Appendix A. The R2 of eachrst-stage model exceeds 60%. This figure overstates the truexplanatory power of the instrument because the control vari-bles contribute to the R2. The partial R2 nets out the effect ofhe control variables and offers a test of whether the model isubject to weak instruments. In the loans market, the partial R2
s 0.4894 compared to the overall R2 of 0.6896. The partial Ftatistic of the first-stage model is 370.66, which rejects the nullypothesis that the instrument is weak. Similarly, the F statistic
or joint significance in the first-stage model is 499.83 satisfieshe requirement that the instrument is not weak. The F statisticsnd significant coefficients on the instruments in each regres-ion show that weak instruments are not a problem (Staiger andtock, 1997; Stock et al., 2002) (see Table A1). The Kleibergen–Paap
M and Wald F statistics show the models are not subject toither under or weak identification. Market power is posited toe endogenous but if it is exogenous then the OLS estimates wille more efficient. The endogeneity test shows this not the case as
taLL
t rejects the null hypothesis that market power is exogenous (seeable 3).
The 2SLSIV estimates refute the quiet life hypothesis for Latinmerican banks. Cost efficiency and market power are positivelynd significantly related in the loans market. The significant andegative relationship found by the OLS estimates in the depositarket loses significance. Lastly, when the conventional Lerner
ndices are replaced by the efficiency adjusted index I find a posi-ive relationship at the 1% significance level between market powernd cost efficiency. In lending and deposit markets, bigger banksnd better capitalised banks are significantly less operating costfficient whereas banks with more loan-intensive balance sheets,anks with higher revenue diversification, and banks with higherarket shares are more cost efficient. In the model of the assetsarket, the dependent variable is variable cost efficiency. The
esults show that larger banks, better capitalised banks, and moreoan intensive banks are relatively more cost efficient whilst moreighly diversified banks are less efficient.
The cost efficiency concept is limited because it fails to considerhe fact that banks may absorb higher costs in exchange for highereturns. Using profit efficiencies remedies this shortcoming and Ie-estimate the models with profit efficiency as the dependent vari-ble. Statistical tests confirm the relevance of the instrument andhe endogeneity of market power. From Table 4 it is not possibleo reject the quiet life hypothesis in loans and deposit markets.he findings suggest that banks with larger margins are willing
o lose more potential profit to inefficiencies. However, when thessumption of efficient banks is relaxed and the efficiency adjustederner is employed as proxy for market power, the sign on theerner index turns positive and significant at the 1% level. This
J. Williams / Journal of Financial Stability 8 (2012) 263– 276 271
Table 4Estimation of the relationship between profit efficiency and market power from Eq. (6); full period 1985–2010.
KP Wald statistic 370.66*** 1116.92 535.54Endogeneity test 5.40** 11.49*** 61.05***
Robust standard errors clustered at the firm level are reported.
fiedp
csbsafi(ad
4
opisfiottacwb
elttthAptiLs
dwaceteL
4m
* Statistical significance at the 10 percentage level.** Statistical significance at the 5 percentage level.
*** Statistical significance at the 1 percentage level.
nding rejects the quiet life hypothesis. In the assets model, profitfficiency is inversely correlated with bank size whereas moreiversified banks, and better capitalised banks are relatively morerofit efficient.
The country level controls show that both cost and profit effi-iency improve when banks supply more credit to the privateector; when market deepening occurs (M2/GDP); and when theusiness cycle is on the up (GDP growth). Increases in the moneyupply (M2 growth) are positively correlated with profit efficiencynd inversely with cost efficiency. From Table 4, the estimated coef-cients on state-owned banks (State) and foreign-owned banksForeign) show that these banks underperform in terms of losing
greater share of potential profit to inefficiencies compared withomestic, privately owned banks.
.2.1. Robustness test of parameter stabilityTo resolve the issue of parameter stability and to take account
f changes in accounting standards made during the restructuringrocess to improve accountability and transparency, the sample
s divided into pre and post-restructuring sub-periods. The firstub-period, 1985–1997, contains the main liberalisation events,nancial crises, and bank resolution programmes. State-ownershipf banks was prevalent although the transfer of bank ownership tohe private sector had started. The second sub-period is from 1998o 2010. A priori I expect any perceived benefits of the reforms
nd resolution programmes to be realised in this period. Economi-ally, the region exhibited strong rates of growth and rising incomeshilst the general lowering of interest rates provided stability to
anks.oo
Table 5 shows the estimated coefficients from the 2SLSIV mod-ls for when cost efficiency is the dependent variable. The quietife hypothesis is rejected by the significant, positive coefficient onhe Lerner index in the loans market over 1985–1997. The magni-ude of the coefficient is nearly twice as big as the coefficient forhe full period. Indeed, we cannot reject nor accept the quiet lifeypothesis in the 1998–2010 period. The results show that Latinmerican banks enjoy a quiet life in deposit markets in both sub-eriods though the significance of the inverse relationship betweenhe Lerner index for deposits and operating cost efficiency weakensn the second sub-period. In contrast, when the efficiency adjustederner index is proxy for market power, the quiet life hypothesis isignificantly rejected but only in the 1998–2010 sub-period.
Table 6 reports the 2SLS estimates using profit efficiency as theependent variable. The estimated coefficients clearly indicate thehilst Latin American banks seems to enjoy a quiet life in loans
nd deposit markets between 1985 and 1997, this is no longer thease in the post bank restructuring period. Furthermore, using thefficiency adjusted Lerner index strongly rejects any support forhe quiet life hypothesis in either sub-period. Taken together, thempirical evidence from the models using the efficiency adjustederner index offer support for the efficient structure hypothesis.
.3. The impact of governance changes on bank efficiency andarket power
I use a difference-in-differences approach to identify the impactf bank privatisation and foreign acquisition following the repealf restrictions on foreign bank entry. For reasons outlined earlier,
272 J. Williams / Journal of Financial Stability 8 (2012) 263– 276
Table 5Estimation of the relationship between cost efficiency and market power from Eq. (6); sub-sample periods 1985–1997 and 1998–2010; 2SLS estimates.
Robust standard errors clustered at the firm level are reported.
petydrarc
odftiitbstaiwmtstptoa
twestspp
bcicetatoaf
sekrkp
* Statistical significance at the 10 percentage level.** Statistical significance at the 5 percentage level.
*** Statistical significance at the 1 percentage level.
olicy makers expected the governance changes to improve bankfficiency. The effects on market power are less certain. Eq. (6) iswice estimated for abridged samples of an interval spanning sixears either side of the beginning of a privatisation process or theecision to repeal restrictions on foreign banks. Table 7 shows theesults for the impact on bank efficiency of privatisation and foreigncquisition, respectively. Consistent with the previous approach,obustness is tested using efficiencies drawn from alternative con-epts.
Columns 1 and 2 show the impact of privatisation whenperating cost efficiency and alternative profit efficiency are theependent variables. Efficiencies are drawn from the two outputrontier model shown in Eq. (2). ı0 quantifies the efficiency trend inhe post-privatisation period for all banks. Although the coefficients positive and implies improvement in both types of efficiencyt lacks statistical significance. The interpretation of ˇ1 revealshe pre privatisation efficiency performance of the pre-privatisedanks when they operated under state control. The coefficient isignificant in both regressions. Under state ownership, the bankshat were later privatised were significantly less profit efficientnd lost a significantly greater proportion of potential profit tonefficiencies. This is not unexpected because state-owned banks
ere beset by asset quality problems that undermined perfor-ance. Furthermore, national governments had to intervene and
ake over some private-owned banks during the crises that weretricken and this type of bank enters the analysis as a privatisa-ion. The coefficient ı1 reveals if the post privatisation efficiency
erformance of the privatised banks is significantly better thanhe trend. The results are interesting and generally accord withur earlier analysis. After privatisation, privatised banks achieven improvement in profit efficiency that is significant greater than
emaw
he trend for the control group. This finding confirms findings else-here in the literature that report significant post-privatisation
fficiency gains. The cost efficiency results are interesting. Undertate ownership, banks are significantly more cost efficient buthis disappears after privatisation. Taken together, the resultsuggest that privatised banks incur additional cost to increaserofit, which arguably is in indication of better managementractices.
Columns 3 and 4 show the impact of foreign bank acquisition onank efficiencies. The results are inconclusive except for a signifi-ant ı0 in the cost efficiency model, which suggests that all banksmprove cost efficiencies following foreign bank penetration. It isonsistent with evidence of spill over efficiency effects due to for-ign bank entry (Jeon et al., 2011). Taken together the results showhat privatisation has produced significant (profit) efficiency gainsnd assume additional cost in the process. The findings confirmhat bank ownership does matter. However, the impact of foreignwnership appears to have stimulated efficiency improvements allround. As a caveat, the interval of six years may not be long enoughor gains to accrue.
Table 8 shows comparative results for the impact of privati-ation and foreign acquisition on market power. In Eq. (6) thefficiency adjusted Lerner index is the preferred indicator of mar-et power and I also specify the marginal cost of assets in separateegressions. Governance changes do not impact as much on mar-et power compared to bank efficiency. In the post-privatisationeriod, the banking sector records a significant increase in the
fficiency adjusted Lerner index due to a significant reduction inarginal costs. Banks that were acquired by foreign banks achieved
significantly lower efficiency adjusted Lerner index before theyere bought.
J. Williams / Journal of Financial Stability 8 (2012) 263– 276 273
Table 6Estimation of the relationship between profit efficiency and market power from Eq. (6); sub-sample period 1985–1997 and 1998–2010; 2SLS estimates.
Notes: (1) After lags the number of banks and observations is 305 and 1630; (2) after lags the number of banks and observations is 351 and 2162.Log HHI is the logarithm of the Herfindahl–Hirschman index on bank assets. Dum signifies a country level dummy variable.
* Statistical significance at the 10 percentage level.** Statistical significance at the 5 percentage level.
*** Statistical significance at the 1 percentage level.
274 J. Williams / Journal of Financial Stability 8 (2012) 263– 276
Table 8Estimation of impact of governance change on market power using Eq. (6).
Notes: (1) After lags the number of banks and observations is 305 and 1630; (2) after lags the number of banks and observations is 351 and 2162; (2) the Lerner index is theefficiency adjusted index shown in Eq. (5); (3) MC equals marginal cost.Log HHI is the logarithm of the Herfindahl–Hirschman index on bank assets. Dum signifies a country level dummy variable.
* Statistical significance at the 10 percentage level.** Statistical significance at the 5 percentage level.
*** Statistical significance at the 1 percentage level.
5
msppaptstitaicvbs
aecisegul
cTtttsepsoertgf
iaiaiAthm
. Conclusions
This paper considers the relationship between efficiency andarket power for Latin American banks using a panel dataset that
pans a quarter of a century. Extensive and sweeping changes tooklace during the period that should condition bank behaviour anderformance. Given the tradition of state involvement in banking,nd the high interest margins observed across the region in com-arison to other regions of the world, there are grounds to assumehat banks enjoyed a quiet life, especially as hyper-inflation yieldedizeable inflationary revenues that may have acted as a disincentiveo control costs. The competitive landscape changed dramaticallyn the mid-1990s when inflationary revenues tended to zero andhe extent of the lack of cost control became apparent particularlyt state-owned banks. The policy response to banking sector crisesn the mid-1990s was to restructure national banking sectors andonsolidate troubled banks. Privatisation and foreign entry are twoehicles that policymakers expect to raise competitive standardsy reducing the role of the state and transforming the governancetructure of banks.
Although more concentrated markets are posited to be char-cterised by less market discipline that in turns lowers bankfficiency, an alternative view suggests that only the most effi-ient banks survive and gain market share. The empirical evidencen this paper firmly rejects the quiet life hypothesis and offersupport for the efficient structure hypothesis. The impact of gov-
rnance changes on banks should not be understated. After initialovernment-driven bank restructurings, the consolidation processnleashed a wave of market-driven M&A. The main players are
ocal banks that were stimulated into improving efficiency by the
A
ompetitive challenge from new entrants, including foreign banks.he difference-in-differences models show the impact of privatisa-ion and foreign acquisition on efficiency and market power. Of thewo modes of governance change, privatisation is the most effectivereatment at least in the medium term as privatised banks makeignificant improvements in profit efficiency above the trend infficiency. For all banks, market power is significantly larger in theost-privatisation period and appears to be strongly influenced by aignificant reduction in marginal cost. However, the market powerf privatised banks falls relative to the trend though the differ-nce is not statistically significant. In contrast, foreign acquisitionealises increases in market power though it is not significant fromhe trend. The evidence shows that privatisation yields efficiencyains at privatised banks and it implies that these banks did notorgo profit efficiencies for a quiet life.
As a caveat, it should be noted that there is some inconsistencyn the results. Whereas the relationship between bank efficiencynd market power as measured by the efficiency adjusted Lernerndex refutes the quiet life hypothesis, the evidence from loansnd deposits market that makes use of conventional Lerner indicess sensitive to the choice of efficiency concept and time period.rguably, this reflects methodological problems associated with
he assumptions of conventional Lerner indices, which in itselfighlights the need for a more robust treatment of the impact ofarket power on bank efficiency.
ppendix A.
See Table A1.
J. Williams / Journal of Financial S
Table A1First stage regressions; full period, 1985–2010.
* Statistical significance at the 10 percentage level.** Statistical significance at the 5 percentage level.
R
A
A
B
BB
B
B
B
B
B
B
C
C
C
C
C
C
C
C
D
D
D
D
D
G
G
G
G
G
H
HH
H
IJ
K
L
L
L
L
M
M
M
M
N
N
N
O
P
P
Rhoades, S.A., Rutz, R.D., 1982. Market power and firm risk. Journal of MonetaryEconomics 9, 73–85.
*** Statistical significance at the 1 percentage level.
eferences
charya, V.V., Hasan, I., Saunders, A., 2006. Should banks be diversified? Evidencefrom individual loan portfolios. Journal of Business 79 (3), 1355–1412.
riff, M., Can, L., 2009. IMF bank-restructuring efficiency outcomes: evidence fromEast Asia. Journal of Financial Services Research 35, 167–187.
arth, J.R., Caprio Jr., G., Levine, R., 2001. Banking systems around the globe: doderegulation and ownership affect performance and stability? NBER ConferenceReport Series, University of Chicago Press, Chicago, London, 31–88.
elaisch, A., 2003. Do Brazilian banks compete? IMF Working Paper WP/03/113.erger, A.N., Mester, L.J., 1997. Inside the black box: what explains differences in
the efficiencies of financial institutions? Journal of Banking and Finance 21,895–947.
erger, A.N., Hannan, T.H., 1998. The efficiency cost of market power in the bank-ing industry: a test of the “quiet life” and related hypotheses. The Review ofEconomics and Statistics LXXX (3), 454–465.
erger, A.N., DeYoung, R., Genay, H., Udell, G.F., 2000. Globalisation of financial insti-tutions: evidence from cross-border banking performance. Brookings–WhartonPapers on Financial Services 3, 23–158.
erger, A.N., Clarke, G.R.G., Cull, R., Klapper, L., Udell, G.F., 2005. Corporate gov-ernance and bank performance: a joint analysis of the static, selection, anddynamic effects of domestic, foreign, and state ownership. Journal of Bankingand Finance 29, 2179–2221.
os, J.W.B., Koetter, M., 2011. Handling losses in translog profit models. AppliedEconomics 43, 307–312.
os, J.W.B., Koetter, M., Kolari, J.W., Kool, C.J.M., 2009. Effects of heterogeneity onbank efficiency scores. European Journal of Operational Research 195, 251–261.
oyd, J.H., De Nicolo, G., 2005. The theory of bank risk taking and competitionrevisited. Journal of Finance 60 (3), 1329–1343.
aprio, G., Laeven, L., Levine, R., 2007. Governance and bank valuation. Journal ofFinancial Intermediation 16, 584–617.
arvalho, F.J.C., 2002. The recent expansion of foreign banks in Brazil: first results.Latin American Business Review 3 (4), 93–119.
arvalho, F.J.C., Paula, L.F., Williams, J., 2009. Banking in Latin America. In: Berger,A., Molyneux, P., Wilson, J. (Eds.), The Oxford Handbook of Banking. Oxford
University Press, pp. 868–902.
laessens, S., Demirgüc -Kunt, A., Huizinga, H., 2001. How does foreign entry affectdomestic banking markets? Journal of Banking and Finance 25, 891–911.
R
tability 8 (2012) 263– 276 275
larke, G.R.G., Crivelli, J.M., Cull, R., 2005. The direct and indirect impact of bank pri-vatization and foreign entry on access to credit in Argentina’s provinces. Journalof Banking and Finance 29, 5–29.
oelho, C.A., de Mello, J.M.P., Rezende, L., 2007. Are public banks pro-competitive?Evidence from concentrated local markets in Brazil. Texto para Discssão. No.551. Pontifícia Universidade Católica do Rio de Janeiro.
ornett, M.M., Guo, L., Khaksari, S., Tehranian, H., 2010. The impact of state owner-ship on performance differences in privately-owned versus state-owned banks:an international comparison. Journal of Financial Intermediation 19, 74–94.
rystal, J.S., Dages, B.G., Goldberg, L., 2002. Has foreign bank entry led to sounderbanks in Latin America? Current Issues in Economics and Finance 8 (1), 1–6.
ages, B.G., Goldberg, L., Kinney, D., 2000. Foreign and domestic bank participation inemerging markets: lessons from Mexico and Argentina. FRBNY Economic PolicyReview, 17–36.
elis, M., Tsionas, E., 2009. The joint estimation of bank-level market power andefficiency. Journal of Banking and Finance 33, 1842–1850.
emsetz, H., 1973. Industry structure, market rivalry, and public policy. Journal ofLaw and Economics 16, 1–9.
etragiache, E., Tressel, T., Gupta, P., 2008. Foreign banks in poor countries: theoryand evidence. Journal of Finance LXII (5), 2123–2160.
omanski, D., 2005. Foreign banks in emerging market economies: changing play-ers, changing issues. BIS Quarterly Review, 69–81.
elos, R.G., Roldós, J., 2004. Consolidation and market structure in emerging marketbanking systems. Emerging Markets Review 5, 39–59.
reene, W., 2005a. Reconsidering heterogeneity in panel data estimators of thestochastic frontier model. Journal of Econometrics 126, 269–303.
reene, W., 2005b. Fixed and random effects in stochastic frontier models. Journalof Productivity Analysis 23, 7–32.
reene, W.H., 2008. The econometric approach to efficiency analysis. In: Fried, H.O.,Lovell, C.A.K., Schmidt, S.S. (Eds.), The Measurement of Productive Efficiency andProductivity Growth. Oxford University Press, pp. 92–251.
uimarães, P., 2002. How does foreign entry affect domestic banking market? TheBrazilian case. Latin American Business Review 3 (4), 121–140.
aber, S., 2005. Mexico’s experiments with bank privatization and liberalization,1991–2003. Journal of Banking and Finance 29, 2325–2353.
icks, J., 1935. The theory of monopoly. Econometrica 3, 1–20.ughes, J.P., Lang, W.W., Mester, L.J., Moon, C.-G., Pagano, M.S., 2003. Do bankers
sacrifice values to build empires? Managerial incentives, industry consolidation,and financial performance. Journal of Banking and Finance 27, 417–447.
umphrey, D., Pulley, L., 1997. Banks’ response to deregulation: profits, technologyand efficiency. Journal of Money, Credit and Banking 29 (1), 417–447.
dence from emerging Asian and Latin American banking markets. Journal ofBanking and Finance 35, 856–875.
oetter, M., Kolari, J., Spierdijk, L., 2012. Enjoying the quiet life under deregulation?Evidence from adjusted Lerner indices for U.S. banks. Review of Economics andStatistics 94 (2), 462–480.
aeven, L., Levine, R., 2009. Bank governance, regulation and risk taking. Journal ofFinancial Economics 93, 259–275.
a Porta, R., Lopez-de-Silanes, F., Schleifer, A., 2002. Government ownership ofbanks. Journal of Finance LVII (1), 265–301.
erner, A., 1934. The concept of monopoly and the measurement of monopolypower. Review of Economic Studies 1 (3), 157–175.
evine, R., 2005. Finance and growth: theory, evidence, and mechanisms. In: Aghion,P., Durlauf, S. (Eds.), The Handbook of Economic Growth. North Holland, Ams-terdam, pp. 865–934.
audos, J., de Guevara, J.F., 2007. The cost of market power in banking: social welfareloss vs. cost inefficiency. Journal of Banking and Finance 31, 2103–2125.
egginson, W., 2005. The economics of bank privatisation. Journal of Banking andFinance 29 (8–9), 1931–1980.
ester, L.J., 1997. Measuring efficiency at US banks: accounting for heterogeneityis important. European Journal of Operational Research 98, 230–424.
ian, A., 2006. Distance constraints: the limits of foreign lending in poor economies.Journal of Finance LXI (3), 1465–1505.
akane, M.I., Weintraub, D.B., 2005. Bank privatization and productivity: evidencefor Brazil. Journal of Banking and Finance 29, 2259–2289.
akane, M.I., Alencar, L.S., Kanczuk, F., 2006. Demand for bank services and marketpower in Brazilian banking. Working Paper Series 107. Banco Central do Brasil,June.
ess, W.L., 2000. Reducing government bank presence in the Brazilian financialsystem: why and how. The Quarterly Review of Economics and Finance 40,71–84.
rea, L., Kumbhakar, S., 2004. Efficiency measurement using a latent class stochasticfrontier model. Empirical Economics 29, 169–183.
aula, L.F., 2002. Expansion strategies of European banks to Brazil and their impactson the Brazilian banking sector. Latin American Business Review 3 (4), 59–91.
aula, L.F., Alves Jr., A.J., 2007. The determinants and effects of foreign bank entryin Argentina and Brazil: a comparative analysis. Investigación Económica LXVI(259), 63–102.
ojas-Suarez, L., 2007. The provision of banking services in Latin America: obstaclesand recommendations. Working Paper No. 124. Center for Global Development,June.
2 ncial S
S
S
S
S
S
S
T
T
V
Y
76 J. Williams / Journal of Fina
anya, S., Wolfe, S., 2011. Can banks in emerging economies benefit from revenuediversification? Journal of Financial Services Research 40 (1–2), 79–101.
ealey, C., Lindley, J.T., 1977. Inputs, outputs and a theory of production and cost atdepository financial institution. Journal of Finance 32, 1251–1266.
olís, L., Maudos, J., 2008. The social costs of bank market power: evidence fromMexico. Journal of Comparative Economics 36, 467–488.
tiroh, K.J., Strahan, P.E., 2003. Competitive dynamics of deregulation: evidence fromU.S. banking, Journal of Money. Credit and Banking 35 (5), 801–828.
taiger, D., Stock, J.H., 1997. Instrumental variables regression with weak instru-ments. Econometrica 65 (3), 557–586.
tock, J.H., Wright, J.H., Yogo, M., 2002. A survey of weak instruments and weak iden-tification in generalized method of moments. Journal of Business and EconomicStatistics 20 (4), 518–529.
Y
tability 8 (2012) 263– 276
sionas, E., 2002. Stochastic frontier models with random coefficients. Journal ofApplied Econometrics 17, 127–147.
urk Ariss, R., 2010. On the implications of market power in banking: evi-dence from developing countries. Journal of Banking and Finance 34,765–775.
asconcelos, M.R., Fucidji, J.R., 2002. Foreign Entry and Efficiency: Evidence fromthe Brazilian Banking Industry. State University of Maringá, Brazil.
eyati, E.L., Micco, A., 2007. Concentration and foreign penetration in Latin American
banking sectors: impact on competition and risk. Journal of Banking and Finance31, 1633–1647.
ildirim, H.S., Philippatos, G.C., 2007. Restructuring, consolidation and competi-tion in Latin American banking markets. Journal of Banking and Finance 31,629–639.
