Do Markets Care about Central Bank Governor Changes? Evidence from Emerging Markets

CHRISTOPH MOSER

AXEL DREHER

Do Markets Care about Central Bank Governor

Changes? Evidence from Emerging Markets

Based on a new daily data set for 20 emerging markets over the period1992–2006, we examine the reactions of foreign exchange markets, do-mestic stock markets, and sovereign bond spreads to central bank governorchanges. We find that the replacement of a central bank governor negativelyaffects financial markets on the announcement day, which is in line withthe hypothesis that newly appointed central bank governors suffer from asystematic credibility problem at the beginning of their tenure. We also findsome evidence that changes in perceived central bank independence affectmarkets.

JEL codes: F30, F34, G14, H63Keywords: central bank governor turnover, monetary policy, emerging

markets, risk premium.

“After arguing behind the scenes with his central bank governor over the directionof interest rates, Prime Minister Thaksin Shinawatra of Thailand . . . dismissed thebanker . . . brought a sharp reaction of financial markets, where it cast doubts over thepolitical independence of the Thai central bank.” (The New York Times [“Chief of ThaiBank Is Dismissed Over Rates,” May 30, 2001, Source: Proquest])

We thank two anonymous referees and the editors, Robert DeYoung and Deborah Lucas, for helpfulcomments. Furthermore, we are grateful to Christian Bjørnskov, Ricardo Caballero, Christian Conrad,Etienne Farvaque, Martin Gassebner, Ashok Kaul, Silke Rath, Jan-Egbert Sturm, Dieter Urban, participantsat the First BBQ Conference, the Verein fur Socialpolitik, the KOF Research Seminar, the Verein furSocialpolitik: Research Committee Development Economics, the Central Banking Conference “DoesCentral Bank Independence Still Matter?” at the Bocconi University 2007, the Brown Bag Seminarat the University of Mainz, the Annual Meeting of the Austrian Economics Association (NOeG), theMacroeconomics Research Meeting 2007, and in particular Helge Berger for suggestions and discussion.We thank Simon Holzhammer, Stefan Keitel, Michaela Lischer, Elisabeth Munch, Nadja Panzer, andChristoph Woodli for excellent research assistance; Jan Schopen for help in compiling the stock marketand foreign exchange market data; and Hendrik van Broekhuizen for excellent proofreading.

CHRISTOPH MOSER is at the ETH Zurich, Swiss Federal Institute of Technology, KOF SwissEconomic Institute (E-mail: [email protected]). AXEL DREHER is a Faculty of EconomicSciences, Georg-August University Goettingen, KOF Swiss Economic Institute, Switzerland,and IZA and CESifo, Germany (E-mail: [email protected]).

Received October 26, 2007; and accepted in revised form July 12, 2010.

Journal of Money, Credit and Banking, Vol. 42, No. 8 (December 2010)C© 2010 The Ohio State University

1590 : MONEY, CREDIT AND BANKING

THE DYNAMIC INCONSISTENCY of low-inflation monetary policycan be overcome by delegating monetary policy to independent and conservativecentral bankers (Rogoff 1985). While many countries have recently granted theircentral banks legal independence, the experience of some countries suggests limitedactual independence, with the head of the central bank frequently being replaced atshort notice and outside the legal schedule. How do financial markets react to changesof central bank governors?1 We expect them to react to such changes if changes conveynew information about future monetary policy. Economic theory suggests that theinflationary bias is determined by the degree of central bank independence and thedegree of the bank’s conservatism.2 Hence, if market participants’ perceptions changewith respect to one of these two dimensions, asset prices should change to the extentof their sensitivity to inflation. There are two transmission channels. First, when thegovernment interferes in the replacement procedure, irregular turnovers are likelyto affect markets’ perceptions about the central bank’s independence. Second, if theperceived inflation aversion of the new central bank governor differs from that of thepredecessor’s, this will alter expected financial market returns.

The literature has shown that the identity of central bank council members hasimportant bearing on economic outcomes. Drawing on a sample of 15 industrializedcountries, Kuttner and Posen (2010) conclude that markets do care about who chairsthe central bank. Central bank governor changes apparently convey new informationabout the future monetary policy, thereby affecting exchange rates and domesticbond yields. Kuttner and Posen (2010) do not, however, find evidence of a genericcredibility problem, that is, a systematic (at least) transitory increase in inflationexpectations at the beginning of a central bank governor’s tenure. This is hardlysurprising, since central bank governor turnovers are mostly predictable and highlydeveloped institutions are likely to reduce the individual governor’s influence inadvanced countries. By contrast, replacing the central bank governor is among themost sensitive decisions for emerging market governments, as these policymakersplay a crucial role in communicating with international markets (Santiso 2003).Credibility is more likely to be an issue here. Surprisingly, whether and to whatextent central bank governor changes in transition countries affect financial marketshas so far not been investigated.

This paper examines the impact of central bank governor changes on domestic andinternational financial markets in emerging economies. Based on a new daily dataset on 20 emerging markets over the period 1992–2006, we study whether foreignexchange rates, domestic stock market indices, and sovereign bond spreads react to theannouncement of a change at the helm of a central bank and whether new governorshave systematic credibility problems. Moreover, we investigate whether changes in

1. We call the heads of the central bank “governors” independent of whether their actual job title is“governor,” “director,” or “president.”

2. By “conservatism” we mean that the central bank has a stronger bias against inflation than thegovernment.

CHRISTOPH MOSER AND AXEL DREHER : 1591

perceived central bank independence or changes in the perceived conservatism of thecentral banker affect markets.

To foreshadow our main results, we find that the replacement of a central bankgovernor negatively affects financial markets on the announcement day. Overall, ourresults are in line with the hypothesis that incoming governors in emerging marketsface a credibility problem. Furthermore, we find evidence that the negative effecton financial markets is primarily driven by central bank governor changes that occurbefore the officially scheduled end of tenure, indicating reactions to perceived changesin central bank independence.

The remainder of the paper is structured as follows. Section 1 derives our hy-potheses, while Section 2 describes the data set. Section 3 discusses the methodologyused, derives the empirical specifications, and describes the results. The final sectionconcludes.

1. HYPOTHESES

Evidence on the impact of who is in charge of economic policies on economicoutcomes is scarce. It is only very recently that studies have started analyzing thisissue. Among them, Jones and Olken (2005) demonstrate that national leaders matterfor economic growth. Dreher et al. (2008) show that the educational and professionalbackground of a head of government matters for the implementation of reforms. Fora sample of Latin American countries, Moser (2006) finds that changing the financeminister had the effect of increasing sovereign bond spreads due to a perceivedworsening of the fiscal policy stance.

Turning to central banks, Gohlmann and Vaubel (2007) provide recent empiricalevidence that the educational qualifications and professions of the central bank’sgoverning council members matter for its effectiveness when attempting to controlinflation. Kuttner and Posen (2010) find that changing a central bank governor conveyssignals about the future course of monetary policy, thereby affecting exchange ratesand financial market returns. However, the direction of these effects subsequent tosuch announcement is not obvious a priori. What are the channels by which turnoversaffect markets and are market reactions positive or, instead, negative? Policymakerscannot credibly commit themselves to low-inflation policy (Kydland and Prescott1977). One approach to overcome this time-inconsistency problem is to establishreputation (Backus and Driffill 1985). While the public may not know central bankers’preferences, policymakers’ behavior conveys information about their characteristics,and the public will adapt their expectations about inflation accordingly. Cukiermanand Meltzer (1986) emphasize the importance of uncertainty about the underlyingpreferences of governors. A less conservative governor (dovish) has an incentive tomimic the behavior of the more conservative one (hawkish) for a while, but sooner orlater, it becomes optimal to behave opportunistically. Uncertainty can be expected tobe highest at the beginning of a governor’s tenure. In a similar vein, Schaumburg and


Tambalotti (2007) and Kara (2007) state that newly appointed central bank governorssuffer from a systematic credibility problem: while incumbent governors can committo policies during their own administration, their successors might deviate and pursuediscretionary policies. We therefore hypothesize the following:

Hypothesis 1. Investors react negatively to central bank governor changes due to asystematic credibility problem at the start of new governors’ tenure.

One way to remedy the inconsistency problem is proposed by Rogoff (1985) andentails the delegation of monetary policy to a conservative central bank. Eijffinger andHoeberichts (1998) and Berger, de Haan, and Eijffinger (2001) develop an argumentin the spirit of Rogoff. The government seeks to minimize the following loss function,representing the preferences of society:

LGov = 1

2π2

t + χ

2(yt − y∗

t )2,

where π t is the rate of inflation at day t, yt is output, y∗t denotes desired output, and

χ is the government’s weight on output stabilization (χ > 0). The loss function ofthe central banker is expected to differ from that of the government in one importantaspect:

LC B = 1 + ε

2π2

t + χ

2(yt − y∗

t )2,

where ε denotes the additional inflation aversion of the central bank governor, that is,his conservatism. Furthermore, Eijffinger and Hoeberichts (1998) argue that centralbankers’ preferences only matter to the extent to which they can pursue monetarypolicy without (much) government interference. This can be captured in the followingway:

Mt = γ LC B + (1 − γ ) LGov,

where γ and Mt denote the degree of central bank independence and monetarypolicy, respectively. Assuming that output is determined by a simplified Lucas supplyfunction and assuming rational expectations, inflation turns out to be:

πt = χy∗t − χ

χ + 1μt and πt = χ

1 + γ εy∗

t − χ

1 + γ ε + χμt .

The equation on the left (right) represents the inflation outcome without (with)delegation of monetary policy.3 Comparing these two outcomes, it becomes clearthat it is the product of central bank independence and conservatism that mattersfor monetary policy. We label this product “effective conservatism” of monetary

3. μt denotes a random shock with zero mean and variance σμ and is part of the simplified Lucassupply function.


policy. For positive values of γ and ε, any increase of either the degree of centralbank independence or the central bank governor’s conservatism will decrease theinflationary bias, ceteris paribus. The same level of monetary policy can be achievedthrough various combinations of the two dimensions.

From the above, changing the central bank governor might affect financial marketexpectations along these two dimensions. First, changes at the head of the central bankmay carry signals about the future stance of the incumbent government on the centralbank’s independence. If central bank governors’ resignations are politically motivatedand/or incoming governors lack political independence, financial markets are likelyto react negatively. Second, a newly appointed central bank governor’s attitude towardinflation might deviate from that of her predecessor. If the new governor is perceivedto be more conservative than the old one, market reactions should be positive. Fromthis, we derive the following two hypotheses:

Hypothesis 2. Investors react to central bank governor changes due to changes inperceived conservatism.

Hypothesis 3. Investors react to central bank governor changes due to changes inperceived central bank independence.

We employ various characteristics of outgoing and newly appointed governors andtheir nominating governments to distinguish between these two channels.4,5

Clearly, governor changes might affect exchange markets, stock markets, and bondmarkets to different degrees and even in different directions. We address each ofthese markets in turn. We implicitly assume that the semistrong form of the efficientmarket hypothesis holds. Under this hypothesis, security prices are assumed to reflectall public information and to adjust swiftly to the arrival of such information. In thisvein, if markets fully anticipate a change of governor, prices will not react on theofficial announcement day.

1.1 Foreign Exchange Markets

Following Kuttner and Posen (2010), our starting point for analyzing the impact ofcentral bank governor changes on the exchange rate is uncovered interest rate parity:

Et�et+1 = i∗t − it ,

4. In the working paper version of this paper, we also investigated whether markets react to personalcharacteristics of the incoming central bank governor. We test whether incoming governors educated inthe United States or United Kingdom are perceived to be more conservative and credible than thosewithout such education, since investors can better anticipate their preferences. We also analyzed incominggovernors with a history in their central bank (insiders) and those without (outsiders). Moreover, weinvestigated whether market reactions depend on the central bank’s degree of independence.

5. To the best of our knowledge, the only other study that seeks to disentangle the two Rogoffdimensions is Berger and Woitek (2005). They find that conservatism does matter for Germany, where theBundesbank enjoyed a virtually unchanged high degree of independence in the postwar period.


with e being the log of the foreign exchange rate (foreign currency/domestic cur-rency), i being the domestic interest rate, and i∗ being the foreign interest rate.Solving forward, we obtain

et = Et

[T −1∑s=0

(it+s − i∗t+s) + eT

],

where eT , the nominal exchange rate at some future date T , can be thought of asthe expected equilibrium exchange rate, determined by—for large enough values ofT—purchasing power parity. Hence, expected changes in monetary policy can affectthe foreign exchange rate either through expected changes in the nominal interest ratedifferentials and/or changes in the expected long-run exchange rate. For an increasein effective conservatism, that is, either an increase in central bank independence orconservatism, we expect the foreign exchange rate to appreciate. Alternatively, thedomestic currency might depreciate during a period of financial crisis (Caballero andKrishnamurthy 2001, 2002).

1.2 Domestic Stock Markets

The effect of a change in the expected monetary policy on stock markets is lessobvious than for exchange markets. Following Campbell et al. (1997), the stock pricecan be expressed as the expected value of future dividends (D) out to the infinitefuture, discounted at a constant rate (R).

Pt = Et

[ ∞∑i=1

(1

1 + R

)i

Dt+i

].

For this classic “Gordon growth model,” changes in monetary policy expectationscan affect stock prices through two different channels. Policy expectations mightsimply affect the discount rate or more subtly affect the expected future dividendstream. In summary, while the direct effect of higher expected inflation on stockprices is ambiguous, to the extent that it is associated with greater economic turmoil,investors may demand a higher risk premium.

1.3 Foreign-Currency-Denominated Bond Markets

The classical approach to model sovereign bond yields dates back to Edwards(1984), where the spread is denoted as a function of the probability of default (pd)and the risk-free interest rate (i*)

s = pd

1 − pd(1 + i∗).

Since our bond data are restricted to foreign-currency-denominated public or pub-licly guaranteed debt, bonds are not sensitive to changing inflation expectations andany possible transmission channel is of an indirect nature via the changing perception


of the probability of default. News that increases the sovereign’s perceived defaultprobability is expected to lower bond prices and, hence, increase sovereign bondspreads.6

2. DATA

Our analysis is based on several types of data. Our main selection criterion isthe availability of reliable daily data on the foreign exchange, and stock and bondmarkets for emerging countries. The resulting sample spans the period 1992–2006for the following 20 economies: Argentina, Brazil, Chile, China, Colombia, Egypt,Hungary, Malaysia, Mexico, Morocco, Pakistan, Peru, Philippines, Poland, Russia,South Africa, South Korea, Thailand, Turkey, and Venezuela. The number of obser-vations varies, however, across countries. The exact data availability and summarystatistics are provided in Appendices A and B.

A central component of our daily data set is the unique sample of exact announce-ment dates of changes of central bank governors, which we have identified through afull-text analysis of the Economist, the Wall Street Journal, and the Financial Times,using the database LexisNexis.7 Overall, this procedure yields 65 events, compris-ing 44 resignations and 21 appointments at the head of a central bank. Appendix Cpresents the number of events per country in our sample, with Argentina and Brazilshowing the greatest number of turnovers.8

We have cautiously sought to infer from newspaper articles whether and to whatextent the respective changes had been anticipated. If the change was largely an-ticipated by the markets, we would expect a softer reaction in market movements.While we can be confident about the hard facts, that is, the name and position of thegovernor and his date of departure, the soft facts about the surprise content should beinterpreted more cautiously. Appendix D gives a precise listing of the timing, nature,and reason of central bank governor changes. We complete our sample with data onthe partisanship of the nominating government, drawn from the Database of PoliticalInstitutions (see Beck et al. 2001).

Our sovereign bond spread data are denominated in U.S. dollars and drawn fromJ.P. Morgan. The Emerging Market Bond Index (EMBI) sovereign bond spreads are

6. We can think of two different channels along which central bank governor changes affect theperceived probability of default. First, if the change at the head of the central bank is interpreted as a signof political interference, investors will be expected to demand a higher yield. A less independent centralbanker (for any given level of conservatism) makes it more likely that fiscal policy ultimately dominatesmonetary policy, driving up public debt and the probability of default. Second, a more conservativeresponse to excess inflation raises (for given level of independence) expected real interest rates, reducingexpected investment and growth rates. This in turn will worsen the debt sustainability, increasing theprobability of default and, hence, the sovereign bond spread. Alternatively, one can argue once more inline with Caballero and Krishnamurthy (2001, 2002).

7. These three financial newspapers are backed by other press sources available through LexisNexis,where necessary. We also compared the dates of resignation with official sources, most of which are listedin the database of central bank governor turnovers provided in Dreher, Sturm, and de Haan (2010); seehttp://www.kof.ethz.ch/centralbankgovernors.

8. Neither the exclusion of Brazil nor Argentina would qualitatively change the results below.


calculated as the yield difference between the (basket of a country’s) emerging marketbond(s) and a comparable U.S. bond.9 These country indices are closely monitoredindicators for perceived country risk in emerging markets and can be interpretedas a default premium charged by investors above the risk-free interest rate. Onlysovereign bonds that comply with well-defined liquidity requirements are eligible forJ.P. Morgan’s bond indices.

Turning to stock markets, we use Morgan Stanley Capital International (MSCI)local market indices, obtained from Datastream. The MSCI data used here are dailyreturns of indices, excluding dividends, and measured in local currency. The in-dices measure market performance for selected securities, capturing the market-capitalization-weighted return of all constituents included.10

The exchange rate data is obtained from Bloomberg. Daily foreign exchange ratesvis-a-vis the U.S. dollar are employed, whereby an increasing foreign exchange ratemeans a depreciation of the domestic currency vis-a-vis the U.S. dollar.

We also employ a number of control variables. We control for U.S. financial marketindicators using the yield of 10-year U.S. Treasury bonds and 3-month U.S. T-bills.Both variables are widely used to control for international liquidity. Finally, we addthe Volatility Index (VIX) of the Chicago Board Options Exchange (CBOE). TheVIX measures the implied volatility from option contracts on the Standard & Poor’s100 (S&P 100) index and, hence, can be interpreted as a forward-looking indicatorof global risk aversion.

3. METHOD AND RESULTS

To test for the effects of central bank governor changes on financial markets, weemploy three different dependent variables: sovereign bond spreads, stock marketindices, and foreign exchange rates vis-a-vis the U.S. dollar.

We start by considering the average market reaction of our variables of interest (�y)to news announcements, following Kuttner and Posen (2010). As the volatility of ourdependent variables varies over time and country, we normalize �y by dividing it byits estimated standard deviation σ over the 90 days preceding the announcement. Thestatistic we use is thus zi ≡ (�yt )/σ . Under the null hypothesis that news regardingthe change of the central bank governor contains no relevant information, �y followsthe preannouncement distribution with zero mean and unit variance (Kuttner andPosen 2010). We test whether the average change in our normalized dependentvariables significantly differs from zero on days where the replacement of a governoris announced. We thus assume that the effect of a governor change materializes on that

9. Henceforth, the notion EMBI is used synonymously for EMBI, Emerging Markets Bond Index PlusEMBI+, and Emerging Markets Bond Global EMBIG. We mainly rely on the EMBI+ due to its largecoverage, its liquidity requirements, and its up-to-date record. Bond spread data from the early 1990s areobtained from EMBI. For Chile and Uruguay, only EMBIG data are available.

10. Note that the returns on MSCI and the respective country index are highly correlated (Pantzalis,Stangeland, and Turtle 2000).


TABLE 1

AVERAGE MARKET REACTION TO EVENTS

Sample Exchange rate Bond yield Stock price

CB resignation (all) Avg. change 0.493 0.297 −0.225N 41 44 44p-value 0.001 0.049 0.136

CB resignation (irregular) Avg. change 0.770 0.248 −0.272N 32 34 34p-value 0.000 0.148 0.113

CB resignation (regular) Avg. change −0.493 0.462 −0.063N 9 10 10p-value 0.139 0.144 0.842

Partisanship Avg. change 0.797 0.097 −0.294N 8 9 9p-value 0.024 0.771 0.378

CB appointment Avg. change 3.217 −0.263 0.766N 20 21 21p-value 0.000 0.228 0.000

NOTES: The table evaluates whether the mean change of our variables of interest (�y) over the sample period, normalized by dividingit by its estimated standard deviation σ over the 90 days preceding the announcement, equals their mean change on the event days. Wetest whether the average change in our normalized dependent variables significantly differs from zero on days where the replacement of agovernor is announced. CB resignation refers to the announcement day of the resignation of a central bank governor, whereby a resignationis coded as irregular if the central bank governor change occurred before the expiration of the central bank’s tenure and as regular otherwise.Partisanship indicates that the current nominating government is less conservative than the government that nominated the previous governor.CB appointment denotes the announcement day of the appointment of a central bank governor.

same day. The average dependent variables are approximately normally distributedwith variance 1/N, where N is the number of events in our sample.11

Table 1 shows the value of the z-statistics, indicating the respective levels ofsignificance. As can be seen, the results show some interesting patterns. Resignationsincrease bond spreads at the 5% level of significance and lead to an exchange ratedepreciation at the 1% level.12 Relating these results to our hypotheses, investors doindeed react to central bank governor changes. In line with Hypothesis 1, the averagemarket reaction to the announcement of the resignation of the central bank governor isnegative in two out of the three financial markets. On average, emerging markets thusseem to react differently than markets in advanced countries. While Kuttner and Posen(2010) report that markets do react significantly to governor changes for their sampleof 15 industrialized countries, their results do not follow a unidirectional pattern:domestic-currency-denominated bond yields do not, on average, rise, nor does theexchange rate depreciate, in response to the announcement of a new governor. Kuttnerand Posen thus reject the widely held belief that markets might react due to a lack ofcredibility at the start of a governor’s tenure. This is contrary to our own results.

The appointments of new governors do not affect bond spreads significantly butlead, on average, to a depreciation of the exchange rate and an increase in stock prices

11. In the working paper version of this paper, we provide two further simple methods, which producequalitatively similar results.

12. In a test for robustness, we increased the event period to include 1 day prior and 1 day after theannouncement day. Our results are sensitive to this change to some extent. While our findings on theexchange rate effects remain very similar, bond spreads are no longer significantly affected.


(at the 1% level of significance).13 The stock market reaction can be interpreted in linewith Brown, Harlow, and Tinic (1988) who show that the resolution of uncertaintyleads to positive market reactions, on average. Once the successor is announced,pending uncertainty about the new head of the central bank is resolved.

What about the impact of effective conservatism on financial markets? Turning tothe first dimension of effective conservatism, we draw on the Database of PoliticalInstitutions (Beck et al. 2001) to construct a proxy for perceived conservatism. Thedatabase classifies the partisanship of an incumbent government as left, center, orright. Our variable takes a value of 1 when partisanship of the government nominat-ing the current governor is to the left of the government who nominated the previousgovernor, –1 when the nominating government is more conservative than the previousone, and 0 if the government’s degree of conservatism remains unchanged. In con-structing this proxy for the central bank governor’s conservatism, we follow Bergerand Woitek (2005), who assume that right-wing governments tend to nominate moreinflation-averse central bank governors, and vice versa. Arguably, the assumptionthat the nominating governments’ political color corresponds to those of the gover-nor is debatable. We therefore checked our coding by employing full-text analysisin LexisNexis. While exclusive reliance on LexisNexis would reduce the number ofevents and would not allow meaningful econometric analysis, those cases where wecould identify the relative conservatism of an incoming governor largely confirm thevalidity of our coding (see Appendix E).

Table 1 reports results for a sample of resignations where the government nomi-nating the incoming governor has a different political partisanship than the one thatappointed the previous governor. More specifically, the table shows market reactionswhen central bank governors are appointed by less conservative governors, as de-fined above. As argued in Hypothesis 2, markets react to changes in the governors’perceived conservatism. When a governor resigns and the incumbent government onthe announcement day is less conservative than the one that nominated the outgo-ing governor, markets should thus react negatively. As can be seen, we find such anegative reaction in favor of Hypothesis 2 only for the inflation sensitive exchangerate. Note, however, that the very small number of (a maximum of nine) observationsmakes this finding rather tentative.

We also gain insight into Hypothesis 3 by separating regular from irregular res-ignations. While irregular events apparently drive the negative announcement effecton the exchange rate, the effects are marginally insignificant for the bond and stockmarket. We do not find evidence for significant market reactions to regular resig-nations. If markets were primarily concerned about the incoming governor lackingcredibility, we would expect regular and irregular events to be equally likely to affectmarkets. If, however, central bank independence is perceived to be at risk, irregularchanges of central bank governors should drive the results. Hence, irregular events

13. Note, however, that the very small sample size might drive the insignificant effect of appointmentson bond spreads.


can be expected to alter perceived central bank independence, which is one of thetwo key determinants of inflationary bias.14

Arguably, stock markets and U.S. dollar-denominated sovereign bonds are ratherinsensitive to changes in expected inflation. This might explain why we cannot findevidence in favor of Hypothesis 3 for these financial markets. However, note onceagain that these conclusions rest on a very small number of observations, and are thusonly tentative.

As a next step, we conduct panel data analysis. Our baseline regression is estimatedby pooled ordinary least squares (OLS) with robust standard errors clustered at thecountry level15 and takes the following form:

�Yi,t = α + λ�Yi,t−1 + βRESIGNi,t + γ APPOINTi,t + η�Xi,t

+ νw Dw + εi,t , (1)

where the subscripts i and t indicate country and time, respectively. Yi,t representsthe respective dependent variables denoted in log-differences. The one-period laggedvalue of the dependent variable also enters the equation as a covariate. Our coefficientsof interest are β and γ , accounting for the impact of resignations, RESIGNi,t, and newappointments, APPOINTi,t, at the head of the central bank.16 The two announcementvariables are 1 on the day of the change and 0 otherwise. We denote the error termεi,t and employ dummy variables Dw, running from Monday to Thursday, in order tocontrol for day-of-the-week effects.

The matrix Xi,t includes the three U.S. financial market indicators introduced above.We employ (log) changes of the volatility index, and (log) changes in 10-year U.S.Treasury bond yields and 3-month U.S. T-bill yields, defined as 100 × log (1 + it,US).

By definition, fixed exchange rate regimes do not allow for daily market reactionsto the announcement of the turnover. For this reason, we interact the resignation orappointment variable with a dummy variable that takes the value of 1 when there is aflexible exchange rate (or at least not a fully pegged exchange rate) and 0 otherwise.17

Finally, as one potential caveat to the analysis, central bank governors might bedismissed as a consequence of economic crises, giving rise to endogeneity. Whengovernors are dismissed due to economic shocks, market reactions might reflect theseshocks rather than the exogenous change in who governs the central bank. While thisargument appears reasonable for quarterly or yearly data, endogeneity is unlikelyto be an issue when data frequency is daily, as is the case in our study. Even if

14. In coding the dummy for irregular changes, we follow Kuttner and Posen (2010) and also classifycases in which the incumbent governor was eligible for reappointment, but did not receive it, as irregular.Note that these kinds of irregular changes constitute a very small number of all irregular resignations.

15. We do not include fixed country effects as they are not jointly significant at conventional levels.Our key results are not changed by their exclusion.

16. On average, the new central bank governor is announced 10 days after the resignation of hispredecessor. We only measure appointments when they constitute an independent event.

17. We code an exchange rate regime as flexible as long as at least some market reactions are observablein the weeks around the day of the event.


TABLE 2

FINANCIAL MARKETS AND GOVERNOR CHANGES, 20 COUNTRIES, 1992–2006

(1) (2) (3) (4) (5)Stock price Stock price IV Exchange rate Bond yield Bond yield IV

� log y, lagged 0.1062∗∗∗ 0.1854∗∗∗ 0.0249 −0.1925∗∗∗ 0.5765∗∗∗(6.55) (3.12) (0.53) (3.38) (14.23)

CB resignation −0.0042∗ −0.0045∗∗ 0.0097∗∗∗ 0.0047 0.0152∗∗∗(1.96) (2.23) (2.89) (1.43) (4.23)

CB appointment 0.0140 0.0129 −0.0073 0.0002 −0.0086(1.40) (1.36) (0.72) (0.02) (0.83)

� log volatility index (VIX) −0.0534∗∗∗ −0.0538∗∗∗ 0.0053∗∗ 0.0825∗∗∗ 0.0801∗∗∗(4.66) (4.70) (2.62) (5.45) (5.44)

� log U.S. T-bond 10 years 0.0682∗∗∗ 0.0658∗∗ 0.0176∗ −0.9960∗∗∗ −0.9299∗∗∗(3.94) (3.73) (1.97) (4.09) (3.91)

� log U.S. T-bill 3 months 0.0158 0.0167 −0.0043 −0.1355 −0.2050∗∗(1.01) (1.01) (0.77) (1.71) (2.22)

Observations 51,423 51,422 51,423 51,399 51,375R2 0.04 0.03 0.003 0.07 —

NOTES: The dependent variable is the (log) change in y. Results are based on clustered standard errors. (absolute) values of t-statistics inparentheses. Week-day effects and a constant are estimated but not reported. The instrumental variable (IV) estimation in columns (2) and(5) uses the second lag of the dependent variable as instrument for the lagged dependent variable. Testing for first-order autocorrelation inthe error term indicates no first-order auto correlation.∗∗∗ , ∗∗ , ∗denote 1%, 5%, 10% level of significance.

the governor is fired as a consequence of macroeconomic crisis, such crises usuallyunfold over a longer period of time, so that daily data can still be used to identify thecausal impact of the turnover itself on market reactions. Endogeneity is thus unlikelyto be an issue here. Still, the effect of central bank governor turnovers might bedifferent during times of crisis, a matter to which we return below.

Table 2 reports the results of the panel data analysis. Column (1) reports thebase specification while the potential bias introduced by the correlation between thelagged dependent variable and the error term is addressed in column (2). We thereforeinstrument the (highly significant) lagged dependent variable with its second lag(Anderson and Hsiao 1982). Our results are hardly changed by this. Column (3)shows the results for the exchange rate. Finally, results for bond spreads are reportedin columns (4) and (5). The lagged dependent variable is completely insignificantin column (3) and we therefore do not instrument it. While the lagged dependentvariable is highly significant according to the OLS specification of column (4), witha negative coefficient, it becomes significantly positive once instrumented (column(5)).

Turning to our variables of interest, Table 2 again shows some support for our hy-potheses and, specifically, that financial markets do react to the resignation of centralbankers. By contrast, the announcement of the appointment of a governor has nosignificant impact. Overall, the results regarding appointments are thus weaker thanthose reported earlier in Table 1. We base our conclusions on the more stringent re-gression analysis. As can be seen in the table, domestic stock markets react negativelyto central bank governor resignations. The estimated coefficients in columns (1) and


(2) imply a small decline in returns of about 0.5 percentage points. While being farfrom a dramatic crash of markets, this effect is, arguably, economically relevant. Thestock market announcement effect is about five times higher than the one triggeredby changes in sovereign ratings and sovereign rating outlooks in emerging markets(Kaminsky and Schmukler 2002).18

Column (3) shows that the resignation of a central banker leads to a depreciation ofthe exchange rate, at the 1% level of significance.19 The estimated coefficient impliesa depreciation of almost 1 percentage point following the resignation of a centralbanker. Finally, columns (4) and (5) show that bond spreads do not increase followingthe resignation of the head of the central bank according to the OLS regressions butdo increase (at the 1% level of significance) once the lagged dependent variable isinstrumented. According to the coefficient, the resignation of a central bank governorincreases bond spreads by more than 1.5 percentage points.

Based on our results, we cannot reject Hypothesis 1, stressing the credibility prob-lem at the start of a governor’s tenure. As potential objection to this interpretation, thenegative market reactions might, of course, simply reflect the presence of a financialcrisis, indicated by the resignation of the governor. We therefore include interactionterms between our proxies for crises and changes of central bank governors. Thefollowing specification accounts for financial crises:

�Yi,t = α + λ�Yi,t−1 + β1RESIGNi,t + β2RESIGNi,t ∗ CRISISi,t

+ γ1APPOINTi,t + γ2APPOINTi,t ∗ CRISISi,t + η�Xi,t + νw Dw

+φCRISISi,t + εi,t , (2)

where our variable CRISISi,t takes the value of 1 in case of an ongoing currencycrisis in country i in month t and 0 otherwise. We follow Kaminsky and Reinhart(1999) and compute a foreign exchange market pressure index as an average ofthe rate of change of the exchange rate and of international reserves, weighted bytheir standard deviations.20 Following Dreher, Herz, and Karb (2006), a country isdefined to be in a currency crisis when its index deviates from the index mean for thatcountry by at least one standard deviation. According to this definition, 11 centralbankers from nine countries have been dismissed during a currency crisis in oursample.

Table 3 shows the results. As can be seen, currency crises never enter significantlyin any of our regressions. Similarly, the interaction of crises and dismissals is notsignificant at conventional levels in any specification, and with the exception of

18. This comparison can only be made for the U.S. dollar-denominated MSCI index (results notshown).

19. Note that this result remains when the insignificant lagged dependent variable is omitted from theregression.

20. Ideally, we would also like to include interest rates that monetary authorities can use to defendparities. However, this leaves us with a drastically reduced sample, and so, following Kaminsky andReinhart (1999), we do not include interest rates.


TABLE 3

FINANCIAL MARKETS, GOVERNOR CHANGES, AND FINANCIAL CRISES, 20 COUNTRIES, 1992–2006

(1) (2) (3) (4) (5)Stock price Stock price IV Exchange rate Bond yield Bond yield IV

� log y, lagged 0.1060∗∗∗ 0.1851∗∗∗ 0.0221 −0.1926∗∗∗ 0.5766∗∗∗(6.52) (3.13) (0.52) (3.38) (14.2)

CB resignation [1] −0.0051∗ −0.0054∗∗ 0.0090∗∗ 0.0051 0.0136∗∗∗(1.97) (2.23) (2.54) (1.44) (4.19)

CB resignation (crisis) [2] 0.0046 0.0046 0.0044 −0.0027 0.0089(0.82) (0.85) (0.37) (0.44) (0.59)

CB appointment [3] 0.0080 0.0075 0.0048 −0.0119 −0.0172(1.68) (1.50) (1.51) (1.12) (1.51)

CB appointment (crisis) [4] 0.0314 0.0284 −0.0607 0.0635∗∗ 0.0451∗(0.68) (0.65) (1.46) (2.75) (1.88)

� log volatility index (VIX) −0.0534∗∗∗ −0.0538∗∗∗ 0.0053∗∗ 0.0824∗∗∗ 0.0801∗∗∗(4.67) (4.70) (2.64) (5.44) (5.44)

� log U.S. T-bond 10 years 0.0682∗∗∗ 0.0658∗∗∗ 0.0176∗ −0.9961∗∗∗ −0.9299∗∗∗(3.93) (3.73) (1.97) (4.09) (3.91)

� log U.S. T-bill 3 months 0.0156 0.0165 −0.0042 −0.1357 −0.2051∗∗(1.01) (1.00) (0.76) (1.71) (2.22)

Currency crisis period −0.0001 −0.0001 −0.0003 −0.0002 −0.0003(1.01) (0.40) (1.49) (0.33) (1.14)

Joint significance [1] + [2] 0.16 0.10 0.03 0.36 0.00Joint significance [3] + [4] 0.19 0.25 0.16 0.04 0.16Observations 51,423 51,422 51,423 51,399 51,375R2 0.04 0.03 0.006 0.07 —

NOTES: The dependent variable is the (log) change in y. Results are based on clustered robust standard errors. The (absolute) values oft-statistics are in parentheses. Week-day effects and a constant are estimated but not reported. The instrumental variable (IV) estimation incolumns (2) and (5) uses the second lag of the dependent variable as instrument for the lagged dependent variable. The coefficients [2] and [4]represent the respective interaction with our crisis dummy. Testing for first-order autocorrelation. The in the error term indicates no first-ordercorrelation. p-values for test of joint significance are reported.∗∗∗ , ∗∗ , ∗denote 1%, 5%, 10% level of significance.

the bond market, the same is true regarding the interaction of appointments andcrises. We actually do find some statistically weak evidence that appointments areperceived more negatively by sovereign bond holders during currency crises than intranquil times. Note, however, that while the point estimates for the stock market,foreign exchange, and bond markets marginally change when we control for financialcrises, overall, there is neither evidence for an increased sensitivity to central bankturnovers in crisis times nor for an additional risk premium following turnovers duringcrises.21

Coming back to Hypothesis 1, we conclude that our results do not reflect a riskpremium. However, the negative market reactions are also in line with a decreasein perceived effective conservatism. Given the results in Table 2, we cannot knowwhether Hypothesis 1 or, alternatively, Hypotheses 2 or 3 hold true. In order to

21. Note that the coefficient on central bank appointments is marginally insignificant in column (1).This is even though the t-statistic of 1.68 with more than 50,000 observations might suggest otherwise.However, the standard errors are clustered at the country level, so the number of clusters (M) becomesthe relevant dimension to determine the degrees of freedom. The confidence values Stata 11 reports whenclustering standard errors are based on (more conservative) critical values from a Student’s t-distributionwith M–1 degrees of freedom.


disentangle the two dimensions of “effective conservatism” we estimate the followingequation:

�Yi,t = α + λ�Yi,t−1 + β1RESIGNi,t + β2RESIGNi,t ∗ PARTISANi,t

+β3RESIGNi,t ∗ IRREGi,t + γ APPOINTi,t + η�Xi,t

+ νw Dw + εi,t , (3)

where PARTISANi,t is our proxy for the governor’s perceived conservatism describedabove and IRREGi,t is a dummy variable that takes the value of 1 when the centralbank governor change occurred before the expiration of the central bank governor’stenure and 0 otherwise.22 According to the results (not shown in a table), almostno coefficient of interest is significant at conventional levels. There is one importantexception, however. Irregular turnovers lead to a depreciation of the exchange rate,with a coefficient significant at the 5% level. This provides some evidence thatinvestors are indeed worried about central bank independence when central bankgovernors resign before the end of their tenure.

4. CONCLUSION

Central bank governor changes in emerging markets may convey important signalsabout future monetary policy. Based on a new daily data set on 20 countries over the1992–2006 period, this paper has examined the reactions of foreign exchange markets,domestic stock market indices, and sovereign bond spreads to the announcement ofa central bank governor change.

Our results show, first, that the resignation of a central bank governor negativelyaffects financial markets on the announcement day, with average market reactionsbetween 0.5 and 1.5 percentage points. By contrast, we find little evidence thatappointments of new governors convey relevant news to investors.

Second, comparing our results with those found in the existing literature, we findthat our results for emerging market economies are distinct from industrialized coun-tries in an interesting and important way. Newly appointed central bank governorsapparently suffer from a systematic credibility problem at the beginning of theirtenure. In contrast to their counterparts in industrialized countries, emerging marketgovernors initially have to face (at least) a transitory rise in inflation expectations,because investors are uncertain about the true type of the central bank governor(hawkish versus dovish).

Third, the negative announcement effect for resignations is mainly driven by irreg-ular changes, that is, changes occurring before the scheduled end of tenure. Foreign

22. Note that we cannot include PARTISAN and IRREG individually, as they, by definition, can onlybe observed when the governor changes. We do not interact the appointment dummy with changes inpartisanship, since the incumbent government’s partisanship, and hence potential changes in ideologyvis-a-vis the government nominating the previous central bank governor, are already priced in on theresignation day.


exchange market participants are apparently sensitive to signals about perceivedcentral bank independence, expecting higher inflationary bias.

Finally, we find little evidence that the governor’s degree of conservatism mattersfor market reactions. Still, this lack of significance might well be due to the verylimited number of observations for changes in conservatism.

Overall, our study complements Santiso (2003), pointing out that key policymakersin emerging markets are crucial for building credibility in international financialmarkets in one important aspect. Investors are apparently sensitive to the way inwhich an incumbent government handles the replacement of key policymakers. Withrespect to central bank governor changes, investors seem to care most about perceivedcentral bank independence. Future research to further disentangle the two dimensionsof effective conservatism is clearly desirable.


APPENDIX A: DATA AVAILABILITY, EMBI (G), FX TO U.S. DOLLAR, ANDMSCI AVAILABLE

Country Start End

Argentina April 30, 1993 July 31, 2006Brazil January 15, 1992 July 31, 2006Chile May 28, 1999 July 31, 2006China December 31, 1997 July 31, 2006Colombia December 31, 1997 July 31, 2006Egypt July 31, 2001 July 31, 2006Hungary January 1, 1999 July 31, 2006Malaysia December 31, 1997 July 31, 2006Mexico December 31, 1991 July 31, 2006Morocco December 31, 1997 July 31, 2006Pakistan June 29, 2001 July 31, 2006Peru May 30, 1997 July 31, 2006Philippines December 31, 1991 July 31, 2006Poland December 31, 1997 July 31, 2006Russia December 31, 1997 July 31, 2006South Africa December 31, 1997 July 31, 2006South Korea December 31, 1997 July 31, 2006Thailand December 31, 1997 March 31, 2006Turkey December 31, 1997 July 31, 2006Venezuela December 31, 1992 July 31, 2006

APPENDIX B: DESCRIPTIVE STATISTICS

Variable Mean Std. dev. Min Max

(log) Spread 5.68 1.07 0.00 8.88(log) MSCI 5.53 1.06 2.55 8.61(log) Exchange rate 3.09 −2.44 7.82 8.00Central banker change 0.00 0.03 0.00 1.00(log) VIX 2.91 0.32 2.23 3.90(log) U.S. T-bond 10 years 1.94 0.22 1.41 2.35(log) U.S. T-bill 3 months 1.62 0.42 0.59 2.31

APPENDIX C: NUMBER OF CENTRAL BANK GOVERNOR CHANGES BYCOUNTRY

Country Number events Country Number events

Argentina 6 (3) Pakistan 1 (1)Brazil 9 (2) Peru 4 (3)Chile 1 (1) Philippines 2 (1)China 1 (0) Poland 1 (2)Colombia 1 (0) Russia 2 (1)Egypt 2 (1) South Africa 0 (1)Hungary 1 (1) South Korea 2 (0)Malaysia 2 (1) Thailand 2 (0)Mexico 1 (0) Turkey 2 (2)Morocco 1 (0) Venezuela 3 (1)

NOTE: This table reports the number of central bank governor resignations (appointments). Our data are for the period of 1992–2006.

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APPENDIX E: CENTRAL BANK GOVERNORS AND CONSERVATISM—SOMEEXAMPLES FROM THE PRESS

Country Date Name

Argentina 2001 Roque Maccarone President De La Rua (of the Alliance forWork, Justice and Education, which isconsidered to be a center party) dismissedPedro Pou on alleged bad administration in2001. The incoming Roque Maccarone wasindeed considered to be less conservativethan Pou (nominated by the conservativePartido Justicialista), as illustrated in TheMiami Herald (April 27, 2001, p. 2):“[Minister of the Economy] Cavallopublicly blamed the current 33-monthrecession on Pou’s failure to loosen liquidityrequirements for banks, saying the policyprevented businesses from obtainingmuch-needed credit. . . . Maccarone said hewill consider changing the level of reservescommercial banks are required to keep ondeposit with the Central Bank.”

Argentina 2002 Mario Blejer Maccarone suddenly resigned in 2002 andMario Blejer became his successor,appointed by the more conservative PartidoJusticialista. Again, the change in ideologyis reflected in the news. According to TheNew York Times (January 18, 2002, p. 1)“Mr. Maccarone’s replacement, MarioBlejer, is a former International MonetaryFund official with a good reputation amonginternational lenders.”

South Africa 1998 Tito Mboweni Another example that fits our coding is theappointment of Tito Mboweni in SouthAfrica in 1998. Being appointed by theAfrican National Congress—defining itselfas force of the left—he is coded to be lessconservative than his predecessor ChristianStals, who was nominated by theconservative National Party in 1989.According to the Guardian (July 7, 1998, p.12) Mboweni was perceived to be “generousto a fault to the unions. He has been heavilydependent on left-wing advisers at the laborministry.”

Turkey 2006 Durmus Yilmaz In Turkey, Durmus Yilmaz followed SureyyaSerdengecti in 2006. Serdengecti had beenappointed by the Democratic Left Party,while Yilmaz was appointed by the moreconservative Justice and Development Party.Consequently, Yilmaz is coded as moreconservative than Serdengecti. According tothe Washington Post (April 19, 2006, p. 5),Yilmaz was indeed perceived to beconservative at the time of his appointment:“Yilmaz’s appointment was seen as anassurance that the bank would not deviatefrom its tight monetary policy.”


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Documents

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