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Published in: Public Choice 3-4/2021

02-01-2020

The limits of central bank independence for inflation performance

Author: Jamus Jerome Lim

Published in: Public Choice | Issue 3-4/2021

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Abstract

The independence of the central bank is routinely regarded as sacrosanct, at least for governments wishing to maintain credible monetary policy to meet inflation objectives. Yet empirical efforts to ascertain that routine economic policy advice are complicated by the endogeneity of inflation and independence. Using a large panel of up to 147 economies between 1970 and 2012, we revisit the claim that central bank independence leads to superior inflation outcomes from the perspective of democratic governance. We deploy a measure of the degree of democratic representation as an instrument for the independence of the monetary authority and obtain estimates of the causal effect of central bank independence on inflation. Our baseline results overturn the standard negative inflation-independence relationship. Further inquiry into parameter heterogeneity indicates that the result is driven by developing economies and is attributable to political-economy factors: insufficient transparency enables the pursuit of non-price-stabilization objectives.

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Footnotes
1
What ties virtually all empirical tests of the three-way relationship has been the use of interaction terms (between independence and democracy) to assess their joint effect on inflation.
 
2
This is an empirical claim: among the 2338 country-year observations in our data where countries are classified as democracies (possessing Polity scores exceeding 5), the weighted central bank independence score is 1.7, compared to a score of 1.5 among the 1034 instances of countries classified as autocracies (Polity score less than 5). In Sect. 4.2, we provide additional evidence of the relevance of democracy as an instrument.
 
3
Those external factors include the amount of domestic slack and private agents’ expectations regarding price formation (Rudd and Whelan 2007), the outstanding burden of public debt (Leeper 1991; Woodford 1995), structural changes in population demographics (Bobeica et al. 2017), and the pass-through from fluctuations in international intermediate input prices (Choi et al. 2018).
 
4
Cukierman (2008) also emphasizes a number of regional motives underlying the move, such as the breakdown of the European Monetary System (which shifted toward mimicking the operational structure of the successful and highly independent Bundesbank), the stabilization of inflation in Latin America (which prompted the search for credible, anti-inflationary institutions), and the end of the Cold War (which led former socialist countries to adopt best-practice, Western-type monetary institutions).
 
5
We also consider system GMM models, applied to five-year averaged data, as robustness checks; in the Online Appendix, we replicate the fixed-effects benchmark that is common to the literature.
 
6
A third source of endogeneity is the problem of measurement error in the key explanatory variable, independence. Such error may be resolved by IV methods, albeit only to the extent that we believe that the instrument itself is subject to less systematic bias relative to the endogenous variable. While that belief is disputable for the case of democracy relative to independence, our strategy of considering alternative democracy measures in the robustness checks serves as an indirect control for the possibility that measurement error may be attenuated by our instrument.
 
7
More formally, the bias in an ordinary least squares (OLS) estimator of \(\beta\) arise when, for instance, reverse causality takes the form \(CBI=\chi \pi +\nu\), with \(\nu \sim N\left( 0,\sigma _{\nu }^{2}\right)\) being the idiosyncratic error term, the direction of bias is given by \({{\,\mathrm{sgn}\,}}\left[ {\chi /\left( {1-\beta \chi }\right) }\right]\). Since \(\chi <0\), so long as \(\beta \chi <1\) (which holds empirically in our case), the bias is negative.
 
8
For example, claims that democracies may pursue spending policies funded by an inflation tax—along the lines of Calvo (1978)—require that government debt be monetized, which can occur only when effected by a pliant monetary authority. Similarly, models wherein inflation results from a political business cycle (Alesina 1987, fn. 9) acknowledge that the party in power can influence inflation only when the central bank is not completely independent. Even in modern economies, where the money supply is led by commercial banks extending loans, the amount of money circulating in the economy ultimately is a decision of the central bank. Indeed, the delegation of currency issuance to a monetary authority in all modern economies ensures, almost by definition, that policy-induced price inflation necessarily must operate through a central bank (McLeay et al. 2014).
 
9
The same weak relationship also holds in more sophisticated statistical analysis: a bivariate fixed-effect regression, including time and country fixed effects, yields a similarly small and marginally significant coefficient (\({\hat{\delta }}=0.03\), \(p=0.05\)).
 
10
The transformation has the advantage that it does not exclude instances of deflation in the data, which is important for our understanding of inflation dynamics.
 
11
The traditional definition of hyperinflations has been an increase in prices of more than 50% per month. Applying that figure would amount to an annual inflation rate of more than 12,000%, a condition that would fully be met by only one observation in our sample (Cote d’Ivoire in 1994, where inflation topped 23,773%) and marginally by another (Bolivia in 1985, when inflation reached 11,750%).
 
12
Since such constraints capture the ability of independent branches of government to perform a check and balance function on policy, they serve as measures of how an economy with such processes in place are more likely to confer policy independence on the monetary authority. In practice, the two measures (of democracy and constraints) are correlated strongly (\(\rho =0.83,p=0.00\)). It is important to note as well that doing so does not mean that we are conditioning independence on constraints (as other authors, such as Moser (1999), do), but rather that we are exploiting how countries with more checks and balances in effect simultaneously tend to afford their central banks more formal independence (and, hence, is a relevant instrument), even if the constraints otherwise do not have any direct influence on inflation (satisfying the exclusion restriction).
 
13
Technically, meaning that the host of regressions we run essentially are 2SLS rather than IV specifications, since the exclusion of the exogenous variables would lead to bias in the estimates (Baltagi 2011). Following convention, however, we continue to label such specifications with democracy explicitly accounted for as IV estimates and retain the label 2SLS for those that include purely exogenous variables in the first stage.
 
14
The principle underlying this measure, which is formally defined the online appendix, is to capture the feasibility of policy change (based on the number of branches of government holding veto power over such change).
 
15
In the final two columns—where the effect remains positive but is estimated imprecisely—the instrument no longer passes the underidentification test and robust inference is threatened by Stock-Wright S statistics that are either only marginally significant or insignificant at the 10% level. Accordingly, we place greater weight on the specifications with controls that satisfy the relevance condition, meet the criteria for strong instruments, and for which inference is robust to weak instrumentation; those are reported in columns (I2) and (I4), which we apply to the robustness checks in Sect. 5.3.
 
16
The elasticity for an arcsinh-logarithm specification is given by \({\hat{\zeta }}_{\pi ,CBI}=\widehat{\frac{\partial \pi }{\partial CBI}}\cdot \frac{CBI}{\pi }=\widehat{\beta _{F}}\frac{\cosh \left( {{{\,\mathrm{arcsinh}\,}}({\hat{\pi }})}\right) }{\pi }=\widehat{\beta _{F}}\cdot \frac{\sqrt{1+\pi ^{2}}}{\pi }\). Since \(\lim _{\pi \rightarrow \infty }\frac{\sqrt{1+\pi ^{2}}}{\pi }=1\), for \(\pi\) sufficiently large, \({\hat{\zeta }}_{\pi ,CBI}\approx \widehat{\beta _{F}}\).
 
17
In the Online Appendix, we also document a number of historical cases wherein independence did not result in lowered inflation outcomes.
 
18
To be clear, not all studies that address the endogeneity issue and focus on developing economies overturn the usual outcome. Crowe and Meade (2008) apply IV to a small cross-sectional sample and recover the typical negative effect, while Jácome and Vázquez (2008) likewise uncover a negative relationship using a sample of developing Latin American economies.
 
19
The coefficient on (C5) also approaches marginal significance (\(p=0.119\)).
 
20
The premise behind doing so stems from the notion that, under partisan theories (Alesina 1987), inflation tends to be higher under left-wing than under right-wing governments (for a survey of the empirical evidence, see Potrafke (2018)). Given the heterogeneity of political systems in our sample, we enter political constraints here as a proxy for the extent of partisan divergence, instead of a direct left-right partisanship index.
 
21
Cahan et al. (2019), for example, find that independent central banks were more likely to be interest-rate sensitive when left-leaning governments are in office.
 
22
In addition to Bodea and Hicks (2015), a number of alternative independence measures have been coded, although all follow the same fundamental principles outlined in Cukierman et al. (1992). The Bodea-Hicks sample is the largest of them, however, which is why we utilize it as a robustness check.
 
23
One possible critique is that it is unfair to rely directly on the level of inflation, but to instead use inflation relative to official targets as a more accurate metric of performance. The problem with such an approach is that inflation targeting is a relatively recent phenomenon adopted by a comparatively limited set of central banks. Nevertheless, as an unreported robustness check, we also consider a specification with deviations (from the time targets were respectively adopted) as the dependent variable. As expected, the sample size shrinks considerably, to only 656 observations. Even so, the coefficient remains unchanged—in fact, it is much larger in magnitude—although, unsurprisingly, the coefficient drops out of statistical significance.
 
24
We cannot utilize the exchange-rate regime directly as an instrument since it is obviously endogenous to independence. Unsurprisingly, regressions that do include the regime in the first stage end up failing the overidentification test, rendering the approach invalid (these additional results are available on request).
 
25
It is worth noting that not all studies make the same case. Oatley (1999), for example, finds remarkable robustness to the inclusion of covariates.
 
26
Our main split adopts the World Bank’s definition of what constitutes high-income versus developing economies, for the year 2012 (our final sample year). The definition corresponds to per capita income thresholds of $1035 for low, $4085 for lower-middle, $12,615 for upper-middle, and $12,615 for high.
 
27
The other results are available on request.
 
28
The results likely are explained by the fact that democratic development does not differ that much between advanced economies; hence, our otherwise-reliable instrument loses the necessary variation needed for adequate identification.
 
29
One reason why advanced economies may exhibit a negative relationship between independence and inflation is because they also turn out to be home to the central banks that are more credible and/or transparent. We probe those possibilities in Sect. 6.2, where we consider channels of transmission.
 
30
That interpretationis easily verified by comparing the simple means of democracy or political constraints by income group, which situates middle-income economies between those of high- and low-incomes.
 
31
It may be argued that, prior to the 1980s, central banks may not have had price stability as their primary objective. Regressions that split the sample into pre- and post-1980s periods reveal, perhaps ironically, that the effects of independence on inflation in before the 1980s period actually are negative (while remaining positive in after the 1980s). That result is is likely explained by the predominance of high- and upper-middle income economies in the subsample, relative to lower-middle income countries, as well as to the small sample size (501 observations). The results are available on request.
 
32
Blinder et al. (2017) have suggested that central bank independence may have been threatened as a result of the global financial crisis, with the effects especially pertinent to the high-income countries that were most afflicted by it. We take the suggestion seriously and repeat our analysis for the pre- and post-crisis period. The results (available on request) indicate that the positive independence-inflation effects are indeed stronger prior to the crisis, with the coefficient becoming negative (but insignificant) after 2008. Given the small size of the latter subsample, and that the sample necessarily excludes the history of hyperinflations, it is unclear how much stock to place on this result. Still, one may speculate as to whether institutional developments since 2008 are now sufficiently advanced that independent central banking may now, belatedly, be fulfilling its promise.
 
33
No universally agreed standard for measuring financial stability is available. As an alternative, we consider changes in the nominal exchange rate, which aligns with a “fear of floating” typical among emerging economies (Calvo and Reinhart 2002) and is often included in alternative specifications of the Taylor rule that consider deviations other than the output gap. Doing so results in a likewise insignificant coefficient on independence; the results are available on request.
 
34
We recognize that using either output or credit growth may introduce endogeneity issues that could invalidate our identification strategy—in particular, the exclusion restriction—which we do not deal with here. Rather, our more modest goal is to simply check whether any statistically significant relationship exists, consistent with a falsification test, rather than undertaking serious attempts to establish causality.
 
35
One ex ante concern that may arise is that central bank transparency may be correlated systematically with a country’s development level and, hence, transparency merely is capturing a development effect, realized through the choice of monetary regime (especially inflation targeting). We believe that that concern is not an issue in practice, for three reasons. First, in our sample, inflation targeters are more or less distributed between advanced and emerging economies. Second, the correlation between transparency and per capita income is fairly low (\(\rho =0.41\)) and specifications controlling for both do not exhibit symptoms of multicollinearity. Third, per capita income is included already as a control in the set of parsimonious independent variables in specification C6; thus, transparency is likely capturing a distinct political economy effect.
 
36
This notion of operational independence is, of course, related to the idea of checks-and-balances—captured by the political constraints index—in the robustness section. But we believe that it is possible to pin down more directly the means by which the exercise (or not) of those checks matters operationally for monetary policy (rather than just the overall policy environment), which motivates our use of turnover.
 
37
One concern with relying on the turnover rate is the possibility of mismeasurement; for example, if turnover is extremely low (suggesting high de facto independence) but that is because of absolute loyalty by the central banker to the political elite, a low turnover rate is paradoxically the result of less independence. That said, the foregoing concern does not appear to be borne out empirically. The correlation between turnover frequency and the Polity2 index is very small (\(\rho =0.05\)); among developing economies that also are autocracies, this correlation is even smaller and even statistically insignificant (\(\rho =0.03\), p = 0.36). Nevertheless, we also consider two alternative central banker tenure measures: the total tenure length of any given central bank governor and the amount of time remaining before a scheduled turnover. The former may better capture the circumstance wherein a loyal central banker is in place because tenure continues to increase as a result of a reappointment regardless of the legally mandated tenure in office. The latter indicator resets with each appointment and could be a better indicator of de facto independence if longer tenures per se reflect greater independence (rather than the inverse relationship wherein more turnovers reflects less independence, regardless of tenure length). Across those variations, the coefficients on the level and interaction terms remain insignificant. The results are available on request.
 
38
Notably, the quality of the instruments is weak in columns C5 and C6, as indicated by poor underidentification and weak instrument tests.
 
39
To verify that finding, we reran C5 and C6 by development status. The coefficients in both subsamples are qualitatively the same as above (if anything, the magnitudes of the coefficients when using the full sample tend to be more modest); the results are available on request.
 
40
In the Online Appendix, we run a decomposition exercise for the independence variable and find that while the positive coefficient remains for other subindexes, the policy independence subindex enters with a negative sign. This suggests that inflation tends to be lower when central bankers are able define their goals and formulate monetary policies to achieve them.
 
41
In the Online Appendix, we use our dataset to offer some speculative results in that regard.
 
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Metadata
Title
The limits of central bank independence for inflation performance
Author
Jamus Jerome Lim
Publication date
02-01-2020
Publisher
Springer US
Published in
Public Choice / Issue 3-4/2021
Print ISSN: 0048-5829
Electronic ISSN: 1573-7101
DOI
https://doi.org/10.1007/s11127-019-00771-8

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