Monetary Policy Rules in Practice: Evidence from Turkey and Israel

Transcription

1 Moneary Policy Rules in Pracice: Evidence from Turkey and Israel M. Ege Yazgan* İsanbul Bilgi Universiy, İsanbul, Turkey Hakan Yılmazkuday** Vanderbil Universiy, Nashville, TN, USA (0 December 004) Absrac We esimae forward looking moneary policy rules for Israel and Turkey. When variable inflaion arges are aken ino consideraion, as opposed o he fixed arges used in prior research ha use daa from developed counries, forward looking Taylor rules seem o provide reasonable descripion of Cenral Bank behavior in boh counries. In general, i can be said ha moneary policy appears o be quie srong in hese counries, and especially so in Turkey, when compared wih developed counries. JELClassificaion: E5, E58 Key Words: Taylor Rule, Moneary Policy Rule, Moneary Policy, Turkey, Israel *Kurulusderesi Caddesi, No: Dolapdere, Isanbul TURKEY Tel: Fax: eyazgan@bilgi.edu.r ** hakan.yilmazkuday@vanderbil.edu 1

2 1. Inroducion Moneary policy rules have been analyzed in considerable volume by researchers boh from descripive and prescripive perspecives 1. As saed by Svensson (003), from a descripive perspecive, research has examined o wha exen moneary policy rules proposed in he lieraure, explain acual Cenral Bank behavior. For developed counries, his quesion has been exensively analyzed (see Clarida, Gali and Gerler, 1998, 1999, 000; Taylor, 1993, 1999; Nelson, 000; Judd and Rudebusch 1998, among ohers). However, for developing counries he empirical lieraure has remained sparse. One reason for his could be he fac ha flexible exchange rae sysem is a relaively new developmen for mos developing counries afer long periods of some form of a fixed exchange rae sysem in place. As saed by Taylor (000), once an emerging marke economy abandons he fixed exchange rae regime (a peg, a currency board, or a common currency dollarizaion) he only sound moneary policy alernaive is he one based on he riniy (Taylor, 1993) of a flexible exchange rae, an inflaion arge, and a moneary policy rule. As a resul many developing counries have recenly sared implemening inflaion argeing regimes wih an accompanying moneary policy rule. In his paper, we conribue o his empirical lieraure by considering wo developing economies, namely, Turkey and Israel. Our aim is o see how reasonable i is o approximae he behavior of he cenral banks of hese counries by our proposed Taylor rule. Turkey adoped a flexible exchange rae regime from February 001 and explicily declared is inflaion arges saring from January 00. Israel, on he oher hand adoped inflaion argeing in 199 along wih a flexible exchange rae regime. The paper esimaes forward looking moneary policy rules for boh Turkey and Israel similar o hose in Clarida, Gali and Gerler, (1998, 1999, 000) (CGG, hereafer) and Taylor (1993). The specificaions in CGG (1998, 1999, 000) are forward looking rules esimaed by he Generalized Mehod of Momens (GMM) wih implici fixed arges over long esimaion periods. Alhough his specificaion is reasonable for developing counries for which he inflaion arges say almos fixed over years, for he wo counries under consideraion in his sudy, i canno be accepable because hey had variable inflaion arges over he period under 1 This erminology belongs o Svensson (003).

3 consideraion. Therefore we will modify he CGG formulaion o ake ino accoun he variable arges. To be able o do his we will use expeced inflaion daa insead of acual inflaion as CGG. This expeced inflaion daa is based on surveys conduced by he Cenral Banks of he wo counries. Like CGG, we use GMM for he esimaion of our moneary policy rules. However, recenly GMM esimaors have been severely criicized on he ground ha inference based on hese esimaors is inconclusive. The relaed economeric lieraure indicaes ha here has been considerable evidence ha asympoic normaliy provides a poor approximaion o he sampling disribuions of GMM esimaors. Paricularly, he -sage leas square (SLS) esimaor becomes heavily biased (in he same direcion as he ordinary leas squares esimaor), and he disribuion of he SLS esimaor is quie far from he normal disribuion (e.g. bimodal). Sock and Wrigh (000) aribue his problem o weak idenificaion or weak insrumens, ha is, insrumens ha are only weakly correlaed wih he included endogenous variables. Sock, Wrigh and Yogo (00) and Dufour (003) provide a comprehensive survey on weak idenificaion in GMM esimaion. We also address his issue in our esimaors by using recenly developed saisics ha are immune o weak idenificaion. The res of he paper is organized as follows: Secion illusraes our modified specificaion developed o ake ino accoun he variables arges in he usual backward looking policy rules. Secion 3 describes he daa and provides he esimaion resuls. Secion 4 concludes.. The model Consider he following moneary policy rule of he form (Taylor, 1993, CGG, 1998, 1999, 000) ( / ) ( / ) * * * i = r+ π+ k + β E π+ k Ω π+ k + γe x Ω (1) See Clarida, Gali, and Gerler (1999) Berumen and Tasci (004) esimaes moneary policy rules for Turkey over he period from They adop he framework of CGG. This approach can be criicized on wo grounds. As saed above, arges are reaed as fixed and inflaion argeing was adoped as a policy a he beginning of

4 where i * denoes he arge rae for nominal ineres rae in period. E is he expecaion operaor, and Ω is he informaion se a he ime he ineres rae is se. π + kdenoes he percen change in he price level beween periods and +k (expressed in annual raes). he (variable) arge for inflaion for period +k formed a period. x is a measure of he average oupu gap, wih he oupu gap being defined as he percen deviaion beween acual oupu and he corresponding arge (poenial). r is he long-run equilibrium real rae 3. As in Clarida, Gali, and Gerler (000), we assume ha he real rae is saionary and is deermined by non-moneary facors in he long-run. * π + k is The policy rule given by (1) has been proven o be useful in boh heoreical and empirical grounds. I has been used exensively in empirical research on developed counries as noed above. Approximae forms of his rule are opimal for a cenral bank ha has a quadraic loss funcion in deviaions of inflaion and oupu from heir respecive arges, given a generic macroeconomic model wih price ineria (see, Clarida, Gali, and Gerler 1999, Svensson, 003). Moreover, as is shown by Clarida, Gali and Gerler (001, 00) his policy rule has he same form as in (1) for a small open economy, wih possibly differen coefficiens han hose of a closed economy 4. However, all of hese and previously menioned sudies consider a fixed inflaion arge over he period of esimaion, which makes sense for developed counries, bu has lile relevance for he experiences of Turkey and Israel wih inflaion argeing. Therefore, we will modify Equaion (1) o esimae i wih variable inflaion arges. Rearranging Equaion (1) yields * ( / ) r = r+ β E π+ k Ω π + k + γx + ε () * ε = γ Ω + µ, r = i π + k, and i is he acual nominal ineres rae. where x E( x / ) The erm µ capures he difference beween he desired and he acual nominal ineres rae, 3 I should be noed ha r is an approximae real rae since forecas horizon for inflaion will generally differ from he mauriy of he shor-erm nominal rae used as a moneary policy insrumen. As noed by Clarida, Gali and Gerler (000), in pracice he presence of high correlaion beween shor-erm raes and a mauriies associaed wih he arge horizon (1 year) prevens his from being a problem. 4 Clarida, Gali, and Gerler (001, 00) indicaes ha openness only affecs he magniude of he coefficiens in he policy rule. 4

5 * i.e. µ = i i 5. This difference, as pu by Clarida, Gali and Gerler (000), may resul from hree facs. Firsly, he specificaion in Equaion () assumes an adjusmen of he acual overnigh raes o is arge level, and hus ignores, if any, he Cenral Banks (CB) s endency o smooh changes in ineres raes (we will address his issue below). Secondly, i reas all changes in ineres raes over ime as reflecing he CB s sysemaic response o economic condiions. Specifically, i does no allow for any randomness in policy acions, oher han associaed wih misforecass of he economy. Third, i assumes ha he CB has perfec conrol over he ineres raes, i.e. i succeeds in keeping hem a he desired level (e.g., hrough open marke operaions). Finally le z be a vecor of variables wihin he cenral bank s informaion se a ime i chooses he ineres rae (i.e. z Ω ) ha are orhogonal o ε. Possible elemens of z include any lagged variables ha help o forecas oupu gap, as well as any conemporaneous variables ha are uncorrelaed wih he curren ineres rae shock µ. Then E ε / = 0, Equaion () implies he following orhogonaliy condiion since [ z ] ( ) * E r r β E π k / π + + k γx / + Ω + z = 0 (3) By using his orhogonaliy condiion we use GMM o esimae he parameer vecor [ r, β, γ ] as in Clarida, Gali and Gerler (1998, 000). CBs may have a endency o smooh changes in ineres rae. Ineres rae smoohing can be inroduced ino he model via he following parial adjusmen mechanism (see Clarida, Gali, and Gerler, 1998, 000) ( 1 ρ) * ρ 1 r = r + r + v (4) where he parameer ρ [ 0,1] capures he degree of ineres rae smoohing. Equaion (4) posulaes ha each period he CB adjuss he funds rae o eliminae a fracion ( 1 ρ ) of he 5 We assume ha µ is idenically and independenly disribued. 5

6 gap beween is curren arge level and pas value. ν is a independenly and idenically disribued error erm. Subsiuing () ino (4) and rearranging yields * { } ( 1 ) ( / ) r = ρ r+ β E π+ k Ω π+ k + γx + ρr 1 + ξ (5) Where ( 1 ) x E( x / ) ξ = ρ γ Ω + ν. Apparenly Equaion (5) implies an orhogonaliy condiion similar o (5). 3. Daa and Time Series Properies For Turkey, annual year-end CPI inflaion arges are incorporaed in he disinflaion program implemened wih he suppor of he IMF: he arges for 00, 003, and 004 respecively 35 %, 0 %, and 1 % were indicaed in he relevan review of he Sandby Arrangemen during he las quarer of 001 and also explicily declared in Cenral Bank of Turkey (CBT) websie. As our inflaion measure looks a he nex 1 monhs, we need o find a soluion o he disconinuiy of hese year-end arges, which we do by a linear ransformaion, whereby he inflaion arge in he formula falls each monh o reach he nex year s inflaion arge a he firs monh of he nex year. For expeced inflaion we use CBT s expecaions surveys ha presen expeced figures for he inflaion for nex 1 monhs. We use he simple average of he wo surveys published by CBT. The equivalen in Turkey of he money marke shor erm ineres rae of he Taylor rule is he overnigh ineres rae on borrowing published daily by he CBT: consensus again exiss in he markes ha his is he relevan indicaor of moneary policy. This daa is available over he period 001M8-004M4. For Israel we use expeced and arge inflaion raes provided by he Cenral Bank of Israel (CBI) over he period 1999M1-00M1. Treasury bill rae is used as a proxy for he money marke shor erm ineres rae. This laer variable is obained from Inernaional Financial Saisics (IFS) CDROM published Inernaional Moneary Fund. Measuring he oupu gap is a rouine calculaion in developed economies wih low oupu volailiy and smooh (relaively slow) changes in he srucure of oupu and he composiion of domesic demand. Especially for Turkey his variable raises several serious issues, furher amplified by crisis condiions wihin he period of observaion. The ideal measure should be 6

7 based, in our opinion, on he comparison of he real domesic demand excluding he change in invenories wih is long-erm rend value, for which daa is only published on a quarerly basis wih a lag of hree monhs. Therefore, we use he seasonally adjused indusrial producion series (IPS) and use he definiion of Khalaf and Kichian (004) for he measure of oupu gap. Tha is, raher han derending he log of IPS using he full sample, T, we proceed ieraively: o obain he value of he gap a ime, we derend IPS wih he daa ending in T. We hen exend he sample by one more observaion and re-esimae he rend. This is used o derend IPS and yields a value for he gap a ime +1. This process is repeaed unil he end of he sample. In his fashion, our gap measures a ime T do no use informaion beyond ha period and can herefore be used as valid insrumens. To derend IPS. Hodrick-Presco (HP) filer. IPS series for boh counries are obained from IFS CDROM. Since he economeric esimaion procedure ha we use here (GMM) requires ha all he variables (including insrumens) used in he esimaion should be saionary, all of he variables are esed by using he Augmened Dickey-Fuller (ADF) ess and we find ha he null of uni roo is rejeced in all variables, a leas a he 10 percen significance level, when ess are applied a differen lags 6 4. Esimaion In his secion we will perform a linear GMM esimaion o obain he esimaors of he parameers of Equaion (). The GMM esimaor ha we use is Limied Informaion Maximum Likelihood (LIML) 7. The resuls are illusraed in Table 1 for Turkey and Israel. The insrumens we use for GMM esimaion consis of one lag of oupu gap and and one lag of oupu growh, boh for Turkey and Israel 8. Table 1 is abou here The firs hree columns of Table 1 and repor he esimaes of r, β, and γ. The esimae of he coefficien on difference beween expeced and argeed inflaion is around 0.95 for 6 These resuls are available upon reques. 7 The usual Two Sage Leas Square esimaors yield exacly he same resuls. For LIML esimaors we used he GAUSS code originally used by Sock and Wrigh (000). 8 By choosing hese insrumens, we implicily assume ha hese wo variables are srong insrumens for predicing oupu gap. 7

8 Turkey and 0.73 for Israel. Tha is, if expeced inflaion were 1 percenage poin above he arge, he Cenral Banks (CB) would se he real ineres rae 95 and 73 basis poin above is equilibrium value. This coefficien also appears o be highly significan for boh counries when we use asympoic normaliy as an approximaion o he sampling isribuion of GMM esimaors. The response of he CBs o he deviaions of expeced oupu gap from is arge (assumed o be zero) is around 0.54 for Turkey and is 0. for Israel. In oher words holding he difference beween expeced and argeed inflaion consan one percen increase in oupu gap induces he CBs o increase he real raes by 54 and percen in Turkey and Israel respecively. These coefficiens are also saisically significan a 5 percen significance (hough no a 1 percen) level. Anoher difference in he wo counries lies in esimaed coefficien of he equilibrium real ineres rae. This is esimaed as 13 and 6 percen for Turkey and Israel respecively. They are also highly significan using normal asympoics. These esimaes indicae he fac ha he CB of Turkey implemened much sronger moneary policy in he period of observaion han ha of he CB of Israel. The Hansen s J-saisic repored in Table 1, does no rejec he null hypohesis ha he overidenifying resricions are saisfied a convenional significance levels for Turkey. However, for Israel J-saisic indicaes ha overidenfying resricions are no saisfied. Despie heir significance (or insignificance), as we menion in he inroducion, one should wary abou GMM-based resuls ha are obained under he asympoic normaliy of he sampling disribuions ha obained under convenional asympoics. However, under weakidenificaion asympoics, he sampling disribuions are quie far from being normally disribued. In his paper we address he problem of weak idenificaion by using Anderson and Roubin (1949), es (AR es) in is general form developed by Dufour and Jasiak (001) (see also Dufour 003). This es is robus in he case of nonlinear models (see Dufour, 003; Sock, Wrigh and Yogo, 00), and perhaps, more imporanly, are even robus o excluded insrumens (see Dufour, 003). Since i is rarely possible o use all possible insrumens, his laer propery is quie imporan from applied poin of view. AR es saisic is used o es he null hypohesis ha (for Turkey s esimaed parameers), H : r = 1.675; β = ; γ = , i.e. given he insrumens ha we used, wheher he 0 esimaed parameers of Equaion (3) are compaible wih he daa or no. Since he es is 8

9 fully robus o weak insrumens (see Sock, Wrigh, and Yogo, 00, pp.5), a non-rejecion of his null hypohesis means ha our esimaes are also daa-admissible even under he case of weak insrumens. The AR-saisics, under he above null hypohesis has an exac Fisher disribuion wih k and T-k-x degrees of freedom (where k is he number of insrumens, x is he number of exogenous variables, and T is he number of observaions), given ha he error erms are i.i.d. normal and he insrumens are sricly exogenous. k AR saisics are asympoically disribued chisquare wih k degrees of freedom, even wihou i.i.d. normal errors under sandard regulariy condiions (see Dufour and Jasiak, 001, pp. 89, and Dufour 003, pp.0). As can be followed from Table 1 given he high p-value of he AR-es, our parameer esimaes canno be rejeced. In oher words our GMM esimaes of moneary policy rules given Equaion (3), canno be refued by neiher he Turkish nor Israelian daa. Table summarizes he esimaion resuls of Equaion (5), i.e. when ineres rae smoohing is inroduced ino he equaion ino Equaion (3). The insrumen se for his esimaion conains firs and second lags of oupu gap and growh 9. Table is abou here The esimaion resuls indicae ha smoohing parameers for boh counries are highly significan, similar o he oher parameers, and heir magniudes equal o 0.51 (Turkey) and 0.78 (Israel). These esimaes imply ha alhough Israel s CB allows less variabiliy in is ineres rae, boh counries CBs pu significan effors for smoohing ineres rae. The inclusion of he ineres rae smoohing parameer also leads o some significan changes in he esimaed parameers of he previous equaion. When his parameer is included, he only negligible change occurs in r (he esimaed value of his coefficien moves from 1.7 o 11.1 for Turkey, from 5.9 o 6.1 for Israel. However, here are considerable changes in he esimaed of he remaining parameers especially in β, and o a lesser exen inγ (for Turkey only). When ineres rae smoohing is presen in he model, he esimae of he coefficien on difference beween expeced and argeed inflaion is considerably higher han in he absence 9 In his case he esimaion mehod is nonlinear LIML wih coninuous updaing algorihm. We again used he GAUSS code of Sock and Wrigh (000). The asympoic sandard errors of he esimaors are calculaed by using dela mehod. 9

10 of smoohing. This coefficien is now esimaed as 1.19 for Turkey and 1.35 for Israel. Tha is, if expeced inflaion were 1 percenage poin above he arge, he CBs would se he real ineres rae 1.19 and 1.35 basis poin above is equilibrium value. The response of he CBs o he deviaions of expeced oupu gap from is arge is now around 0.73 for Turkey, i.e., wih smoohing Turkish CB appears o be more reacive o he deviaions in he oupu gap. There is no much significan change in he esimaed value of his variable compared o he oher case. Similar o he above case J-Saisics indicaes ha he validiy of our insrumens is no rejeced by he daa and AR saisics confirms he validiy of he model even in he case of weak insrumens. 5. Conclusion When variable arges are aken ino accoun, forward looking Taylor rules seem o provide reasonable descripion of CBs behavior, in boh Turkey and Israel, even wih only wo response variables such as deviaion from arges and oupu gap. I should be poined ou ha we also include some oher variables in Taylor rules, such as money growh, real exchange rae, deviaion of he real exchange from an equilibrium level, nominal exchange rae growh. None of hese variables urn ou o be significan in hese counries. In general, i can be said ha moneary policy appears o be quie srong, especially in Turkey, in hese counries when compared wih he developed counries policy funcion as oulined in he above menioned papers. 10

11 Tables Table 1 Resuls of Equaion (3) for Turkey and Israel Counry r β γ AR-sa J-sa Sample size (n) F(;9) χ () χ (1) Turkey (1.1839) (0.1518) (0.833) [0.087] [0.4373] [0.469] [0.3446] 33 F(;44) χ () χ (1) Israel (0.195) (0.1303) 0.17 (0.0494) [0.434] [0.471] [0.001] 48 Noes: Sandard errors are in paranheses and p-values are in brackes. 11

12 Table Resuls of Equaion (5) for Turkey and Israel Counry r β γ ρ AR-sa J-sa Sample size (n) F(4;5) χ (4) χ (3) Turkey (1.965) (0.759) (0.3439) [0.0164] (0.1458) [0.0003] [0.4578] [0.440] [0.04] 3 F(4;40) χ (4) χ (3) Israel (0.717) (0.346) 0.37 (0.0650) [0.0003] (0.0390) [0.394] [0.3801] [0.031] 47 Noes: Sandard errors are in paranheses and p-values are in brackes. 1

13 References: Anderson TW. and Rubin H Esimaors of he Parameers of a Single Equaion in a Complee Se of Sochasic Equaions. The Annals of Mahemaical Saisics 1: Berumen H. and Tasci H Moneary Policy Rules in Pracice: Evidence from Turkey. Inernaional Journal of Finance and Economics 9: Clarida R., Gali J. and Gerler M Moneary Policy Rules in Pracice: Some Inernaional Evidence. European Economic Review 4: Clarida R., Gali J. and Gerler M The Science of Moneary Policy: A New Keynesian Perspecive. Journal of Economic Lieraure XXXVII: Clarida R., Gali J. and Gerler M. 000 Moneary Policy Rules and Macroeconomic Sabiliy: Evidence and Some Theory. The Quarerly Journal of Economics 115(1): Clarida R., Gali J. and Gerler M Opimal Moneary Policy in Open versus Close Economies: An Inegraed Approach. American Economic Review, 91(): 48-5 Clarida R., Gali J. and Gerler M. 00. A simple Framework for Inernaional Moneary Policy Analysis. Journal of Moneary Economics. 49: Dufour JM Idenificaion, Weak Insrumens, and Saisical Inference in Economerics. Canadian Journal of Economics 36(4): Dufour JM. and Jasiak J Finie Sample Limied Informaion Inference Mehods for Srucural Equaions and Models wih Generaed Regressors. Inernaional Economic Review. 4(3): Judd JP. and Rudebusch GD Taylor s Rule and he Fed: Federal Reserve Bank of San Francisco Economic Review. 3:

14 Hansen L Large Sample Properies of Generalized Mehod of Momens Esimaors. Economerica. 50: Khalaf L. and Kichian M Esimaing New Keynesian Phillips Curves Using Exac Mehods. Bank of Canada Working Paper. 11. Nelson E UK Moneary Policy : A Guide Using Taylor Rules. Bank of England Working Papers. 10 Sock JH., Wrigh, JH. and Yogo M. 00. A Survey of Weak Insrumens and Weak Idenificaion in Generalized Mehod of Momens. Journal of Business and Economic Saisics. 0(4): Sock JH. and Wrigh JH GMM wih Weak Idenificaion. Economerica. 68(5): Svensson LE Wha is Wrong wih Taylor Rules? Using Judgmen in Moneary Policy hrough Targeing Rules. Journal of Economic Lieraure. XLI: Taylor JB Using Moneary Policy Rules in Emerging Marke Economies. (available a hp:// Taylor JB Discreion Versus Policy Rules in Pracice. Carnegie-Rocheser Conference Series on Public Policy. 39: