# Euro-Dollar Real Exchange Rate Dynamics in an Estimated Two-Country Model: What Is Important and What Is Not - WP/06/177

←

**Page content transcription**

If your browser does not render page correctly, please read the page content below

WP/06/177 Euro-Dollar Real Exchange Rate Dynamics in an Estimated Two-Country Model: What Is Important and What Is Not Pau Rabanal and Vicente Tuesta

© 2006 International Monetary Fund WP/06/177 IMF Working Paper Middle East and Central Asia Department Euro-Dollar Real Exchange Rate Dynamics in an Estimated Two-Country Model: What Is Important and What Is Not Prepared by Pau Rabanal and Vicente Tuesta1 Authorized for distribution by Miguel Savastano July 2006 Abstract This Working Paper should not be reported as representing the views of the IMF. The views expressed in this Working Paper are those of the author(s) and do not necessarily represent those of the IMF or IMF policy. Working Papers describe research in progress by the author(s) and are published to elicit comments and to further debate. We use a Bayesian approach to estimate a standard two-country New Open Economy Macroeconomics model using data for the United States and the euro area, and we perform model comparisons to study the importance of departing from the law of one price and complete markets assumptions. Our results can be summarized as follows. First, we find that the baseline model does a good job in explaining real exchange rate volatility but at the cost of overestimating volatility in output and consumption. Second, the introduction of incomplete markets allows the model to better match the volatilities of all real variables. Third, introducing sticky prices in Local Currency Pricing improves the fit of the baseline model but does not improve the fit as much as introducing incomplete markets. Finally, we show that monetary shocks have played a minor role in explaining the behavior of the real exchange rate, while both demand and technology shocks have been important. JEL Classification Numbers: F41, C11 Keywords: Real exchange rates, Bayesian estimation, model comparison Author(s) E-Mail Address: pau.rabanal@gmail.com, vtuesta@bcrp.gob.pe 1 Pau Rabanal was an economist in the Middle East and Central Asia Department of the IMF when this paper was written. Vicente Tuesta is an economist in the Research Department of the Central Reserve Bank of Peru. The views expressed in this Working Paper are those of the authors and do not necessarily represent those of the IMF or IMF policy, or the Central Reserve Bank of Peru. The authors thank Philippe Bachetta, Roberto Chang, Jordi Galí, Gian Maria Milesi-Ferretti, Juan Rubio, Miguel Savastano, and seminar participants at various institutions for their helpful suggestions.

2 Contents Page I. Introduction...................................................................................................................4 II. The Model.....................................................................................................................6 A. Households............................................................................................................6 B. Asset Market Structure, Budget Constraint, and the Real Exchange Rate ...........8 C. Intermediate Goods Producers and Price-Setting ...............................................10 D. Closing the Model...............................................................................................11 E. Symmetric Equilibrium.......................................................................................12 F. Dynamics ............................................................................................................12 III. Extensions to the Baseline Model...............................................................................14 A. Incomplete Markets with Stationary Net Foreign Assets ...................................14 B. LCP by Intermediate Goods Producers...............................................................15 C. Incomplete Markets and Sticky Prices in LCP ...................................................17 IV. Estimation and Model Comparison ............................................................................18 A. Data .....................................................................................................................18 B. Bayesian Estimation of the Model’s Parameters ................................................20 V. Results.........................................................................................................................23 A. Posterior Distributions for the Parameters..........................................................23 B. Model Comparison..............................................................................................27 C. Second Moments.................................................................................................28 D. Shocks and Real Exchange Rate Dynamics........................................................30 E. Forecasts .............................................................................................................33 VI. Concluding Remarks...................................................................................................35 Appendix: The Metropolis-Hastings Algorithm...................................................................36 References .............................................................................................................................37 Tables 1. Variables in the Home and Foreign Countries..............................................................7 2. Definitions and Functional Forms ................................................................................9 3. Linearized Equations ..................................................................................................13 4. Properties of the Data for the United States and the Euro Area .................................19 5. Prior Distributions of the Model’s Parameters ...........................................................21 6. Posterior Distributions ................................................................................................24 7. Selected Second Moments in the Data and in the Models..........................................28 8. Contributions of the Shocks to Selected Second Moments in the Preferred Model...31 9. Mean Squared Errors of One Period Ahead Forecasts ...............................................34

3 Figures 1. Impulse Responses, U.S. Technology Shock .............................................................32 2. Impulse Responses, U.S. Demand Shock ...................................................................33

4 I. INTRODUCTION Most puzzles in international macroeconomics are related to real exchange rate dynamics. Fluctuations in real exchange rates can be very large and persistent, when compared to other real variables. In addition, there is clear evidence of lack of consumption risk-sharing across countries, which is at odds with the assumption of complete markets. In order to replicate these features of the data, the New Open Economy Macroeconomics (NOEM) literature has incorporated either nominal rigidities, alternative structures of assets markets, or both. The real exchange rate (qt) between two currencies is defined as the ratio of the two countries’ price levels expressed in a common currency.2 When all the components of the price level, namely domestically produced and imported goods, are sticky, it can be possible to explain some empirical features, like the high correlation between nominal and real exchange rates, and real exchange rate volatility. In the literature, pricing of imported goods are assumed to be governed either by Producer Currency Pricing (PCP), where the law of one price holds and there is perfect pass-through; or Local Currency Pricing (LCP), where the pass-through is zero in the short run. Under complete markets, the real exchange rate should be equal to the ratio of the marginal utility of consumption across countries, because it reflects the relative price of foreign goods in terms of domestic goods. For example, assuming separable preferences and log utility, the following relationship should hold as an equilibrium condition: q =ct-ct*, where ct and ct* are the levels of domestic and foreign consumption. This relationship, which implies a correlation of one between the real exchange rate and the ratio of consumption levels in two countries, does not hold for many bilateral relationships in general. For the bilateral euro-U.S. dollar exchange rate in particular, the correlation between these variables (HP-filtered) is -0.17. Hence, models that incorporate complete markets are bound to perform poorly, even when they allow for other nominal or real rigidities. One possibility to get around this problem is to assume that agents do not have access to complete markets to ensure their wealth against idiosyncratic and country-specific shocks. Another possibility is to introduce preference shocks that affect the marginal utility of consumption, as in Stockman and Tesar (1995). A recent paper by Chari, Kehoe, and McGrattan (2002—hereafter CKM) attempts to explain the volatility and persistence of the real exchange rate by constructing a model with sticky prices and LCP. Their main finding is that monetary shocks and complete markets, along with a high degree of risk aversion and price stickiness of one year, are enough to account for real exchange rate volatility and, to a lesser extent, for its persistence. However, their model found it difficult to account for the observed negative correlation between real exchange rates and relative consumption across countries, a fact that they labeled the consumption-real 2 In log-linear terms, the real exchange rate is defined as qt=st + pt*-pt, where st is the (logarithm of the) nominal exchange in units of domestic currency per unit of foreign currency, pt* is the (log of the) foreign price level, and pt is the (log of the) domestic price level.

5 exchange rate anomaly. In addition, CKM showed that some conventional ways of modeling of asset market incompleteness and habit persistence do not eliminate the anomaly.3 We use a Bayesian approach to estimate and compare two-country NOEM models using different assumptions of imports goods pricing and asset markets structures, thereby testing some of the key implications of CKM. Unlike them, we find that monetary policy shocks have a minor role in explaining real exchange rate volatility, and that both demand and technology shocks have had some importance. Using the Bayes factor to compare between competing alternatives, we find that what is crucial to explaining real exchange rate dynamics and the exchange rate-consumption anomaly is the introduction of incomplete markets with stationary net foreign asset positions. Somewhat surprisingly, we find that in a complete markets set up, the introduction of LCP improves the fit of the model. However, when incomplete markets are allowed for, LCP actually lowers the overall fit, overestimating real exchange rate volatility and implying a lower correlation between the real exchange rate and the ratio of relative consumptions than in the data. We contribute to the existing literature on estimation of NOEM models. First, we focus on the relationship between relative consumption and the real exchange rate by introducing data on consumption for the United States and the euro area. Second, although our model is quite rich in shocks (we need nine shocks because we try to explain nine variables), we have left aside uncovered interest-rate parity (UIP)-type shocks, which tend to explain a large fraction of real exchange rate variability. We do so because under complete markets these shocks, at least conceptually, should not be included and also because we want to study more carefully the role of “traditional” shocks (technology, demand, monetary, and so on) in explaining real exchange rate fluctuations.4 Third, we believe this is the first paper to evaluate the merits of the incomplete markets assumption with stationary net foreign assets in a two-country NOEM model. Last, but not least, we perform an in-sample forecast exercise and find that the preferred model does a good job in forecasting compared to the other NOEM models, but still lags behind the performance of a vector autorregression (VAR) model. The literature on estimating NOEM models in the spirit of CKM and Galí and Monacelli (2005) has grown rapidly, with the adoption of the Bayesian methodology to an open economy setting.5 For example, Lubik and Schorfheide (2006) estimate small open 3 Alternative ways to explain this anomaly typically include models with traded and nontraded goods. Selaive and Tuesta (2003a) and Benigno and Thoenissen (2005) have shown that this anomaly can be successfully addressed by models with incomplete markets and nontraded goods, with the traditional Balassa-Samuelson effect and sector-specific productivity shocks. Similarly, Ghironi and Melitz (2005) rely on aggregate productivity shocks and also find that the Balassa-Samuelson effect help to explain the consumption real exchange rate anomaly. Finally, Corsetti, Dedola, and Leduc (2004) have shown that distribution services can help to account for the real exchange rate-consumption correlation by lowering import demand elasticity. 4 Our benchmark model, unlike the International Real Business Cycle (IRBC) literature, always includes nominal rigidities, because we want to evaluate the relative importance of monetary shocks in explaining real exchange rate fluctuations. 5 Examples of closed economy applications of this methodology are Rabanal and Rubio-Ramírez (2005), and Galí and Rabanal (2004) for the United States, and Smets and Wouters (2003) for the euro area.

6 economy models with data for Australia, Canada, New Zealand, and the United Kingdom, to examine whether the monetary policy rules of those countries have targeted the nominal exchange rate. Justiniano and Preston (2004) estimate and compare small open economy models analyzing the consequences of introducing imperfect pass-through. Adolfson et al. (2005) estimate a medium-scale (15 variable) small open economy model for the euro area, while Lubik and Schorfheide (2005) and Batini et al. (2005) estimate a small-scale two-country model using U.S. and euro-area data. The rest of the paper is organized as follows. In the next section, we outline the baseline model and describe the LCP and the incomplete market extensions. In Section III, we explain the data and, in Section IV, the econometric strategy. The estimation results are reported in Section V. First, we present the parameter estimates of the baseline model. Then, we analyze the parameter estimates of all the extensions along with the second moments implied by each model. We select our preferred model based on the comparison of Bayes factors, and analyze its dynamics by studying the impulse response functions. Finally, we evaluate the importance of shocks through variance decompositions. We also compare the forecasting performance of all the Dynamic Stochastic General Equilibrium (DSGE) models with respect to VAR models. Section VI concludes. II. THE MODEL In this section, we present the stochastic two-country NOEM model that we will use to analyze real exchange rate dynamics.6 We first outline a baseline model with complete markets and where the law of one price holds, in the spirit of Clarida, Galí, and Gertler (2002); Benigno and Benigno (2003); and Galí and Monacelli (2005). We add features that are known to improve the model’s empirical properties, namely: home bias and habit formation in consumption, and staggered price setting with backward looking indexation. In the next section, we introduce two extensions that we are interested in comparing: incomplete markets and sticky prices of imported goods in local currency. The model assumes that there are two countries, home and foreign, of equal size. Each country produces a continuum of intermediate goods, indexed by h ∈ [0,1] in the home country and f ∈ [0,1] in the foreign country. Preferences over these goods are of the Dixit-Stiglitz type, implying that producers operate under monopolistic competition, and all goods are internationally tradable. Table 1 lists all the variables of the model. The model contains nine shocks: a world technology shock that has a unit root, and country-specific stationary technology, monetary, demand and preference shocks. All stationary shocks are AR(1), except for the monetary shocks that are iid. A. Households In each country there is a continuum of infinitely lived households in the unit interval, who obtain utility from consuming the final good and disutility from supplying hours of labor. It 6 This type of model has been the workhorse of the NOEM literature after Obstfeld and Rogoff (1995).

7 Table 1. Variables in the Home and Foreign Countries Home Foreign Quantity Price Quantity Price Consumption goods Aggregate Ct Pt Ct* Pt* Imports C F ,t PF ,t C H* ,t PH* ,t Domestically produced C H ,t PH ,t C F* ,t PF*,t Intermediate goods Imports ct ( f ) pt ( f ) ct* (h) p t* (h) Domestically produced ct ( h) p t ( h) ct* ( f ) pt* ( f ) Production Aggregate (GDP) Y H ,t PH ,t YF*,t PF*,t Intermediate goods y t ( h) p t ( h) y t* ( f ) pt* ( f ) Home Foreign Labor markets Hours worked Nt N t* Real wage ωt ω t* Firms’ labor demand N t ( h) N t* ( f ) Terms of trade Tt Tt* Interest rates Rt Rt* Bonds Bt Bt* Real exchange rate Qt Nominal exchange rate St Shocks World technology At Country technology Xt X t* Preference Gt Gt* Monetary zt z t* Demand ηt η t* is assumed that consumers have access to complete markets at the country level and at the world level, which implies that consumers’ wealth is insured against country-specific and world shocks, and hence all consumers face the same consumption-savings decision.7 7 Baxter and Crucini (1993) have used the same assumption in an IRBC model in order to explain the saving-investment correlation.

8 In the home country, households’ lifetime utility function is: ∞ N t1+γ E 0 ∑ β Gt [log(C t − bC t −1 ) − t ]. (1) t =0 1+ γ E0 denotes the rational expectations operator using information up to time t=0. β ∈ [0,1] is the discount factor. The utility function displays external habit formation. b ∈ [0,1] denotes the importance of the habit stock, which is last period’s aggregate consumption. γ>0 is inverse elasticity of labor supply with respect to the real wage. Table 2 contains additional variable definitions and functional forms. Ct denotes the consumption of the final good, which is a CES aggregate of consumption bundles of home and foreign goods. The parameter 1 − δ is the fraction of home-produced goods in the consumer basket, and denotes the degree of home bias in consumption. Its analogous in the foreign country is 1 − δ * . The elasticity of substitution between domestically produced and imported goods in both countries is θ, while the elasticity of substitution between types of intermediate goods is ε>1. In our baseline case, we assume that the law of one price holds for each intermediate good. This implies that PH ,t = St PH* ,t , and PF ,t = St PF*,t . Note, however, that purchasing power parity (a constant real exchange rate) does not necessarily hold because of the presence of home bias in preferences. The home-bias assumption allows to generate real exchange rate dynamics in a model, like this one, with only tradable goods. From previous definitions, we can express the real exchange rate as a function of the terms of trade: 1 S t Pt ⎡ δ * + (1 − δ * )Tt1−θ ⎤ 1−θ * Qt = =⎢ 1−θ ⎥ (2) Pt ⎣ (1 − δ ) + δTt ⎦ B. Asset Market Structure, Budget Constraint, and the Real Exchange Rate We model complete markets by assuming that households have access to a complete set of state contingent nominal claims which are traded domestically and internationally. We represent the asset structure by assuming a complete set of contingent one-period nominal bonds denominated in home currency. 8 Hence, households in the home country maximize their utility (1) subject to the following budget constraint: Et {ξ t ,t +1 Bt +1 } − Bt 1 Ct = ωt N t + + ∫ Π t (h)dh, (3) Pt 0 8 Given these assumptions, it is not necessary to characterize the current account dynamics in order to determine the equilibrium allocations, and the currency denomination of the bonds is irrelevant.

9 Table 2: Definitions and Functional Forms θ θ −1 θ −1 θ −1 ⎡ 1 1 ⎤ C t ≡ ⎢(1 − δ ) θ (C H ,t ) θ + δ θ (C F ,t ) θ ⎥ ⎣⎢ ⎦⎥ Consumption θ θ −1 θ −1 θ −1 ⎡ 1 1 ⎤ C t* ≡ ⎢(δ * ) θ (C H* ,t )θ + (1 − δ ) (C ) θ * θ * F ,t ⎥ ⎢⎣ ⎥⎦ ε ε ε −1 ε −1 ⎧ 1 ⎫ ε −1 ⎧ ⎫ ε −1 ∫ [c ] ⎪ ⎪ ⎪ ⎪ ≡ ⎨ [ct (h)] dh ⎬ , C H* ,t ≡ ⎨ 1 ∫ ε * ε C H ,t t ( h) dh ⎬ ⎪⎩ 0 ⎪⎭ ⎪⎩ 0 ⎪⎭ Consumption components ε ε ε −1 ε −1 ⎧ 1 ⎫ ε −1 ⎧ ⎫ ε −1 ∫ [c ] ⎪ ⎪ ⎪ ⎪ ≡ ⎨ [ct ( f )] df ⎬ , C F* ,t ≡ ⎨ 1 ∫ ε * ε C F ,t t ( f) df ⎬ ⎪⎩ 0 ⎪⎭ ⎪⎩ 0 ⎪⎭ 1 Consumer price indices [ Pt ≡ (1 − δ )( PH ,t ) 1−θ + δ ( PF ,t ) 1−θ ]1−θ , 1 ≡ [δ ] 1−θ Pt* * ( PH* ,t )1−θ + (1 − δ * )( PF*,t )1−θ . 1 1 ∫ [p ] 1−ε 1−ε ⎧ ⎫ 1−ε ⎧ ⎫ 1−ε ∫ [ p ( h)] 1 1 PH ,t ≡⎨ t dh ⎬ , PH* ,t ≡⎨ * t ( h) dh ⎬ ⎩ 0 ⎭ ⎩ 0 ⎭ Price sub indices 1 1 ∫ [p ] 1−ε 1−ε ⎧ 1 ⎫ 1−ε ⎧ ⎫ 1−ε ≡ ⎨ [ pt ( f )] df ⎬ , PF*,t ≡ ⎨ 1 ∫ * PF ,t t ( f) df ⎬ ⎩0 ⎭ ⎩ 0 ⎭ Terms of trade Tt = PF ,t / PH ,t , Tt* = PH* ,t / PF*,t Net exports NX t = PH ,t YH ,t / Pt − C t S t Pt* Real exchange rate Qt = Pt Production functions y t (h) = At X t N t (h) , y t* ( f ) = At X t* N t* ( f ) World technology shocks log( At ) = Γ + log( At −1 ) + ε ta * Country technology shocks log( X t ) = ρ x log( X t −1 ) + ε tx , log( X t* ) = ρ x* log( X t*−1 ) + ε tx * Preference shocks log(Gt ) = ρ g log(Gt −1 ) + ε tg , log(Gt* ) = ρ g* log(Gt*−1 ) + ε tg

10 where Bt +1 denotes nominal state-contingent payoffs of the portfolio purchased in domestic currency at t , and ξ t ,t +1 is the stochastic discount factor.9 The real wage is deflated by the country’s consumer price index (CPI). The last term of the right hand side of equation (3) denotes the profits from the monopolistically competitive intermediate goods firms, which are ultimately owned by households in each country. Combining optimality conditions of consumption in both countries under complete markets, we arrive at the following expression for the real exchange rate, that equals the ratio of marginal utilities of the two countries: (C t − bC t −1 ) Gt* Qt = ν * , (4) (C t − b *C t*−1 ) Gt where ν is a constant that depends on initial conditions (see CKM, and Galí and Monacelli, 2005). The risk-sharing condition (4) differs from the one in CKM because of the presence of both preference shocks and habit persistence. C. Intermediate Goods Producers and Price-Setting In each country, there is a continuum of intermediate goods producers, each producing a type of good that is an imperfect substitute of the others. As shown in Table 2, the production function is linear in the labor input, and has two technology shocks. The first one is a world technology shock, that affects the two countries the same way: it has a unit root, as in Galí and Rabanal (2004) and Ireland (2004), and it implies that real variables in both countries grow at a rate Γ. In addition, there is a country-specific technology shock that evolves as an AR(1) process. Firms face a modified Calvo (1983)-type restriction when setting their prices. When they receive the Calvo-type signal, which arrives with probability 1 − α in the home country, firms reoptimize their price. When they do not receive that signal, a fraction τ of intermediate goods producers index their price to last period’s inflation rate, and a fraction 1 − τ indexes their price to the steady-state inflation rate. This assumption is needed to incorporate trend inflation, as in Yun (1996). The equivalent parameters in the foreign country are 1 − α * and τ * . Cost minimization by firms implies that the real marginal cost of production is ω t /( At X t ) . Since the real marginal cost depends only on aggregate variables, it is the same for all firms 9 ξ t ,t +1 is a price of one unit of nominal consumption at time t+1, expressed in units of nominal consumption at t, contingent on the state at t+1 being st+1, given any state st in t. The complete market assumption implies that there exists a unique discount factor with the property that the price in period t of the portfolio with random value Bt +1 is E t {ξ t ,t +1 Bt +1 } .

11 in each country. The overall demand for an intermediate good produced in h comes from optimal choices by consumers at home and abroad: −ε −θ ⎛ p ( h) ⎞ ⎛ PH ,t ⎞ Dt (h) = ct (h) + c (h) = ⎜⎜ t * t P ⎟ ⎟ ⎜⎜ ⎟⎟ [(1 − δ )C t + δ *C t*Qtθ . ] ⎝ H , t ⎠ ⎝ Pt ⎠ Hence, whenever intermediate-goods producers are allowed to reset their price, they maximize the following profit function, which discounts future profits by the probability of not being able to reset prices optimally every period: ∞ ⎧ ⎡ p t ,t + k ( h ) ωt +k ⎤ ⎫ Max Et ∑ α k ξ t ,t + k ⎨⎢ − MCt + k ⎥ Dt ,t + k (h)⎬. (5) ⎩⎣ Pt + k At + k X t + k ⎦ pt ( h ) k =0 ⎭ where pt ,t + k (h) is the price prevailing at t+k assuming that the firm last reoptimized at time t, and whose evolution will depend on whether the firm indexes its price to last period’s inflation rate or to the steady-state rate of inflation, Dt ,t + k (h) the demand associated to that price, and ξ t ,t + k is the k periods ahead stochastic discount factor. The evolution of the aggregate consumption bundle price produced in the home country is: τ 1−ε ⎡ ⎛ PH ,t −1 ⎞ 1−τ ⎤ PH ,t = (1 − α )( PˆH ,t ) + α ⎢ PH ,t −1 ⎜⎜ 1−ε 1− ε ⎟ ΠH ⎥ ⎟ . (6) ⎢ ⎝ P ⎠ ⎥ ⎣ H , t − 2 ⎦ ˆ where PH ,t is the optimal price set by firms in a symmetric equilibrium. D. Closing the Model In order to close the model, we impose market clearing conditions for all home and foreign intermediate goods. For each individual good, market clearing requires y t (h) = ct (h) + ct* (h) for all h ∈ [0,1]. Defining aggregate real GDP as YH ,t = ⎡ ∫ pt (h) y t (h)dh⎤ / PH ,t , the following 1 ⎢⎣ 0 ⎥⎦ market clearing condition holds at the home-country level: −θ ⎛ PH ,t ⎞ YH ,t [ = (1 − δ )C t + δ C Qt .⎜⎜ * * t θ ] ⎟⎟ + ηt (7) ⎝ Pt ⎠ The analogous expressions for the foreign country are, y t* ( f ) = ct ( f ) + ct* ( f ) , for all f ∈ [0,1] and for aggregate foreign real GDP: −θ ⎛ PH ,t ⎞ Y * F ,t [ = δC t + (1 − δ )C Qt .⎜⎜ * * t θ ] ⎟⎟ + η t* (8) ⎝ Pt ⎠

12 We also introduce an exogenous demand shock for each country (η t ,η t* ) that can be interpreted as government purchases, and/or trade with third countries. The model is closed by assuming that each country follows a monetary policy rule of the Taylor-type. We present the rules for each country in Section II.F. below. E. Symmetric Equilibrium Since we have assumed a world-wide technology shock that grows at a rate Г, output, consumption, real wages, and the level of exogenous demand in the two economies grow at that same rate. In order to render these variables stationary, we divide them by the level of world technology At . Hours, inflation, interest rates, the real exchange rate, and the terms of trade are stationary. F. Dynamics We obtain the model’s dynamics by taking a linear approximation to the steady state values at zero inflation. We impose a symmetric home bias, such that δ = δ * . We denote by lower case variables percent deviations from steady state values. Moreover, variables with a tilde have been normalized by the level of technology to render them stationary. For instance, ~ ~ ~ ~ c~t = (C t − C ) / C , where C t = C t / At . The relationship between the transformed variables in the model (normalized by the level of technology) and the first-differenced variables is as follows: c~t = c~t −1 + ∆ct − ε ta , ~ yt = ~ y t −1 + ∆y t − ε ta , c~t* = c~t*−1 + ∆ct* − ε ta , and ~ y t* = ~ y t*−1 + ∆y t* − ε ta . where ∆ denotes the first difference operator. These relationships are used in the estimation strategy, since we include first-differenced real variables in the set of observable variables. In this subsection, we focus the discussion on the equations that influence the behavior of the real exchange rate, and that will be affected by the introduction of imperfect pass-through and incomplete markets. Table 3 presents the rest of the model’s equations, which are fairly standard given our assumptions. The only exception are the Taylor rules, which modify the original formulation by reacting to output growth instead of the output gap, incorporating interest rate smoothing, and an iid monetary shock. The risk sharing condition delivers the following relationship between consumption in the two countries, the preference shocks, and the real exchange rate: ⎡ (1 + Γ)c~t − bc~t −1 ⎤ ⎡ (1 + Γ)c~t − b c~t −1 ⎤ ⎛ 1+ Γ 1+ Γ ⎞ a * * * qt = ⎢ ⎥ − ⎢ ( ⎥ − gt − gt + ⎜ * ) − ε . * ⎟ t (9) ⎣ 1+ Γ − b 1+ Γ − b + Γ − + Γ − * ⎦ ⎣ ⎦ ⎝ 1 b 1 b ⎠ As in CKM, the real exchange rate depends on the ratio of marginal utilities of consumption, which in our case include the habit stock in each country, and the preference shocks. Note

13 Table 3: Linearized Equations Euler b∆ct = −(1 + Γ − b)(rt − Et ∆pt +1 ) + (1 + Γ) Et ∆ct +1 + (1 + Γ − b)(1 − ρ g ) g t , equations b ∆ct* = −(1 + Γ − b * )(rt* − Et ∆pt*+1 ) + (1 + Γ) Et ∆ct*+1 + (1 + Γ − b * )(1 − ρ g* ) g t* * Labor ⎡ (1 + Γ)c~t − bc~t −1 + bε ta ⎤ ~ * ⎡ (1 + Γ)c~t* − b * c~t*−1 + b *ε ta ⎤ ω~t = γnt + ⎢ ⎥ , ω t = γnt + ⎢ * ⎥ supply ⎣ 1+ Γ − b ⎦ ⎣ 1 + Γ − b* ⎦ Goods ~ ⎡ 2δ (1 − δ ) ⎤ ⎡ 2δ (1 − δ ) ⎤ market y H ,t = θ ⎢ ⎥ qt + (1 − δ )c~t + δc~t* + η t , ~ y F* ,t = −θ ⎢ ⎥ qt + δc~t + (1 − δ )c~t* + η t* . clearing ⎣ 1 − 2δ ⎦ ⎣ 1 − 2δ ⎦ Production y H ,t = xt + nt , ~ ~ y F* ,t = xt* + nt* functions rt = ρ r rt −1 + (1 − ρ r )γ p ∆p H ,t + (1 − ρ r )γ y ∆y H ,t + z t Taylor rules ( ) ( ) rt* = ρ r* rt*−1 + 1 − ρ r* γ *p ∆p F* ,t + 1 − ρ r* γ *y ∆y F* ,t + z t* Terms of ∆t t = ∆st + ∆p F* ,t − ∆p H ,t trade that the innovation to world growth enters as long as the effect on the ratio of marginal utilities is different in the two countries, due to differences in the habit formation parameters. Inflation dynamics for domestically produced goods in each country are given by: ∆p H ,t = γ b ∆p Ht −1 + γ f Et ∆p Ht +1 + κ [ω~t − xt + δt t )], (10) ∆p F* ,t = γ b* ∆p F* ,t −1 + γ *f Et ∆p F* ,t +1 + κ * [ω~t* − xt* − δ *t t ]. (11) where for the home country, the backward and forward looking components are γ b ≡ τ /(1 + βτ ) , γ f ≡ β /(1 + βτ ) , and the slope is given by κ ≡ (1 − αβ )(1 − α ) /[(1 + βτ )α ] . Similar expressions with asterisks deliver the coefficients γ b* , γ *f , and κ * . Domestic inflation is determined by unit labor costs (the real wage), productivity shocks, and the terms of trade. This last variable appears because wages are deflated by the CPI: an increase in imports prices will cause real wages to drop, and households will demand higher wages. As a result, domestic inflation will also increase. When the law of one price holds, the real exchange rate and the terms of trade are linked as follows: qt = (1 − 2δ )t t . The symmetric home bias assumption implies a positive comovement between the real exchange rate and the terms of trade which is consistent with the data. Thus, in this model, the real exchange rate inherits the properties of the terms of trade. With no home bias (δ=1/2), the real exchange rate is constant and purchasing power

14 parity holds. The degree of home bias is crucial to account for the volatility of the real exchange rate: the larger the degree of home bias (smaller δ), the larger the volatility of the real exchange rate.10 Finally, the CPI inflation rates are a combination of domestic inflation and imported goods. Since prices are set in the producer currency, and the law of one price holds, the nominal exchange rate has a direct inflationary impact on CPI inflation: ∆pt = (1 − δ )∆p H ,t + δ∆p F* ,t + δ∆st (12) and ∆pt* = δ * ∆p H* ,t − δ * ∆st + (1 − δ * )∆p F* ,t . (13) III. EXTENSIONS TO THE BASELINE MODEL A. Incomplete Markets with Stationary Net Foreign Assets In this section, we introduce the incomplete markets assumption. We assume that home-country households are able to trade in two nominal riskless bonds denominated in domestic and foreign currency, respectively. These bonds are issued by home-country residents in the domestic and foreign currency to finance their consumption. Home-country households face a cost of undertaking positions in the foreign bonds market.11 For simplicity, we further assume that foreign residents can only allocate their wealth in bonds denominated in foreign currency. In each country, firms are still assumed to be completely owned by domestic residents, and profits are distributed equally across households. The real budget constraint of home-country households is now given by: Bt S t Bt* Bt −1 S t Bt*−1 1 Ct + + = ω t N t + + + ∫ Π t (h)dh, (3’) Pt R t ⎛S B ⎞ * Pt Pt 0 Pt Rt*φ ⎜⎜ t t ⎟⎟ ⎝ Pt ⎠ where the φ (.) function depends on the real holdings of the foreign assets in the entire economy, and therefore is taken as given by individual households.12 10 In a model with nontradable goods, this proportionality is broken down so that the real exchange rate will depend upon the relative price of tradable to nontradable goods across countries. 11 This cost is needed to obtain stationarity in the net foreign asset position. See Schmitt-Grohe and Uribe (2001) and Kollman (2002) for applications in small open economies, and Benigno (2001) and Selaive and Tuesta (2003a) for applications in two-country models. Heathcote and Perri (2002) have used the same transaction cost in a two-country IRBC model. 12 In order to achieve stationarity, φ (.) has to be differentiable and decreasing in a neighborhood of zero. We further assume that φ (.) equals zero when Bt* =0.

15 We further assume that the initial level of wealth is the same across households in each country. This assumption combined with the fact that households within a country equally share the profits of intermediate goods producers, implies that within a country all households face the same budget constraint. In their consumption decisions, they will choose the same path of consumption. Dynamics Under incomplete markets, the net foreign asset (NFA) position for the home country consists of the holding of foreign bonds (since domestic bonds are in net supply in the symmetric equilibrium). By definition, the NFA position of the foreign country equals the stock of bonds outstanding with the home country. The risk sharing condition holds in expected first difference terms and depends on the NFA position and preference shocks: ⎡ (1 + Γ) Et ∆ct +1 − b∆ct ⎤ ⎡ (1 + Γ) Et ∆ct +1 − b ∆ct ⎤ * * * Et q t +1 − q t = ⎢ − ⎥ + (1 − ρ g ) g t − (1 − ρ g ) g t + χbt * * * ⎥ ⎢ ⎣ 1+ Γ − b ⎦ ⎣ 1+ Γ − b * ⎦ (9’) ⎛S B ⎞* where χ = −φ ' (0 )YH and bt* = ⎜⎜ t t ⎟⎟ , which substitutes equation (9) in section II.F. ⎝ YH Pt ⎠ The net foreign asset position becomes a state variable—its evolution depends on the stock of previous debt and on the trade deficit (or surplus):13 ⎡ 2θ (1 − δ ) − 1⎤ βbt* = bt*−1 + δ ⎢ 1 − 2 δ ⎥ ( ) qt − δ c~t − c~t* . (14) ⎣ ⎦ Note that the effect of the real exchange rate on the NFA critically depends on the size of the elasticity of substitution: with a low elasticity, a real depreciation will imply that volumes increase less than prices decline, and hence the value of net exports declines after a real devaluation. B. LCP by Intermediate Goods Producers We assume price stickiness in each country’s import prices in terms of local currency. Each firm chooses a price for the domestic market and a price for the foreign market under the same conditions of the modified Calvo lottery with indexation described above. This assumption generates deviations from the law of one price at the border, and nominal 13 This expression comes from defining the law of motion of the NFA position as β bt* = bt*−1 + nxt , where nxt = NX t / YH , and we make use of the expression of net exports in Table 2, the goods markets clearing condition in Table 3, and the consumer’s optimizing conditions.

16 exchange rate movements generate ex-post deviations from the law of one price.14 Importantly, under the assumption of LCP, even without home bias, it is possible to generate real exchange rate fluctuations. The overall demand (from domestic and foreign households) for an intermediate good produced in h, is given by: −ε −θ −ε −θ ⎛ p ( h) ⎞ ⎛ PH ,t ⎞ ⎛ p * ( h) ⎞ ⎛ PH* ,t ⎞ ct (h) = (1 − δ )⎜⎜ t ⎟ ⎜⎜ ⎟⎟ C t and ct* (h) = δ ⎜ t * ⎟ ⎜ ⎟ C t* . P ⎟ ⎜ P ⎟ ⎜ P* ⎟ ⎝ H , t ⎠ ⎝ Pt ⎠ ⎝ H ,t ⎠ ⎝ t ⎠ Hence, whenever domestic intermediate-goods producers are allowed to reset their prices in the home and the foreign country, they maximize the following profit function: ∞ Max Et ∑ α ξ t ,t + k ⎨ k [ ] ⎧⎪ [ pt ,t + k (h) − ω t /( At X t )]ct ,t + k (h) + pt*,t + k (h) S t + k − ω t /( At X t ) ct*,t + k (h) ⎫⎪ ⎬. pt ( h ), pt* ( h ) k =0 ⎪⎩ Pt + k ⎪⎭ where pt ,t + k (h) and pt*,t + k (h) are prices of the home good set at home and abroad prevailing at t+k assuming that the firm last reoptimized at time t, and whose evolution will depend on whether the firm indexes to last period’s inflation rate (a fraction τ of firms) or to the steady-state rate of inflation (a fraction 1- τ of firms) when it is not allowed to reoptimize. ct ,t + k (h) and ct*,t + k (h) are the associated demands for good h in each country. To obtain the log-linear dynamics, we first need to redefine the terms of trade: t t ≡ p F ,t − p H ,t , and t t* ≡ p H* ,t − p F* ,t . These ratios represent the relative price of imported goods in terms of the domestically produced goods expressed in local currency, for each country.15 Dynamics The following new equations arise with respect to the baseline (PCP) case. The inflation equations for home-produced goods are: ∆p H ,t = γ b ∆p Ht −1 + γ f Et ∆p Ht +1 + κ [ω~t − xt + δt t )], (10’) 14 Monacelli (2005) assumes that retail importers are subject to sticky prices, rather than the exporting firms in the country of origin. In his model, the law of one price holds at the border, but the pass-through is slow. 15 Note that if the law of one price holds, t t = −t t* , but now it is no longer the case.

17 ∆p H* ,t = γ b ∆p H* ,t −1 + γ f Et ∆p H* ,t +1 + κ [ω~t − xt − (1 − δ )t t* − qt )], (10b’) ∆p F* ,t = γ b* ∆p F* ,t −1 + γ *f Et ∆p F* ,t +1 + κ * [ω~t* − xt* + δt t* ], (11’) ∆p F ,t = γ b* ∆p F ,t −1 + γ *f Et ∆p F ,t +1 + κ * [ω~t* − xt* − (1 − δ )t t + q t ], (11b’) Similarly to the baseline case, real wages are deflated by the CPI which causes the terms of trade for each country, as well as the real exchange rate, to matter in the determination of unit labor costs and of domestic inflation. The CPI inflation rates under LCP do not include the nominal exchange rate as a direct determinant of imported goods inflation, because the pass-through is low and import prices are sticky in domestic currency: ∆pt = (1 − δ )∆p H ,t + δ∆p F ,t (12’) and ∆pt* = δ∆p H* ,t + (1 − δ )∆p F* ,t (13’) which substitute equations (10)–(13) of the baseline model. In addition, the market-clearing conditions in Table 3 become: y% H ,t = (1 − δ ) θδ ( tt − tt* ) + (1 − δ )c%t + δ c%t* + ηt , and y% F* ,t = − (1 − δ ) θδ ( tt − tt* ) + δ c%t + (1 − δ ) c%t* + ηt* . C. Incomplete Markets and Sticky Prices in LCP Under incomplete markets and LCP, the equations of the model are given by those in Section II.F, Table 3, and modified by those in Section III.B. The additional change is that while the behavior of the real exchange rate is the same as under incomplete markets (Equation 9’ in Section III.A), the NFA position dynamics are given by: βbt* = bt*−1 + δ (1 − δ )(θ − 1)(t t − t t* ) + δqt − δ (c~t − c~t* ). (14’) which substitutes (14) in Section III.A.

18 IV. ESTIMATION AND MODEL COMPARISON A. Data We use data for the United States and euro area to estimate the model. For the United States, we use the following series (mnemonics as they appear in the Haver USECON database): quarterly real GDP (GDPH), the GDP deflator (DGDP), real consumption (CH), and the 3-month T-bill interest rate (FTB3). Since we want to express real variables in per capita terms, we divide real GDP and consumption by total population of 16 years and over (LN16). Data for the euro area as a whole comes from the Fagan, Henry, and Maestre (2001) dataset. (This dataset is a synthetic dataset constructed by the Econometric Modeling Unit at the European Central Bank, and should not be viewed as an “official” series.) We extract from that database real consumption (PCR), real GDP (YER), the GDP deflator (YED), and short-term interest rates. The euro zone population series is taken from Eurostat. Since it consists of annual data, we transform it to quarterly frequency by using linear interpolation. The convention we adopt is that the home country is the euro area, and the foreign country is the United States. The real exchange rate consists of the nominal exchange rate in euros per U.S. dollar, converted to the real exchange rate index by multiplying it by the U.S. CPI and dividing it by the euro area CPI. The “synthetic” euro/U.S. dollar exchange rate prior to the launch of the euro in 1999 also comes from Eurostat; the U.S. CPI comes from the Haver USECON database (PCU) and the euro area CPI comes from the Fagan, Henry, and Maestre database (HICP). Our sample period goes from 1973:1 to 2003:4, at quarterly frequency, which is when the euro area dataset ends. To compute per capita output and consumption growth rates and inflation, we take natural logs and first differences of per capita output and consumption, and the GDP deflator, respectively. We divide the short-term interest rate by four to obtain its quarterly equivalent. We also take natural logs and first differences of the euro/dollar real exchange rate. Table 4 presents some relevant statistics. Interestingly, the raw data show that per capita output growth rates in the United States and the euro area are not that different (0.48 percent versus 0.47 percent), while per capita consumption and output in the euro area grow at the same rate (0.47 percent). Consumption growth in the United States displays a higher sample mean growth rate (0.53 percent) than in the euro area, which is not surprising given recent trends. Interestingly, growth rates in the euro area are less volatile than in the United States. The real exchange rate displays a small appreciating trend mean during the sample period and is much more volatile than any other series. Real exchange rare volatility also stands out in the HP-filtered series: the bilateral real exchange rate has a standard deviation of 7.83 percent, while output and consumption in the United States have a standard deviation of 1.58 percent and 1.28 percent, respectively. Output and consumption in the euro area are less volatile, with a standard deviation of about 1 percent. Interest rates and inflation rates display high persistence, and so do all real variables when they are HP-filtered. Interestingly, only consumption in the euro area displays

19 Table 4: Properties of the Data for the United States and the Euro Area (1973:1–2003:4) Raw Data, Quarterly Growth Rates Consumption Consumption Output Real Exch. Euro Output Euro USA USA Rate Mean 0.47 0.47 0.53 0.48 -0.14 Std. dev. 0.57 0.58 0.67 0.85 4.59 Raw Data, Quarterly Rates Interest Rate Inflation Interest Rate Inflation Euro Euro USA USA Mean 2.08 1.44 1.59 1.00 Std. dev. 0.83 0.93 0.73 0.67 First autocorr. 0.96 0.89 0.94 0.90 HP-Filtered Data Consumption Consumption Output Real Exch. Euro Output Euro USA USA Rate Std. dev. 0.91 1.01 1.28 1.58 7.83 Corr. with RER -0.26 -0.06 -0.02 -0.08 1.00 First autocorr. 0.84 0.86 0.87 0.87 0.83 Relative Consumption Output Relative Outputs, Euro, USA Euro, USA Cons., RER RER Other Correlations 0.33 0.47 -0.17 0.04 Source: Haver Analytics, Eurostat, and Fagan, Henry, and Maestre (2001). Note: Relative variables are the ratio between the euro area variable and its U.S. counterpart. a nonzero correlation of -0.26 with the real exchange rate. The correlation of output in the euro area, and output and consumption in the United States with the real exchange rate is essentially zero. Finally, it is worth noting that the correlation between consumptions is smaller than between outputs (0.33 versus 0.47), although the size of the two correlations are smaller than those obtained using shorter sample periods, as in Backus, Kehoe, and Kydland (1992).16 The correlation of relative output with the real exchange rate is fairly small, while the correlation between the real exchange rate and relative consumptions across countries is negative (-0.17), which is at odds with efficient risk-sharing.17 16 Heathcote and Perri (2004) show that in recent years the U.S. economy has become less correlated with the rest of the world. 17 All the facts related to the U.S. economy are very similar to the ones presented in CKM.

20 B. Bayesian Estimation of the Model’s Parameters According to Bayes’ rule, the posterior distribution of the parameters of any given model is proportional to the product of the prior distribution of the parameters and the likelihood function of the data. An appealing feature of the Bayesian approach is that additional information about the model’s parameters (i.e., micro-data evidence, features of the first moments of the data) can be introduced via the prior distribution. To implement the Bayesian estimation method, we need to numerically evaluate the prior and the likelihood function. The likelihood function is evaluated using the state-space representation of the law of motion of the model, and the Kalman filter. We then use the Metropolis-Hastings algorithm to obtain random draws from the posterior distribution, from which we obtain the relevant moments of the posterior distribution of the parameters. Let ψ denote the vector of parameters that describe preferences, technology, the monetary policy rules, and the shocks in the two countries of the model. The vector of observable variables consists of zt = {∆yt , ∆ct , rt , ∆pH ,t , ∆yt* , ∆ct* , rt* , ∆pF* ,t , ∆qt }' . The assumption of a world technology shock with a unit root makes the real variables stationary in the model in first differences. Hence, we use consumption and output growth per country, which are stationary in the data and in the model. We first-difference the real exchange rate, while inflation and the nominal interest rate in each country enter in levels.18 We express all variables as deviations from their sample mean. We denote by L({zt }Tt=1 | ψ) the likelihood function of {zt }Tt=1 . Priors Table 5 shows the prior distributions for the model’s parameters, which we denote by Π(ψ). For the estimation, we decide to fix only two parameters. The first one is the steady-state growth rate of the economy. Based on the evidence presented in Section IV.A, we set Γ=0.5 percent, which implies that the world growth rate of per capita variables is about 2 percent per year. In order to match a real interest rate in the steady state of about 4 percent per year, we set the discount factor to β=0.995. For reasonable parameterizations of these two variables, the parameter estimates do not change significantly. For the other parameters, gamma distributions are used as priors when nonnegativity constraints are necessary, and uniform priors when we are mainly interested in estimating fractions or probabilities. Normal distributions are used when more informative priors seem to be necessary. Unlike other two-country model studies (e.g., Lubik and Schorfheide, 2005; and CKM), we do not impose the constraint that parameter values be the same in the two countries. However, we do use the same prior distributions for parameters across countries. We use normal distributions for the coefficients of habit formation and inverse elasticity of labor supply with respect to the real wage, centered at values commonly used in the literature (0.7 18 In this way, we avoid having to choose a particular detrending method (linear, quadratic, or HP-filter).

21 Table 5: Prior Distributions of the Model’s Parameters Parameter Distribution Mean Std. Dev. Habit formation b , b* Normal 0.70 0.05 Labor supply γ ,γ * Normal 1.00 0.25 Average duration between optimal price changes (1 − α ) −1 , (1 − α * ) −1 Gamma 3.00 1.42 Indexation τ ,τ * Uniform(0,1) 0.50 0.29 Fraction of imported goods δ Normal 0.20 0.03 Elasticity of substitution between home and foreign goods θ Normal 1.50 0.25 Elasticity of the real exchange rate to χ Gamma 0.02 0.014 the NFA position Taylor rule: inflation γ p , γ *p Normal 1.50 0.25 Taylor rule: output growth γ y , γ *y Normal 1.00 0.20 Taylor rule: smoothing ρr , ρ * Uniform(0,1) 0.50 0.29 r AR coefficients of shocks ρ x , ρ x* , ρ g , Uniform(0,0.96) 0.48 0.28 ρ g* , ρη , ρη* Std. dev. technology shocks σ x , σ x* , σ a Gamma 0.007 0.003 Std. dev. preference shocks σ g , σ g* Gamma 0.010 0.005 Std. dev. monetary shocks σ z , σ z* Gamma 0.004 0.002 Std. dev. demand shocks σ η , σ η* Gamma 0.010 0.005 and 1, respectively). We truncate the habit formation parameter to be between 0 and 1. We assume that the average duration of price contracts has a prior mean of 3 in the two countries, following empirical evidence reported in Taylor (1999). In this case, a gamma distribution is used.19 The prior on the fraction of price setters that follow a backward looking indexation rule is less informative and takes the form of a uniform distribution between zero and one. The priors over the parameters that incorporate the open economy features of the model are as follows: (i) the parameter δ, which captures the implied home bias, has a prior distribution with mean 0.2 and standard deviation 0.03, implying a smaller home-bias than suggested by Heathcote and Perri (2002) and CKM; (ii) the elasticity of substitution between home and foreign goods (θ) is a source of controversy, so we center it at a value of 1.5 as suggested by CKM, but with a large enough standard deviation to accommodate other feasible parameters, 19 To keep the probability of the Calvo lottery between 0 and 1, the prior distribution is specified as average duration between optimal price changes minus one: D=1/(1-α)-1. The shape of the prior is not much different than assuming a beta distribution for α.

22 even those below one;20 and (iii) the parameter χ, that measures the elasticity of the risk premium with respect to the net foreign asset position, is assumed to have a gamma distribution with mean of 0.02 and a standard deviation of 0.014, following the evidence in Selaive and Tuesta (2003a and 2003b). For the coefficients of the interest rate rule, we center the coefficients to the values suggested by Rabanal (2004), who estimates rules with output growth for the United States. Hence, γ p has a prior mean of 1.5, and γ y has a prior mean of 1. The same values are used for the monetary policy rule in the euro area, and we use uniform priors for the autoregressive processes between zero and one. We also truncate the prior distributions of the Taylor rule coefficients such that the models deliver a unique and stable solution. We use uniform priors on the autoregressive coefficients of the six AR(1) shocks. We truncate the upper bound of the distribution to 0.96, because we want to examine how far the models can go in replicating persistence. We choose gamma distributions for the priors on the standard deviations of the shocks, to avoid negative values. The prior means are chosen to match previous studies. For instance, the prior mean for the standard deviation of all technology shocks is set to 0.007, close to the values suggested by Backus, Kehoe, and Kydland (1992), while the prior mean of the standard deviation of the monetary shocks comes from estimating the monetary policy rules using OLS. Drawing from the Posterior and Model Comparison We implement the Metropolis-Hastings algorithm to draw from the posterior. The results are based on 250,000 draws from the posterior distribution.21 The definition of the marginal likelihood for each model is as follows: L({z t }Tt=1 ) = ∫ L({z t }Tt=1 | ψ )Π (ψ )dψ (17) ψ ∈Ψ The marginal likelihood averages all possible likelihoods across the parameter space, using the prior as a weight. Multiple integration is required to compute the marginal likelihood, making the exact calculation impossible. We approximate it by using the modified harmonic mean estimator.22 20 Trade studies typically find values for the elasticity of import demand with respect to price (relative to the overall domestic consumption basket) in the neighborhood of five to six, see Trefler and Lai (1999). Most of the NOEM models consider values of 1 for this elasticity, which implies Cobb-Douglas-type preferences in aggregate consumption. 21 See the appendix for some details on the estimation. Lubik and Schorfheide (2005, 2006) also provide useful details on the estimation procedure. 22 See Fernández-Villaverde and Rubio-Ramírez (2004) for computational details.

23 Then, for two different models (A and B), the posterior odds ratio is P( A | {zt }Tt=1 ) Pr( A) L({zt }Tt=1 | model = A) = . P( B | {zt }Tt=1 ) Pr( B ) L({zt }Tt=1 | model = B) If there are m ∈ M competing models, and one does not have strong views on which model is the best one (i.e., Pr(A)=Pr(B)=1/M) the posterior odds ratio equals the Bayes factor (i.e., the ratio of marginal likelihoods). V. RESULTS We report the results of our estimation in five stages. First, we present the posterior estimates obtained for a closed economy vis-à-vis the four specifications considered for the open economy model. Second, we perform a model comparison by evaluating the marginal likelihood for each model. Third, we compute the standard deviations and correlations of each model at the mode posterior values. Fourth, we discuss the dynamics of our preferred model by analyzing the importance of the structural shocks for real exchange rate fluctuations. And finally, we look at the one-step ahead in-sample forecast performance of all models, and compare their performance to VARs. A. Posterior Distributions for the Parameters Table 6 presents relevant moments of the posterior parameters of all the models. In order to have a benchmark for the open economy estimates, we first provide the results from estimating each country as a closed economy. Column I reports the mean and standard deviation of the posterior distributions of the parameters for the euro area and the United States, when both are estimated as a closed economies. We assume that within each country agents only consume home produced goods (δ=θ=0), and are not allowed to trade bonds internationally. In addition, the real exchange rate is dropped from the set of observed variables. Overall, the estimates in Column I are in line with those obtained in the literature, (e.g., Galí and Rabanal (2005), Rabanal and Rubio-Ramírez (2005) and Lubik and Schorfheide (2005)), hence we do not discuss them any further. Our benchmark open economy model assumes complete markets and PCP. The parameter estimates are displayed in Column II. The results differ in important ways with respect to the closed economy case. First, the proportion of firms that index their prices to the lagged inflation rate increases to almost one in the euro area, while inflation remains almost purely forward looking in the United States. The average duration of price contracts decreases for the euro area to 4.77 quarters and increases significantly for the United States to 14.74 quarters, which is a fairly large number.23 The habit persistence parameters increase 23 This result is a consequence of having assumed a production function that is linear in the labor input. If we had assumed, as Galí, Gertler, and López-Salido (2001) that the production function is concave in labor, we would obtain smaller average price durations. Introducing firm-specific capital or real demand rigidities, as in Altig, et. al. (2005) or Eichenbaum and Fischer (2004), would also have lowered average price duration.

You can also read