Download Does globalization alter the monetary transmission mechanism?

Does globalization alter the monetary transmission mechanism?∗
Tobias Cwik
Goethe University Frankfurt
Gernot J. Müller
Goethe University Frankfurt
Maik Wolters
Goethe University Frankfurt
April 2, 2008
Abstract
In this paper we quantify effects of globalization - measured by increased goods markets integration - on the monetary transmission mechanism within a two country general equilibrium
model. The model is specified to give a quantitatively realistic account of the actual transmission
mechanism. Most notably, it features strategic complementarities in price setting such that relative import prices directly impact on domestic inflation. In a first step, we pin down parameter
values of the model by matching empirical impulse responses to a monetary policy shock. We
compute the latter running vector autoregressions on U.S. time series and an aggregate of industrialized countries. We find that the model is able to mimic the empirical impulse responses quite
closely for plausible parameter estimates. In a second step, we use the model to perform counterfactual experiments in order to assess possible changes in the transmission mechanism induced
by globalization.
∗
Keywords:
International monetary policy transmission, Globalization,
Goods market integration, Openness, Pricing to market
JEL-Codes:
F41, F42, E52
First draft.
Comments welcome.
We thank Zeno Enders and Keith Kuester for helpful discussions. The usual disclaimer applies. Please address correspondence to [email protected],
[email protected] or [email protected]
1 Introduction
The consequences of globalization for monetary policy are widely discussed both in academia and
among policy makers. In this discussion, globalization is understood as a trend towards increased international integration of goods and financial markets. Several observers have pointed to direct effects
of this trend on domestic prices in major industrialized countries, notably disinflationary tendencies
due to increased global capacity and competition, see, e.g., Mishkin (2007). In addition, globalization may alter the dynamics of the inflation process which are frequently analyzed on the basis of a
Phillips curve relationship, see e.g., Sbordone (2007) and Guerrieri, Gust, and López-Salido (2008).
To the extent that globalization changes this relationship at a fundamental level, it alters the way in
which monetary policy influences inflation and output, i.e. the monetary transmission mechanism.1
While such changes do not necessarily reduce central banks’ control over the economy, they may
nevertheless require, as Yellen (2006) puts it, “some recalibration of policy responses.”
In this paper, we try to quantify changes in the monetary transmission mechanism which are likely
consequences of further globalization. To do so, we abstract from the effects of increasing financial
integration and focus on two specific aspects of increasing goods market integration. First, we consider increasing market integration reflected in increased openness of an economy measured by the
average import-to-GDP ratio. If openness is related to ‘home bias’ in preferences and globalization
induces an alignment of preferences, as discussed in Corsetti, Meier, and Müller (2007), economies
are likely to become more open in due course.2 Second, we consider increasing goods market integration resulting from a declining ability of firms to price discriminate across markets (‘local currency
pricing’ or ‘pricing to market’). As pricing to market requires barriers preventing spatial arbitrage,
further integration is likely to reduce these barriers, see Senay (1998).
We study the consequences of both changes for the international monetary transmission mechanism
within a two country dynamic stochastic general equilibrium (DSGE) model of the global economy.
In each country a continuum of imperfectly competitive producers specializes in the production of
intermediate goods. Monetary policy affects the real economy, because firms are adjusting prices
infrequently. Global goods markets are incompletely integrated, because i) a fraction of intermediate
goods producers is able to price discriminate across markets and invoices its exports in buyer’s currency; ii) final goods production is biased towards domestic intermediate goods such that the average
import-to-GDP ratio falls short of 50 percent which would be observed in the absence of home bias
given that countries are symmetric and of equal size.
1
In contrast, the notion that financial globalization via increased capital flows seriously impedes a central bank’s control
over interest rates is rejected by most observers, see Bernanke (2007).
2
In the limiting case of full globalization, consumption bundles may be identical across countries, see Corsetti et al.
(2007). Alternatively, openness may increase, because trade costs fall. While we assume no trade costs throughout, our
results are likely to hold in a setup where trade costs rather than home bias is reduced, see Obstfeld and Rogoff (2000a).
2
As in Gust, Leduc, and Vigfusson (2006), we assume that final goods are assembled on the basis of
an aggregation technology which induces demand functions for intermediate goods to display a nonconstant elasticity of substitution. This property induces strategic complementarities in price setting
of intermediate goods producers and determines the slope of the Phillips curve for any given level of
nominal rigidities. Moreover, as a result of these strategic complementarities import prices become
an important factor in the pricing decisions of domestic firms and the more so, the more open an
economy. As a consequence, increasing openness is bound to alter inflation dynamics.3 In addition,
we assume that prices, if reoptimized, are predetermined relative to monetary policy innovations.
Moreover, we assume that private expenditures for consumption and investment are predetermined as
well. In both countries monetary policy is characterized by an interest rate feedback rule whereby the
nominal short term interest rate is adjusted in response to producer price inflation (PPI) and domestic
absorption.
Given that we are interested in quantifying the effects which increased globalization exerts on the
monetary transmission mechanism, we require our DSGE model to give a quantitatively realistic
account of the actual transmission process. As a benchmark, we therefore compute impulse responses
to a monetary policy shock within an estimated vector autoregression (VAR) model. Specifically, we
use quarterly time series data for the U.S. relative to an aggregate of industrialized countries for the
period 1973 to date. Our VAR model includes consumption, investment, PPI-inflation, a short term
interest rate, consumer price inflation (CPI) and net export. To identify monetary policy shocks, we
assume that - in line with the predictions of our DSGE model - consumption, investment and PPIinflation are predetermined relative to short term interest rates, while consumer price inflation and net
exports are allowed to respond to monetary policy innovations. In other words, we employ a recursive
identification scheme such that monetary policy shocks are innovations to short term interest rates not
accounted for by current realizations of consumption, investment and PPI-inflation as well as by
past values of all the other variables included in the VAR model. We treat the impulse responses to
monetary policy shocks as a characterization of the actual monetary transmission mechanism.
We estimate the structural parameters of the DSGE model employing the minimum distance estimation strategy suggested by Rotemberg and Woodford (1997) and Christiano, Eichenbaum, and Evans
(2005). Specifically, we find values for the key parameters of the DSGE model by matching the implied impulse response functions to those obtained from the VAR model. We find that our model,
evaluated at the parameter estimates, is able to reproduce quite well the shape and magnitude of the
empirical impulse responses. For this to be the case, the model requires parameters to take values
3
Sbordone (2007) focuses on the effects of such a technology on the slope of the New Keynesian Phillips curve and finds
that increasing the number of traded goods has a non-monotonic effect on the slope of the Phillips curve. Guerrieri et al.
(2008) also focus on the New Keynesian Phillips curve and estimate it on the basis of single equation techniques. They find
that incorporating this mechanism improves its empirical performance considerably.
3
which are plausible given the evidence reported by other studies. Specifically, in line with results
reported by Bergin (2006) our estimates suggest that local currency pricing (LCP) is pervasive. Similarly, we find evidence that demand functions for intermediate goods are strongly curved and thus
quite distinct from what one would obtain under the special case of a constant elasticity of substitution.
Given that the model provides an empirically successful account of the actual transmission mechanism, it is well suited to counterfactual experiments. We study how the transmission of an exogenous
increase in the short term interest rate changes given an increase in openness and a decrease in the
fraction of firms engaged in LCP. Specifically, we consider an increase in the import-to-GDP ratio
from 12 percent, which corresponds to the average value for the U.S. in our sample period, to 30
percent, which constitutes a large, yet not implausible increase. The fraction of LCP firms which we
estimate to be close to 90 percent is assumed to decline to 60 percent.
We find that increasing openness only has little bearing on the effects of the monetary innovation
on domestic demand and inflation. This is the result of a high LCP share in the baseline scenario,
which isolates import prices from the exchange rate movements triggered by the monetary impulse.
Apart from preventing the expenditure switching mechanism from operating, the pervasiveness of
LCP also limits the possible effect of relative import prices on inflation dynamics. Consequently, it
is only by simultaneously increasing openness and lowering the extent of LCP, that we find fairly
strong changes in the monetary transmission mechanism. We find that the effect of the policy shock
on domestic absorption is reduced by about 15 percent, while the response of producer price inflation
is increased by about 25 percent. In a last experiment, we turn to the recalibration of monetary policy,
which would be necessary if monetary policy were to have the same effect on domestic output and
inflation in the counterfactual scenarios as in the baseline estimation. Our tentative results point
to considerable changes either in the systematic components of monetary policy or the size of the
monetary impulse.
The remainder of this paper is organized as follows. In section 2 we introduce the details of the model
economy. Section 3 presents time series evidence from the estimated VAR model and discusses the
estimation of the DSGE model. In section 4 we simulate the effects of globalization on the basis of
model-based counterfactual experiments. Section 5 offers a brief conclusion.
2 Model
In this section we suggest a two country business cycle model to study monetary policy transmission
in open economies. Most of the model features are standard. We assume that in each country there
is a continuum of intermediate good producers operating under monopolistic competition and being
constrained in price setting à la Calvo. We assume that a fraction of these firms invoice exports in
4
domestic currency (PCP-firms), while the remaining firms invoice exports in foreign currency (LCPfirms). There is a representative household in each country owning the capital stock which is rented
together with labor services to intermediate goods producers on a period-by-period basis. Adjusting
investment is costly. International financial markets are assumed to be complete.4
In each country final goods firms assemble domestic and imported goods to provide final goods
which are used for consumption and investment. At the final good level we assume an aggregation
technology which induces home bias in the composition of final goods and a non-constant elasticity
of substitution (NCES) in the demand for intermediate goods.5 This technology has recently been
put forward by Gust et al. (2006), Sbordone (2007) and Guerrieri et al. (2008) in order to introduce
strategic complementarities in price setting arising from the degree of competition in the intermediate
goods markets. While these authors stress the importance of this channel in accounting for the consequences of the ongoing globalization process and its impact on inflation dynamics, its implications
for monetary policy transmission have not yet been analyzed within a sticky price two country general
equilibrium framework.
Our model is suited to perform this task. It is outlined in the following by focusing, in turn, on the
problems of final goods firms, intermediate good firms and the representative household. We close
the model with a characterization of monetary policy in terms of an interest feedback rule. As both
countries i ∈ {1, 2} are symmetric, of equal size, and have isomorphic structures, our exposition
focuses on country 1, i.e. the ‘home’ country.
2.1 Final Good Firms
Let F1t denote final goods in country 1 produced at time t and used for private consumption, investment and government consumption (domestic absorption), i.e. F1t = C1t + X1t + G1t . Final goods
firms are perfectly competitive. The problem of a representative final goods firm is to assemble domestically produced intermediate inputs, A1t (j), as well as imported intermediate goods, B1t (j), to
produce a given amount of final goods. These inputs are produced by a continuum of intermediate
A (j) denote the price in country 1 of
goods firms in each country; we assume j ∈ [0, 1]. Letting P1t
B (j) denote the price in country 1 of a generic good
a generic good produced in country 1 and let P1t
produced in country 2, the final goods firm’s problem is given by
Z 1
Z 1
A
B
min
P1t (j)A1t (j)dj +
P1t
(j)B1t (j)dj
0
(1)
0
4
In setting up the model we draw on earlier work by Chari, Kehoe, and McGrattan (2002), Kollmann (2002), Galí and
Monacelli (2005), Corsetti and Pesenti (2005), Bergin (2006) and Schmidt (2006), among others.
5
The second feature is in contrast with earlier open macro models which often employ an Armingtion aggregator as, for
instance, in Backus, Kehoe, and Kydland (1994).
5
subject to
·
σ
h σ−1
σ−1 i
σ−1
σ
σ
Vdt + Vmt
−
¸
1
− 1 = 1,
(1 + η)υ
(2)
where Vdt is an aggregate of domestically produced individual goods and Vmt an aggregate of imported individual goods. These aggregates, in turn, are defined as follows
¸υ
·
Z 1
σ
1
(1 + η) A1t (j)
σ−1
− η dj,
Vdt =
ω
(1 + η)υ
ω
F1t
0
¸υ
·
Z 1
σ
1
(1 + η) B1t (j)
− η dj.
Vmt =
(1 − ω) σ−1
(1 + η)υ (1 − ω) F1t
0
(3)
(4)
This structure defined by (2), (3) and (4) specifies the technology available to the final goods firm and
follows Gust et al. (2006).6 A few remarks concerning the key parameters are in order. The trade price
elasticity, i.e. the elasticity which measures the extent of substitution from goods produced at home
to those produced abroad for a given change in relative prices, is a key parameter for the international
transmission mechanism. In our setup it is a function of several parameters and given by
σ̃ =
−σ
.
(σ(υ − 1) − υ)(1 + η)
(5)
The elasticity of substitution between goods produced within the same country is time varying. In the
steady state this elasticity is given by
²=
1
1
.
1−υ1+η
(6)
Note that the parameter η plays a crucial role for both elasticities. It provides a measure of how
strongly our setup deviates from the special case where the elasticity of substitution is constant
(CES).7 Finally, the parameter ω is often referred to as providing a measure for ‘home bias’ as it
measures the steady state weight of domestically produced goods in final goods. 1 − ω mesures the
import-to-GDP ratio in steady state.
Solving equation (1) gives rise to domestic demand functions for domestically produced intermediate
goods
1
AD
1t (j) = ω
1+η
"µ
A (j)
P1t
A
P1t
1
¶ υ−1
µ
A
P1t
Γ1t
#
σ
¶ σ(υ−1)−υ
+ η F1t .
(7)
In the same manner, final goods producers in the foreign country minimize expenditures, which in turn
implies the following demand function for domestically produced intermediate goods in the foreign
country
AD
2t (j)
6
7
1
= (1 − ω)
1+η
"µ
A (j)
P2t
A
P2t
1
¶ υ−1
µ
A
P2t
Γ2t
σ
¶ σ(υ−1)−υ
#
+ η F1t ,
(8)
The original closed economy formulation goes back to Dotsey and King (2005) or more generally to Kimball (1995).
The CES case is nested in our setup for η = 0.
6
where Γit are price indices defined below. Note that as in Dotsey and King (2005) and the corresponding open economy model of Gust et al. (2006) the demand curves in equations (7) and (8) include a
linear term if η 6= 0 implying the demand elasticities are not constant.8 Global demand for a generic
good j produced in country 1 is given by
D
Y1tD (j) = AD
1t (j) + A2t (j).
(9)
To understand the price indices implicitly defined by the cost minimization problem of the final goods
firm, note that pricing behavior at the intermediate good level will be different depending on whether
A,LCP
(j) denote the price in country 1 of
intermediate goods firms engage in LCP or PCP. Let P1t
A,P CP
a generic good produced in country 1 by a LCP-firm and P1t
(j) the price set by a PCP-firm.
Letting α measure the fraction of LCP-firms and (1 − α) the fraction of PCP-firms, then the producer
A and the import price index P B in the home country are given by
price index (PPI) P1t
1t
µZ
A
P1t
=
0
and
µZ
B
P1t
α
=
0
α
Z
υ
A,LCP
P1t
(j) υ−1 dj
+
α
Z
υ
B,LCP
P1t
(j) υ−1 dj
1
1
+
α
υ
A,P CP
P1t
(j) υ−1 dj
υ
B,P CP
P1t
(j) υ−1 dj
¶ υ−1
υ
(10)
¶ υ−1
υ
,
respectively. Cost minimization also implies the following price index for the final good
µZ α
¶
Z 1
1
η
A,LCP
A,P CP
F
P1t =
Γ1t +
ω
P1t
(j)dj +
P1t
(j)dj
1+η
1+η
0
α
µZ α
¶
Z 1
η
B,LCP
B,P CP
+
(1 − ω)
P1t
(j)dj +
P1t
(j)dj ,
1+η
0
α
(11)
(12)
(13)
where Γ1t is given by
h
i σ(υ−1)−υ
(σ−1)υ
(σ−1)υ
(σ−1)υ
A σ(υ−1)−υ
B σ(υ−1)−υ
.
Γ1t = ω(P1t )
+ (1 − ω)(P1t )
(14)
Letting St denote the nominal exchange rate (the price of home currency in terms of foreign currency)
and assuming that the law of one price holds for PCP-firms, we have
B,P CP
B,P CP
P1t
(j) = St P2t
(j);
A,P CP
A,P CP
P1t
(j) = St P2t
(j).
(15)
2.2 Intermediate Goods Firms
In each country, there is a continuum of firms each of which produces a unique intermediate good
and engages in monopolistic competition. Production is Cobb-Douglas:
Y1t (j) = K1t (j)θ H1t (j)1−θ ,
(16)
8
As a result, the optimal markup set by intermediate good firms will be time-varying. These strategic complementarities
have import implications for inflation dynamics, see Sbordone (2007).
7
where H1t (j) and K1t (j) respectively denote labor and capital employed by firm j . Let W1t and
R1t denote the nominal wage rate and the rental rate of capital, respectively. Minimizing the costs of
producing intermediate goods implies for (nominal) marginal costs
M C1t (j) =
W1t H1t (j)
R1t K1t (j)
=
,
(1 − θ)Y1t (j)
θY1t (j)
(17)
where marginal costs are independent of the level of production and identical across firms, because
both factors of production can be adjusted freely across firms.
We assume that price setting is constrained exogenously by a discrete time version of the mechanism
suggested by Calvo (1983). Each firm has the opportunity to change its price with a given probability
1−ξ . Moreover, we assume as, for instance, in Christiano et al. (2005) that when a firm has the oppor-
tunity, it sets the new price in order to maximize the expected discounted value of net profits before
the realization of shocks in a given period, i.e. period t prices are set conditional on the information
available in period t − 1. Firms that do not reoptimize in a certain period index their price to last
period’s producer price inflation, where the degree of indexation is given by the parameter κ ∈ [0, 1].
A,P CP
In setting the new price P1t
(j), the problem of a generic intermediate good PCP-firm j in country
1 is given by
max
∞
X
Ã
ξ k Et−1
"
A,P CP
P1t
(j)
D
Qt,t+1 Y1t+k
(j)
!
#
k
Y
F
ΠκP P I,1t+s−1 − M C1t+k /P1t+k
(18)
s=1
k=0
subject to demand functions defined by (9), the production function (16), the optimality condition on
factor inputs (17) as well as the constraint that the demand of intermediate good j is satisfied, i.e.
A /P A
Y1tD (j) = Y1t (j). ΠP P I,1t = P1t
1t−1 denotes producer price inflation. Profits are discounted with
the stochastic discount factor, Qt,t+1 .
A,LCP
The pricing problem of a generic intermediate good LCP-firm j is twofold. The firm sets P1t
(j)
for the domestic market by solving
max
∞
X
"
A,LCP
ξ k Et−1 Qt,t+1 AD
(j)
1t+k (j) P1t
k
Y
#
F
ΠκP P I,1t+s−1 − M C1t+k /P1t+k
,
(19)
s=1
k=0
A,LCP
subject to the demand function (7). P2t
(j) is set to maximize
max
∞
X
"
ξ
k
Et−1 Qt,t+1 AD
2t+k (j)
A,LCP
St+k P2t
(j)
k
Y
s=1
k=0
subject to the demand function (8).
8
#
ΠκP P I,2t+s−1
F
− M C1t+k /P1t+k
(20)
2.3 Households
A representative household in country 1 allocates consumption expenditures on final goods, C1t , and
supplies labor, H1t , to intermediate goods firms. The objective of the household is given by
E−1
∞
X
[(C1t − bC1t−1 )µ (1 − H1t )1−µ ]1−γ
β
,
1−γ
(21)
t=0
where β is a time discount factor and b ∈ [0, 1) measures the extent of consumption habits. The
parameter γ measures the degree of risk aversion and the parameter µ measures the weight of consumption in the utility function relative to leisure. We assume that decisions concerning period t are
conditional on the information available in period t − 1.
Labor and capital are internationally immobile; households in country 1 own the capital stock K1t
of that country. We follow Christiano et al. (2005) and assume that it is costly to adjust the level of
investment, X1t . Specifically, the law of motion for capital is given by
K1t+1 = (1 − δ)K1t + [1 − Ψ(Iit /Iit−1 )]Iit ,
(22)
where δ denotes the depreciation rate; restricting Ψ(1) = Ψ0 (1) = 0 and Ψ00 (1) = χ > 0 ensures that
the steady state capital stock is independent of investment adjustment costs captured by χ.
A complete set of state-contingent securities is traded at an international level. We let Ξ1t+1 denote
the period t + 1 payoff of the portfolio held at the end of period t in units of country 1’s currency.
For future reference it is useful to define the gross short-term nominal interest rate, it , given as the
solution to the following relationship: i−1
t = Et Qt,t+1 ; in other words, it is the inverse of the nominal
value of a security paying one unit of account in each state of the world in period t + 1. The budget
constraint of the household reads as follows
F
W1t H1t + R1t K1t + Υ1t + T1t − P1t
(C1t + X1t ) = Et {Qt,t+1 Ξ1t+1 } − Ξ1t ,
(23)
where Υ1t denotes nominal profits earned by monopolistic firms and transferred to households and T1t
denotes lump-sum taxes. Because we assume that government spending is financed entirely through
F G . The household problem is to maximize (21) subject to the
lump-sum taxes, we have T1t = P1t
1t
constraints (22) and (23).
2.4 Monetary Policy
To close the model, we assume that monetary policy is characterized by an interest rate feedback rule
as in Clarida, Galí, and Gertler (2000). Specifically, we assume for the interest rate
µ
µ
¶¶
F1t − F1
i1t = ρi1t−1 + (1 − ρ) i1 + β −1 φπ (ΠP P I,1t − ΠP P I,1 ) + 0.25β −1 φy
+ ν1t (24)
F1
where ρ ∈ [0, 1] captures interest rate smoothing, φπ captures the long-run adjustment of the interest
rate to producer price inflation and φy captures stabilization of domestic absorption.
9
2.5 Model solution
We solve the model numerically by applying standard techniques. Specifically, we linearize the
equilibrium conditions of the model, which are given by the constraints and the first order conditions
which characterize the problems described above, around a deterministic and symmetric steady state.
We assign values to the structural parameter of the model and solve the linear model numerically. We
then study the transmission of monetary policy shocks, ν1t . To keep the model tractable, we focus on
country differences, i.e. the behavior of a domestic variable relative to its foreign counterpart. Our
strategy to assign parameter values is described in the following section.
3 Estimation
Throughout the paper we focus on impulse responses to monetary policy shocks to characterize the
international monetary transmission mechanism. Our objective is that our model provides a quantitatively realistic account of the transmission mechanism as apparent from the data. To achieve this
objective, we proceed in two steps. First, we estimate the VAR model on U.S. time series relative
to an aggregate of industrialized countries and compute empirical impulse response functions to a
monetary policy shock. In a second step, we pin down the values of key parameters of the DSGE
model by matching the implied impulse response functions to those obtained from the VAR model.
3.1 Empirical impulse response functions
We estimate the VAR model on quarterly time series for the period 1973-2006. We focus on relative
variables, i.e. the difference of a variable in the U.S. and its counterpart for an aggregate of industrialized countries, which we treat as the rest of the world (‘ROW’ for short), see also Clarida and Gali
(1994) and Rogers (1999).9 This focus allows us to economize on the degrees of freedom, but is also
convenient given that we focus on country differences in the DSGE model as well.
Specifically, we consider the log of relative consumption, the log of relative investment, the difference
in PPI-based inflation rates, the difference in short term interest rates, the difference in CPI-based
inflation rates as well as the trade balance for the U.S., where the trade balance is defined as the log
difference in deflated exports and imports. Letting Yt denote the vector of endogenous variables, we
estimate the structural model
A(L)Yt = εt ,
where A(L) =
P4
i
i=0 Ai L , LYt
(25)
= Yt−1 and E(εt ε0t ) = I . To achieve identification, we assume
that A0 is lower triangular, i.e. we impose the recursive identification which is frequently employed
9
We provide a detailed description of the data in the appendix. We remove a constant linear trend from consumption and
investment before computing relative variables.
10
to study the effects of monetary policy shocks, see Kim (2001) for an open economy context. As
we only attach a structural interpretation to the innovation in relative short term interest rates, what
matters for identification is how the other variables in Yt are ordered relative to this variable, see
Christiano, Eichenbaum, and Evans (1999). We order relative consumption, relative investment as
well as the inflation differential based on PPI-inflation before and the inflation differential based
on CPI-inflation and net exports after the short term interest rate differential. Given this ordering,
our identification assumptions are consistent with the implications of the general equilibrium model
outline above. Consumption, investment and therefore domestic absorption as well as producer prices
are predetermined relative to monetary policy shocks, while consumer prices and the trade balance are
free to adjust immediately. Implicitly, as in the theoretical model, we are thus assuming that monetary
policy adjusts the short term interest rate in response to contemporaneous changes in producer price
inflation and domestic absorption, but not in response to contemporaneous changes in CPI-inflation
or the trade balance.10
Figure 1 displays the impulse responses to a (relative) monetary policy shock, i.e. an increase by
100 basis points in U.S. short rates relative to the aggregate of industrialized countries. The solid
line shows the point estimate. The shaded area shows 90 percent confidence bounds obtained from
bootstrap sampling. The upper row shows the responses of consumption and investment in relative
terms; for both we find a protracted and hump-shaped decline. While consumption falls by about 0.3
percent, investment falls about 1.25 percent, with the maximum effect occurring between three to six
quarters after the shock.
Producer price inflation (in relative terms) responds somewhat sluggishly; the maximum decline of
about 8 basis points is observed five quarters after the shock. According to our point estimate, it
takes another 3 to 4 years for inflation to return to its pre-shock level. The shock to the short rate
is somewhat persistent and the short rate returns to its pre-shock level after about one year. The
response of consumer price inflation is remarkably close to that of producer price inflation, both
from a quantitative and a qualitative point of view. Finally, the real net exports, measured as the
log difference of real exports and imports for the U.S. displays a hump-shaped increase with the
maximum effect of about 0.2 percent occurring after about a year.
10
Alternative approaches to identify monetary policy shocks in open economy frameworks focus on monetary aggregates and non-recursive identification schemes, see Eichenbaum and Evans (1995), Cushman and Zha (1997) and Kim and
Roubini (2000). More recently, Faust and Rogers (2003) and Scholl and Uhlig (2007) use sign restrictions to achieve identification. A common focus of this literature is the behavior of the real exchange rate in the face of monetary policy shocks
and how important are the latter to account for fluctuations in the former. In the present paper, we are not taking up these
issues. Instead, we use the VAR responses as a key statistic to pin down parameter values of our general equilibrium model.
11
Consumption
Investment
0.5
3
2
1
0
0
−1
−0.5
−2
−3
−4
−1
0
5
10
15
20
0
5
PPI-inflation
10
15
20
Short rate
2
0.2
1.5
0.1
1
0
0.5
−0.1
0
−0.2
−0.5
0
5
10
15
20
0
CPI-inflation
5
10
15
20
15
20
Real net exports
0.2
0.6
0.1
0.4
0
0.2
−0.1
0
−0.2
−0.2
0
5
10
15
20
0
5
10
Figure 1: Effect of monetary policy shock. Notes: Responses are in relative terms (U.S. vs. ROW), except for net
exports which is the log difference of U.S. exports and imports. Solid line: point estimate; shaded areas: bootstrapped 90
percent confidence intervals; dashed-dotted line: responses of estimated general equilibrium model; Vertical axes: percent,
except for inflation and interest rate (percentage points). Horizontal axes: quarters.
3.2 Estimation of general equilibrium model
The second step of the analysis consists in matching empirical (VAR) and theoretical impulse responses in order to obtain estimates for the parameters of our model. This approach has gained popularity in closed economy studies of monetary policy transmission following the pioneering work of
Rotemberg and Woodford (1997) and Christiano et al. (2005), see, for instance, Amato and Laubach
12
(2003), Bovin and Giannoni (2006) and Meier and Müller (2006).
To illustrate this approach, define Ψe to be the empirical impulse response function characterizing
the data. The model itself, in turn, assigns to each admissible vector of structural parameters θ a
theoretical impulse response function Ψt = Ψ (θ). The binding function, Ψ(), must be assumed to
be injective to ensure identification. We obtain an estimate for the parameter vector of interest, θb, by
minimizing the weighted distance between empirical and theoretical impulse response functions, i.e.,
Ψe and Ψt :
θb = arg min (Ψe − Ψ (θ))0 W (Ψe − Ψ (θ)) ,
(26)
where W represents a positive definite weighting matrix.
As the relationship between structural parameters and the implied impulse response functions is nonlinear, we obtain theoretical impulse response functions by applying standard numerical techniques.
Note that our procedure only admits saddle path stable solution and thus rules out by construction
any parameterization of the model which would give rise to equilibrium indeterminacy. Basically,
Ψ (θ) is evaluated repeatedly for different parameter vectors θ until the closest fit with the empirical
impulse responses, Ψe , has been obtained.
Our choice of the weighting matrix W is guided by the idea of giving greater weight to impulse
responses that are more precisely estimated. Thus we opt for the diagonal matrix W diag whose
diagonal entries are the reciprocal values of the variance of the empirical impulse responses. Using
this weighting matrix ensures that the theoretical impulse responses are made to be as close to the
empirical ones as possible, in terms of point-wise standard deviations. Finally, regarding the length
of the impulse response functions, we consider 20 quarters starting from the second quarter.
Standard errors for θb are computed using the following expression for the asymptotic variance of our
estimator, taken from Wooldridge (2002):
³ ´ ¡
´¡
¢ ³
¢
b G G0 W G −1 .
[ θb = G0 W G −1 G0 W ΣW
Avar
(27)
where G = ∇θ Ψt represents the Jacobian of the impulse response function generated from the model
b denotes the bootstrap-estimated variance matrix of the impulse responses.
and Σ
In practice, given the number of the structural parameters, it is not possible to identify all of them
simultaneously. We therefore fix a number of parameters prior to the estimation which are either given
by first moments of the data or largely uncontroversial in the literature. First we set ω = 0.88 which
implies an import-to-GDP ratio of 12 percent, about the average value for the U.S. in our sample
period. We set the discount factor β to 0.99. Following Backus et al. (1994), we assume γ = 2 and
µ = 0.34. Moreover, we set the capital share in intermediate goods’ production θ to 0.36. For the
depreciation rate we assume δ = 0.025. We assume that government spending accounts for 20 percent
of GDP, again close to the average in our sample period. Regarding price rigidities, we set ξ = 0.75,
13
Parameter
σ̃
Table 1: Estimated parameter values of DSGE model
Description
Trade price elasticity
0.50
χ
Investment adjustment costs
κ
Price indexation
1.00
φπ
Inflation coefficient in policy rule
1.00
φy
Output coefficient in policy rule
ρ
Interest rate smoothing
b
Habits
α
Share of firms with local currency pricing
η
NCES-parameter
(0.74)
1.01
(0.64)
(−)
(0.52)
0.01
(0.14)
0.67
(0.09)
0.89
(0.05)
0.89
(0.15)
−11.11
(15.23)
Notes: Parameter estimates obtained from matching DSGE and VAR impulse response functions; standard errors are reported in parentheses. Those parameter values which have been
estimated to be at their bounds have been assumed to take this value prior to estimation; in this
case no standard error is reported.
which implies an average duration of prices of one year which is relatively long given the micro
evidence discussed in Nakamura and Steinsson (2008), but still lower than most macroeconometric
evidence suggests. We set υ such that the markup earned by intermediate goods firms in steady state
is 20 percent. Finally, note that we estimate the trade price elasticity, σ̃ , by adjusting σ .11
3.3 Results
Table 1 provides the results of our estimation exercise, i.e. the solution to (26). We find quite plausible point estimates and fairly narrow confidence bounds implied by the standard errors reported in
parentheses. The estimated trade price elasticity is below the values often used or found in the literature. Yet, several recent studies suggest that that a low trade price elasticity may help to account for
a larger set of macroeconometric observations, see Lubik and Schorfheide (2006), Kollmann (2006)
and de Walque, Smets, and Wouters (2005). Also χ, the parameter capturing investment adjustment
costs is somewhat below the value reported in Christiano et al. (2005) for U.S. data. This is likely to
be the result of the aggregation function of final goods, see the discussion in Backus et al. (1994).
In line with earlier research we also find full indexation of prices, see, for instance, Meier and Müller
11
That is in the estimation we impose the following relationships:
υ=
η + (1/1.22)
σ̃υ(1 + η)
and σ =
.
1+η
1 + σ̃(υ − 1)(1 + η)
14
10
0
−10
−20
−30
−40
−50
−60
−3
−2
−1
0
1
2
Figure 2: Demand function for intermediate goods. Notes: Solid line: CES case (η = 0). dashed-dotted line:
NCES case (η = −11.1); Vertical axes: relative demand in percent; Horizontal axes: relative price in percent.
(2006) for the U.S. Regarding monetary policy we find parameter values which imply a fairly lose
monetary stance and a degree of interest rate smoothing which is in line with previous findings in the
literature, see, for instance, Clarida et al. (2000) for the U.S. Note that while we do not impose a bound
on the parameter φπ , our solution procedure rules out equilibrium indeterminacy by construction. We
find a considerable amount of habits in consumption, somewhat above the values reported in Smets
and Wouters (2005) both for the Euro area and the U.S.
The share of firms engaged in LCP is estimated to be quite high, in line with results reported in Bergin
(2006) who estimates this parameter to be at its upper bound.12 Finally, the estimate for the parameter
η provides a measure for the curvature of our demand functions. Our estimate is somewhat higher
than the values used in Gust et al. (2006) and Guerrieri et al. (2008). To assess the implication of this
estimate in terms of deviation from the standard CES case (η = 0), figure 2 displays the change in
demand for a generic good (vertical axis) resulting from a deviation in its relative price (horizontal
axis). The solid line shows the induced change in demand in case η = 0, which is constant. The
dashed line shows the induced change in demand for our estimate of η . Relative to the CES case,
our estimate implies strong non-linearities in the demand functions. If the relative price increases,
demand falls considerably more strongly, while, if the relative price falls, demand increases more
mildly. This induces strategic complementarities in price setting, which ceteris paribus, provides
firms with an incentive to adjust prices by small amounts if given the opportunity to do so.
Given the estimated parameter values, we compute the impulse response of the model and compare
them to those obtained from the VAR model. The dashed-dotted lines in the panels of figure 1 show
12
The extent of LCP relative to PCP has been the topic of a considerable controversy which is beyond the scope of the
present paper, see Betts and Devereux (2000) and Obstfeld and Rogoff (2000b).
15
0.4
0.35
0.3
0.25
0.2
0.15
0.1
0.05
0
1975
1980
1985
1990
1995
2000
2005
Figure 3: Import-to-GDP ratio in U.S. 1973. Notes: Solid line displays nominal imports divided by nominal
GDP; dashed-dotted line gives the average value; dotted line gives counterfactual scenario with import share of 30 percent.
that the model responses track the empirical responses quite closely. All the responses are within
the confidence bounds of the VAR responses, except for the impact response of CPI-inflation and net
exports. Also the theoretical response of investment is somewhat less pronounced than its empirical
counterpart. The response of the consumption differential, as well as those of PPI-inflation and the
short rate are matched particularly closely. Overall, we conclude that the DSGE model - if evaluated
at the point estimates - can quantitatively account for the international monetary transmission mechanism as apparent for the VAR model estimated on time series from 1973-2006. The DSGE model
seems thus well suited to study counterfactual outcomes in a more globalized world.
4 Model-based counterfactual analysis
We now turn to the question which motivates our investigation, namely, whether and to what extent
globalization alters the international monetary transmission mechanism. Given that our two country
general equilibrium model has been shown to give an empirically plausible account of the international transmission mechanism, it is well suited to address these questions. Notably, given that the
structural model parameters estimated above are fairly ‘deep’, we can perform counterfactual experiments which are not prone to the Lucas critique.
As discussed in the introduction we make the notion of globalization operational by considering two
specific parameter variations in the model. First, we consider an increase in trade integration resulting
in an increase in the import-to-GDP ratio from 12 to 30 percent. We implement this by lowering the
‘home bias’ parameter ω from 0.88 to 0.7.13 Figure 3 displays the evolution of actual U.S. imports
13
We maintain the assumption that both countries are symmetric.
16
relative to GDP for the period 1973-2006. The average value is close to 12 percent and indicated by
the dashed-dotted horizontal line. The import share is increasing over time, from about 6 to about 16
percent. Except for the last 10 years, however, the increase has been rather modest - in line with the
common perception that globalization has been accelerating only recently. The horizontal dotted line
indicates our counterfactual scenario of an import-to-GDP ratio of 30 percent. Given current trends,
this corresponds to a rather strong increase in trade integration, but is not utterly implausible.
The second variation concerns the extent of pricing to market (LCP). Currently, in order to mimic
the transmission mechanism apparent from the VAR impulse responses the model requires that a
considerable fraction of intermediate good firms is price discriminating across markets and invoices
exports in the currency of the destination market. As argued in the introduction, one manifestation
of globalization is an increase in possibilities to arbitrage spatially. Hence, we may expect that fewer
firms will be able to price discriminate across markets; put differently, we may expect the fraction
of PCP firms to increase. We therefore consider the counterfactual scenario that the fraction of firms
which engages in PCP increases from about 10 to 40 percent. We implement this scenario by lowering
α accordingly.
4.1 Counterfactual transmission
Figure 4 displays the results of our model-based counterfactual experiments. We focus on the response of key variables in country 1 to an increase in domestic interest rates by 100 basis points.14
As a baseline scenario we plot the response of the baseline economy, i.e. the model evaluated at the
parameter estimates obtained above (solid line). We consider three counterfactual scenarios. First,
we assume that imports account for 30 percent of GDP (dashed line, CF1); second, we assume that
40 percent of firms engage in PCP rather than LCP (dashed-dotted line, CF2). Finally, we assume
that both changes occur at the same time (dotted line, CF3). For these counterfactual simulations we
keep all the other parameters at the baseline values.
The upper row displays the responses of domestic absorption and output. Note that as government
spending is constant, the response of domestic absorption is the weighted sum of the responses of
consumption and investment. While we did not include output in the VAR model in order to achieve
model consistent identification through a recursive scheme, we can now use the estimated model
to assess the impact of monetary policy shocks on output as well. In the second row, we display
responses of PPI-inflation as well as the response of consumer prices. In the latter case, we focus
on the level response, as CPI inflation responses strongly undershoot on impact thereby preventing a
clear-cut interpretation of the counterfactual outcomes. In the third row we display the response of
14
While cross country differentials generally display a very similar pattern, focusing on home variables allows to derive
quantitative results which presumably matter most from the perspective of domestic policy making.
17
Domestic absorption
Output
0.05
0
0
−0.05
−0.05
−0.1
−0.1
−0.15
−0.15
−0.2
−0.2
−0.25
−0.25
−0.3
−0.3
0
5
10
15
−0.35
0
20
PPI-inflation
0
−0.02
−0.2
−0.04
−0.4
−0.06
−0.6
−0.08
−0.8
5
10
15
−1
0
20
Real exchange rate
0.2
2.5
0
2
−0.2
1.5
−0.4
1
−0.6
0.5
−0.8
0
−1
5
10
15
20
15
5
10
15
20
15
20
Real net exports
3
−0.5
0
10
Consumer price level
0
−0.1
0
5
−1.2
0
20
5
10
Figure 4: Effect of monetary policy shock in theoretical model.
Notes: Impulse response of domestic
variables to exogenous increase in short term interest rate (100 bp). Solid line: baseline estimate; dashed line: import-toGDP ratio is 30 percent (CF1); dashed-dotted line: 40 percent PCP (CF2); dotted line: import-to-GDP ratio 30 percent and
40 percent PCP (CF3); Vertical axes: percent, except for inflation and interest rate (percentage points) and real net exports
(percentage points of GDP). Horizontal axes: quarters.
the real exchange rate for which an increase indicates an appreciation as well as real net exports.15
From a qualitative point of view, the counterfactual scenarios imply a transmission of monetary policy
shocks which is fairly close to the one implied by the estimated model. Yet, important quantitative
15
We do not report the response of the short interest rate, because it is virtually unaffected by the counterfactual experiments.
18
Domestic absorption
Ex ante real interest rate
−0.18
0.5
−0.2
0.45
−0.22
0.4
−0.24
0.35
−0.26
0.3
−0.28
−0.3
0
0.2
0.4
0.6
0.8
1
0.25
0
0.2
0.4
0.6
0.8
1
Figure 5: Peak responses of domestic absorption to monetary policy shock and corresponding response of ex ante real interest rate. Notes: vertical axis measures peak response of domestic absorption (left) and
response of ex ante rate rate in third quarter (right); solid line corresponds to import share of 12 percent; dashed line corresponds to import share of 30 percent; dashed dotted line corresponds to import share of 40 percent; horizontal axes measures
extent of PCP; dashed vertical line corresponds to estimated extent of PCP; dotted vertical line corresponds to extent of PCP
assumed in counterfactuals computed above.
differences can be observed. Considering the first counterfactual, one observes small differences
relative to the baseline responses. With the exception of output, which falls less than in the baseline
case, increasing trade openness has little bearing on the transmission mechanism. Similarly, if only
the extent of LCP is reduced the quantitative changes are contained, except for net exports (CF2).
In contrast, under the third counterfactual scenario, we observe considerable changes in the response
of domestic absorption; its peak response is reduced by some 15 percent. Also the response of PPIinflation increases by some 25 percent (similarly the consumer price level falls much more strongly).
Finally, net exports tend to fall, rather than rise in the counterfactual scenarios (CF2 and CF3).
To build up intuition for the mechanisms which bring about these changes and to assess the quantitative differences more systematically, we plot in figure 5 the peak responses of domestic absorption
against the increasing fraction of PCP firms (left panel) together with the corresponding response of
the ex ante real interest rate. More specifically, on the horizontal axis we increase PCP from zero to
one. On the vertical axis we display the peak responses for three different assumptions on the importto-GDP ratio. The solid line corresponds to our baseline case with imports accounting for 12 percent
of GDP, while the dashed (dashed-dotted) line displays results obtained assuming 30 (40) percent.
We observe that the higher the fraction of PCP firms, the lower the increase in the ex ante real rate.
One way to think about this finding is by recalling that the ex ante real interest rate differential
corresponds to the expected change in the real exchange rate, see Clarida and Gali (1994). As apparent
from figure 4, the real exchange rate appreciates less strongly, the more PCP and the more open the
economy. Intuitively, the more open the economy and the more responsive prices are in buyers’
currency, the less strong the response of international relative prices to bring about an reallocation of
19
expenditure. Also the change in the real exchange rate and, hence, the response of the ex ante real
rate is smaller in the counterfactuals with more PCP and a higher import-to-GDP ratio. Intuitively,
the appreciation lowers the price of imports and thus, ceteris paribus, lowers the price of consumption
today relative to tomorrow. Corsetti et al. (2007) provide a lengthy discussion of this mechanism
in the context of fiscal policy transmission. As a consequence, monetary policy’s leverage over real
interest rates decreases with PCP and the more so, the more open the economy. In other words,
it appears that with the ongoing globalization process, the effects of monetary policy on domestic
absorption become more contained.
A similar consideration serves to rationalize the change in the response of real net exports displayed in
the lower right panel of figure 4. In the period of the shock, domestic absorption is predetermined such
that any change in real net exports reflects substitution from domestic and foreign goods triggered by
an exchange rate appreciation. However, with PCP being pervasive, the exchange rate pass-through
and, hence, expenditure switching is limited. Only if PCP increases, one observes substantial changes
in the international transmission mechanism as far as the response of net export is concerned.16
A key relationship taht shows how inflation dynamics are triggered by the monetary shock is given
by the Phillips curve. It is also pivotal for interpreting the results of our counterfactual analysis.
Drawing on earlier work by Guerrieri et al. (2008), we start from the price stetting problem of a
generic intermediate good firm and derive a dynamic relationship between PPI-inflation, marginal
cost and relative import prices (with first order accuracy expressed in deviations from steady state).
In other words, our model gives rise to the following variant of the the new Keynesian Phillips curve
¸
·
σ̃ B
(28)
πt − κπt−1 = βEt−1 [πt+1 − κπt ] + λEt−1 (1 − Ψ)st + Ψ(1 − ω) qt
²
where
λ=
(1 − βξ)(1 − ξ)
−ηµ
and Ψ =
.
ξ
1 − ηµ
Here πt , st and qtB measure PPI-inflation, real marginal costs and the relative import price relative
to their steady state values. As stressed by Guerrieri et al. (2008), if η > 0, i.e. in case the price
elasticity of substitution implied by the demand functions (7) is non-constant, relative import prices
directly determine inflation in addition to expected future inflation rates and real marginal costs.
Intuitively, with η > 0 if imported goods are relatively cheap those domestic firms which are able to
adjust their prices will tend to reduce them because of strategic complementarities. In contrast, in the
CES case, i.e. if η = 0 we have Ψ = 0, such that inflation dynamics are governed by marginal costs
only, while the slope coefficient λ depends only on the extent of price rigidities, ξ , as in the standard
new Keynesian Phillips curve. As a consequence, changing the import share, i.e. lowering ω alters
16
B
F
Formally, we observe that the relative price of imports, P1t
/P1t
, falls much more strongly if LCP is reduced; it falls
less, the more open an economy (quantitatively the first effect is much stronger. Results are available on request).
20
the slope coefficients in the Phillips curve only in the NCES case. Against, this background, a large
value of η as implied by our estimates is likely to enhance the possible effects of increased openness,
i.e. lower values of ω , on inflation dynamics.
PPI inflation
Relative import price term
−3
−0.04
0
−0.05
−1
−0.06
−2
−0.07
−3
−0.08
−4
−0.09
−5
−0.1
−6
−0.11
−7
−0.12
0
0.2
0.4
0.6
0.8
1
x 10
−8
0
0.2
0.4
0.6
0.8
1
Figure 6: Peak responses of inflation (PPI) to monetary policy shock and corresponding response of
relative import price term. Notes: vertical axis measures peak response of inflation (left) and response of relative
import price in fifth quarter (right); solid line corresponds to import share of 12 percent; dashed line corresponds to import
share of 30 percent); dashed dotted line corresponds to import share of 40 percent horizontal axes measures extent of PCP;
dashed vertical line corresponds to estimated extent of PCP; dotted vertical line corresponds to extent of PCP assumed in
counterfactuals computed above.
Our counterfactual responses suggest that the effects of monetary policy shocks on PPI-inflation tend
to increase in all three scenarios. The left panel of figure 6 illustrates this more systematically. It
shows that the peak effects become stronger in the fraction of PCP starting from the point estimate
(dashed vertical line) and more so, the more open an economy. It is noteworthy, however, that increasing openness only matters for the peak response if there is a considerable amount of PCP firms.
Increasing only the import-to-GDP ratio while maintaining the assumption on the fraction of PCP
firms (the first counterfactual), induces little change in the peak effect on inflation.
The Phillips curve relationship (28) provides some guidance in interpreting these findings. Specifically, investigating how the responses of marginal costs and relative import price changes in PCP and
openness (not shown) reveal that they respond less strongly and are thus of little help in accounting for
the stronger inflation response.17 Therefore, we investigate the behavior of the relative import price
term, i.e. the response of relative import prices multiplied with the slope coefficient in the Phillips
curve. The result which is displayed in the right panel of figure 6 supports the notion that increasing openness induces a stronger response of PPI-inflation to monetary policy shocks, because the
cheaper import prices exert a disinflationary pressure on domestic prices in the presence of strategic
complementarities. It is interesting to note that as relative import prices per se do not change much
17
A suggestive explanation for the weaker effects on marginal costs is that the more open the economy, the more strongly
tends the real appreciation to lower real marginal costs because domestic factor prices are deflated with the CPI.
21
in the PCP share or the degree of openness, this finding is driven by the slope coefficient on relative
import prices in the Philips curve relationship. Moreover, note that the effect of increasing openness,
i.e. lowering ω , is magnified by the parameter Ψ the absolute size of which depends on η . Hence, it
appears that relative import prices will gain increasing importance in our globalization scenario and
alter inflation dynamics, because we find strong complementarities in prices setting resulting from
the NCES specification.
4.2 Recalibrating monetary policy
We finally address the issue whether in the counterfactual globalization scenario monetary policy can
maintain control over key variables and to what extent monetary policy needs to be recalibrated if it
wishes to induce the same effects on these variables. Clearly, our framework allows us to study only
the transmission of monetary policy shocks and not how monetary policy impacts on the transmission
of other sources of business cycle fluctuations. Note also that we take an entirely positive perspective,
because we currently limit simulations to have first order accuracy only and are therefore not in the
position to study questions of optimal policy.18
Nevertheless, with the limitations of our framework in mind, we can assess quantitatively, which
adjustment in the monetary policy rule would bring about the same effects of a monetary policy shock
by 100 basis points in our counterfactual scenarios as in the baseline parametrization. We focus on the
maximum effect on inflation and domestic absorption by a monetary policy shock under the baseline
scenario, which is at -0.07 percentage points for inflation -0.29 percent for domestic absorption. For
all three counterfactual scenarios we compute parameter values for the interest rate feedback rule (24),
φπ and φy , such that the monetary impulse brings about the same effect on inflation and domestic
absorption as in the baseline scenario. In other words, we ask what changes in the systematic part of
monetary policy would ensure monetary policy shocks to have the same effects in a more globalized
world as in the baseline scenario.19 Columns 2 and 3 of table 2 report the results.
We find that the coefficient on inflation in the interest rate feedback rule, φπ increases strongly, while
the coefficient on domestic absorption φy declines. Intuitively, as inflation tends to respond more
strongly in our counterfactuals monetary policy has to be more aggressively in order to bring about
the same effect on inflation. In contrast, domestic absorption is more stable in the globalized economy
compared to the baseline case due to reasons highlighted above. Therefore, the monetary authority
even needs to act destabilizing to achieve a domestic absorption reaction as in the baseline case.
18
Note, moreover, that to take up questions of optimal monetary policy we would need to take a stance on the precise
nature of business cycle fluctuations which we can currently avoid.
19
Our procedures is similar in spirit to our estimation exercise discussion in section 3. We minimize the distance between
the baseline impulse response function peaks and the respective counterfactual impulse response peaks for PPI-inflation and
domestic absorption. Holding fix the interest rate smoothing parameter ρ at the baseline estimation we achieve a unique
solution for φπ and φy that recovers the peaks of the baseline responses.
22
Baseline
CF1
CF2
CF3
φπ
1.00
1.86
2.41
5.76
Table 2: Monetary Control Parameters
φy
σν1t (PPI-Peak)
0.01
1
−0.28
0.97
−0.45
0.91
−1.39
0.80
σν1t (DA-Peak)
1
1.03
1.04
1.13
Notes: CF1: import share is 30 percent; CF2: 40 percent PCP; CF3: 30 percent import share and 40 percent PCP.
Column 2 and 3 report variation in policy rule coefficients to bring about
The parameter values confirm that globalization requires adjustments (‘recalibration’) of monetary
policy, see, for example Yellen (2006). Our results complement the findings by Woodford (2007)
who analyzes three additional features of globalization and concludes that monetary authorities do
not loose any control over inflation. In a sense, our finding suggest that central bank obtain better
control over inflation. Another way to assess this change, we report in column 4 of table 2: the size
of the monetary policy shock which induces the same peak response of PPI-inflation in a globalized
economy as a 100 basis points shock in the baseline economy. We find that the size of the necessary
monetary impulse declines. Note, however, that as domestic absorptions responds less in the counterfactual scenario, monetary policy would need to engineer a larger stimulus in order to generate the
same effect as in the baseline case, see column 5 of table 2.
5 Conclusion
In this paper, we ask whether globalization alters the monetary transmission mechanism and if so,
to what extent monetary policy responses need to the recalibrated in order to engineer the same effects on domestic absorption and inflation for a given monetary policy shock. We proceed as follows.
First, we suggest a general equilibrium model featuring two symmetric countries and several frictions which recent business cycle research has found to key to account for several macroeconometric
observations. Most importantly, we embed a final good aggregation technology into our general equilibrium model, which induces non-constant demand elasticities for intermediate goods. This feature
has been put forward by Gust et al. (2006), Sbordone (2007) and Guerrieri et al. (2008), drawing on
earlier closed economy work by Kimball (1995) and Dotsey and King (2005). In contrast to these
studies, we explore its consequences in a full-fledged open economy general equilibrium model with
sticky prices.
To pin down parameter values of the key parameters of the model, we first estimate a VAR model on
U.S. time series data relative to an aggregate of industrialized countries for the post Bretton-Woods
period 1973-2006. We identify monetary policy shocks in the VAR model imposing an identification
23
scheme which is consistent with our theoretical model and trace out the transmission mechanism by
studying impulse response functions. In a second step, we find parameter values of the general equilibrium model by matching its implied impulse responses to those obtained from the VAR. We find
that the estimated model is generally able to mimic the empirical response functions quite closely
and conclude that it is able to give an empirically plausible account of the actual transmission mechanism. Most importantly, we find that the model requires that most of the firms engage in LCP and
that demand elasticities are very responsive to deviations from steady state.
Hence, the model is well suited to perform counterfactual experiments. Specifically, we study three
experiments. First, we increase the import-to-GDP ratio from 12 percent (the assumption maintained
during the estimation) to 30 percent, as increased trade integration is a likely manifestation of globalization. Second, we lower the fraction of firms engaged in PCP, thus reducing the extent of price
discrimination across countries. Our estimates suggest that currently LCP is quite pervasive; however, globalization through reducing barriers which prevent arbitrage may induce a rise in the number
of PCP firms. In a third counterfactual scenario we study the effect of both changes occurring at the
same time.
Focusing on the effects of a monetary policy shock on domestic absorption and PPI-inflation in the
counterfactual scenarios relative to the estimated model, we find that globalization alters the transmission mechanism quantitatively. Most strongly so in our third counterfactual, i.e. if both LCP is reduced and the openness increased simultaneously. Intuitively, only if the exchange rate pass-through
into import prices increases (as LCP falls), the extent of trade openness becomes more important for
monetary transmission. Most interestingly, if a considerable amount of firms engages in PCP, the role
of relative import prices for inflation dynamics increases considerably with openness.
Finally, we show that monetary policy may still induce the same response of inflation and domestic
absorption to a given monetary impulse (in terms of peak responses). First, by adjusting its endogenous response to the economy, monetary policy may at the same time achieve the same effects on
inflation and domestic absorption in the counterfactual scenarios as in the baseline scenario. Alternatively, it may alter the size of the monetary impulse and thus achieve the same effect on either
inflation or output. Clearly, our investigation is limited to the transmission of monetary policy shocks
and globalization is likely to alter the transmission of other sources of business cycle fluctuations as
well. This may, in turn, require other adjustments in the conduct of monetary policy. We leave these
questions as well as questions of optimal monetary policy for future research.
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