kydland zarazaga 1997 businesscyclearg
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FEDERAL RESERVE BANK OF DALLAS 21 ECONOMIC REVIEW FOURTH QUARTER 1997
Societies w ould prefer a steady grow th
path for their national incom e of, say, 3 p ercent
every year to one that delivers a 3 percent
grow th rate on average, but w ith zigzags from ,
say, 12 percent one year to –6 percent the next.
Consequently, they typically dem and that
policym akers elim inate undesired econom ic
fluctuations.1 It is not surprising, then, that the
understanding of business cycles has alw ayscaptured the interest of econom ists and has
inspired som e of their best w ork.
The w ork of John M aynard K eynes and
M ilton Friedm an w ent a long w ay in defining
the term s and identifying the issues that a suc-
cessful theory of econom ic fluctuations ought to
address. D espite the m uch-advertised difference
betw een the schools of thought inspired by
these scholars, their w ork agrees on som ething
very im portant: nom inal factors, such as the
m oney supply, interest rates, and price rigidi-
ties, play the m ost im portant role in explaining
econom ic fluctuations.As is w ell know n, the 1970s w ere not kind
to the K eynesian interpretation of business cycles.
This interpretation predicts that the rising in-
flation rates of that decade should have been
associated w ith declining unem ploym ent rates,
not w ith the rising rates actually observed.
Em pirical and theoretical research did not treat
the “rival”school m uch better. Sim s (1980), for
exam ple, show ed evidence that seem s to con-
tradict som e versions of the m onetarist theory.
Initially, the theoretical developm ents
inspired by these failures kept nom inal factors
as the param ount force behind econom ic fluc-
tuations. In fact, in Lucas (1972), the first and
perhaps m ost celebrated application of the
novel approach to m acroeconom ic analysis for
w hich Robert Lucas received the 1995 N obel
Prize, the m oney supply still plays a crucial role
for the business cycle. Thus, econom ists w ere
surprised w hen K ydland and Prescott (1982)
show ed that one could account for tw o-thirds of
the U .S. econom ic fluctuations w ith a dynam ic
stochastic general equilibrium m odel from
w hich nom inal variables w ere totally absent—
that is, a m odel w ithout any m oney in it.K ydland and Prescott obtained this result
using a variation of the sam e basic theoretical
m odel econom ists had been using tim e and
again to study econom ic grow th issues.
U nifying theories—that is, theories that can
sim ultaneously explain seem ingly unrelated
phenom ena—are usually w elcom e in science.
W hat m any econom ists found attractive about
the Real B usiness Cycle (RBC) theory proposed
by K ydland and Prescott w as that, for the first
Is the BusinessCycle of Argentina
“ Dif ferent” ?
Finn E. KydlandProfessor
Carnegie Mellon University
and
Research Associate
Federal Reserve Bank of Dallas
Carlos E. J. M. Zarazaga
Senior Economist and Executive Director
Center for Latin American Economics
Federal Reserve Bank of Dallas
Nominal factors do not seem
to be able to account for any
significant fraction of the
business cycles of Latin
American countries in general,
and of Argentina in particular.
Perhaps for this reason it is time
to give real factors their fair
chance to do the job.
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tim e, a business-cycle theory pointed to the
possibility that the sam e analytical tools used to
address econom ic grow th issues could be used
to address business-cycle questions as w ell. This
m ay explain w hy these econom ists regarded
K ydland and Prescott’s findings persuasive
enough to b egin seriously exploring the
hypothesis that “real”factors, rather than nom i-
nal ones, are a prevalent driving force behindeconom ic fluctuations.2 Although real or supply-
side factors, such as the am ount of resources
used by the governm ent, tax p olicies, techno-
logical changes, governm ent regulations, m odi-
fications of financial interm ediation rules, and
even political shocks signaling p ossible changes
in property rights, m ay appear to be the obvi-
ous candidates to explain business cycles, this
w as not that clear a short w hile ago.
The process of verifying, sharpening, or
refuting the real-shock account of business
cycles has generated a large body of theoretical
and em pirical research concentrated, so far,on developed countries. This is unfortunate, be-
cause the evidence suggests that econom ic fluc-
tuations are particularly severe in developing
countries. U nderstanding w hy this occurs could
lead to w ays to m ake the business cycles of
these countries at least as sm ooth as those of
developed ones. W hat m akes the study of Latin
Am erican countries’business cycles p articularly
interesting is the claim that econom ic fluctua-
tions in those countries have been driven by
nom inal factors. Science m akes progress pre-
cisely w hen it encounters observations that the
prevailing paradigm cannot explain. Therefore,
there seem s to be a com pelling need to confirm
the alleged anom alies by answ ering the ques-
tion, Are business-cycle regularities in Latin
Am erica really all that different from those in
the U nited States and in O rganization for Eco-
nom ic Cooperation and D evelopm ent (O EC D )
and other European countries?
This article focuses this question on
Argentina, w ith the hope of m aking a m odest
contribution to the understanding of the busi-
ness cycles of Latin A m erican countries in
general. For exam ple, if Argentina’s business-cycle regularities are sim ilar to those of the
U nited States or Europe, then the business
cycles of all these countries m ay be m anifesta-
tions of essentially the sam e phenom enon.
Therefore, real factors could play an im portant
role in accounting for Argentina’s business
cycles, just as, according to recent research,
they do in the U nited States and Europe.
By contrast, if Argentina’s business cycles
show im portant anom alies w ith respect to the
evidence available for other countries, then the
possibility of real factors playing an im portan
role in its business cycle dim inishes. In this
case, existing interpretations em phasizing the
role of nom inal variables in Latin Am erica m ay
regain the prom inence they had in business
cycle theories until the 1970s. Allow ing for
com parisons w ith the em pirical evidence fo
other countries, this article exam ines theA rgentinean business-cycle regularities w ith
the sam e m ethodological approach used in
previous studies for the U nited States and sev
eral European countries.
In the follow ing section, w e present the
evidence other authors have used to support the
contention that nom inal factors have driven the
business cycles in Latin Am erica and provide
reasons to doubt the robustness of those find
ings. W e also suggest that the data require fur
ther system atic scrutiny before econom ists can
conclude w ith som e confidence that busines
cycles in Latin A m erican countries, and particularly in Argentina, differ in nature from those
observed in the U nited States and in O EC D and
other European countries. N ext, w e undertake
one such system atic study by presenting, as the
availability of data perm its, the A rgentinean
counterpart of the statistics researchers have
used to describe the business cycles of the
U nited States and several European countries
W e then com pare the statistics for Argentina
w ith those of other countries and state the
im plications that result from analysis of cross
country sim ilarities and differences. The las
section sum m arizes our conclusions.
The state of the business-cycledebate in Latin America
The understanding of the Latin A m erican
business cycles has not escaped the view tha
nom inal shocks are the predom inant cause o
econom ic fluctuations. This view still influences
the thinking on m any Latin A m erican econom ic
problem s. This thinking is particularly notice
able in the inflation stabilization literature.
O ne of the m ost serious econom ic prob
lem s m any Latin Am erican countries have facedin past decades has been persistent, high infla
tion.3 Therefore, the quest to find the best anti
inflation policies has inspired a large body o
research on this problem . The m onetarist influ
ence in that literature is evident in its contention
that nom inal factors (such as changes in the
nom inal exchange rate regim e) w ere the only
system atic force driving econom ic fluctuations
around the tim e the stabilization program s w ere
im plem ented. For exam ple, the conventiona
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FEDERAL RESERVE BANK OF DALLAS 23 ECONOMIC REVIEW FOURTH QUARTER 1997
w isdom in Latin Am erica is that anti-inflation
program s using the exchange rate as a nom inal
anchor (exchange-rate-based stabilization, or
ER BS, program s) have been able to reduce the
inflation rate w ithout causing the initial output
losses associated w ith program s that use som e
m onetary aggregate as a nom inal anchor
(m oney-based stabilization program s).4
O f course, a theory for stabilization pro-gram s is not the sam e as a theory for the
business cycle. But there should be som e con-
sistency am ong them . For exam ple, a finding
that nom inal shocks do not have im portant real
effects during Latin Am erican stabilization pro-
gram s w ould m ake it harder to m aintain the
m onetarist view that such factors m ay have
been im portant at any other point of the busi-
ness cycle. A nd this is precisely w hat w e find
problem atic: a reexam ination of the evidence
on ERBS program s show s that it is far from clear
that the adoption of the exchange rate as a
nom inal anchor has been responsible, as theliterature claim s, for the econom ic fluctuations
observed during those program s.
Figure 1 illustrates the consum ption
grow th rates for the ten ERBS program s studied
by V égh (1992). The vertical line indicates the
year or quarter in w hich the ERBS program
started.5 Casual inspection of the plots suggests
that only in the first four cases did consum ption
exp erience the upw ard jum p that theory pre-
dicts should occur upon announcem ent of ERBS
program s.6 H ow ever, this theoretical prediction
did not m aterialize in the rem aining six cases. In
particular, in none of these six did consum ption
grow faster than in the im m ediately preceding
period. Instead, in four of the six cases, con-
sum ption grow th w as basically the sam e im m e-
diately before and im m ediately after the
announcem ent of the corresponding ERBS pro-
gram . In tw o of the four, the so-called con-
sum ption boom preceded the announcem ent.
In the other tw o, there w as no consum ption
boom w hatsoever: consum ption continued
falling at approxim ately the sam e rate as before
the ERB S program s began. Furtherm ore, in the
last tw o cases, the ERB S program w as follow edinstead by a consum ption bust.
Therefore, the tim ing, intensity, or direc-
tion of consum ption grow th for the countries in
Végh’s study, after m ost ERBS program s began,
appears to differ from that im plied by the ERBS
theory.
In this sense, at least four of the p lots in
Figure 1 (Chile, February 1978; Argentina, D e-
cem ber 1978; Argentina, June 1985; and Israel,
July 1985) could be interpreted using the non-
m onetarist approach: the dynam ics of output
im m ediately after the announcem ent of an ERB S
program w ere m ere continuations of upsw ings or
dow nturns that had begun earlier. In these four
cases, forces other than the adoption of a fixed
or pegged exchange rate w ere already driving
the business cycle w hen the ERBS program s
began. But such conclusions from the casual
reading of tw o-dim ensional plots w ould be pre-m ature.7 W e are m ore persuaded, instead, by
the m ore thorough em pirical effort of Rebelo
and V égh (1995), w ho conclude that m onetarist-
inspired theoretical m odels of ERBS program s
are quantitatively incapable of rep licating any
significant fraction of the econom ic fluctuations
associated w ith such program s.
The evidence on ERBS program s, both
from casual plot readings and from the w ork of
Rebelo and Végh, poses a serious challenge to
m onetarist theories of Latin Am erican business
cycles: if nom inal exchange rate shocks in Latin
Am erica seem to have failed to produce thenoticeable and consistent effects on consum p-
tion and other real variables predicted by m one-
tarist-inspired theories precisely w hen they
w ere given the best shot at it, how could they
have significant real effects at other tim es?8
A natural next step in the research agenda
is to pay m ore attention to real shocks as a
potentially im portant source of the econom ic
fluctuations observed in Latin A m erican coun-
tries, including fluctuations observed during
inflation stabilization program s.9 In principle,
there is no reason the assessm ent of the quanti-
tative im portance of such shocks in Latin
Am erica could not be accom plished w ith the
sam e kind of dynam ic stochastic general equi-
librium m odels the RB C tradition has used to
that effect for the U nited States and other devel-
oped countries.
But such a research program m ust start by
describing the data w ith a system atic, atheoreti-
cal m ethodology.10 The rem aining sections of
this article m ake a m odest attem pt in that direc-
tion by describing the business-cycle regularities
of Argentina w ithout im posing theoretical priors
to the data.11
Business-cycle regular it ies f or Argenti naSome caveats about the data. N ational
account data in Latin Am erica are not as reliable
as their U .S. and O ECD counterparts.12 In fact,
because of frequent m ethodological changes
and corrections of previous errors, the reported
series m ay change substantially from one
national account estim ate to the next. This is
indeed the case for Argentina. For exam ple,
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Figure 1
ERBS Programs
Argentina, March 1967 Uruguay, June 1968
–20
–15
–10
–5
0
5
10
15
’83:1’82:1’81:1’80:1’79:1’78:1
Uruguay, October 1978 Brazil, February 1986
SOURCE: Table 10 in Végh (1992). SOURCE: Table 12 in Végh (1992).
SOURCE: Table 5 in Végh (1992). SOURCE: Table 7 in Végh (1992).
Argentina, June 1985 Israel, July 1985
Brazil, March 1964 Mexico, December 1987
SOURCE: Table 6 in Végh (1992). SOURCE: Table 14 in Végh (1992).
Chile, February 1978 Argentina, December 1978
SOURCE: Table 9 in Végh (1992). SOURCE: Table 8 in Végh (1992).
SOURCE: Table 11 in Végh (1992). SOURCE: Table 13 in Végh (1992).
–6
12
10
8
6
4
2
0
–2
–4
’62 ’63 ’64 ’65 ’66 ’67 ’68 ’69 ’70 ’71 ’72
–4
10
8
6
4
2
0
–2
’67 ’68 ’69 ’70 ’71 ’72 ’73
–10
20
15
10
5
0
–5
’84:1 ’85:1 ’86:1 ’87:1
–20
20
15
10
5
0
–5
–10
–15
’77:1 ’78:1 ’79:1 ’80:1 ’81:1 ’82:1 ’83:1
–15
–10
–5
0
5
10
15
20
’84:1 ’85:1 ’86:1 ’87:1 ’88:1
–20
20
15
10
5
0
–5
–10
–15
’78:1 ’79:1 ’80:1 ’81:1 ’82:1
–15
25
20
15
10
5
0
–5
–10
’84:1 ’85:1 ’86:1 ’87:1 ’88:1 ’89:1 ’90:1
–10
20
15
10
5
0
–5
’62 ’63 ’64 ’65 ’66 ’67 ’68 ’69 ’70 –4
8
6
4
2
0
–2
’87:1 ’88:1 ’89:1 ’90:1 ’91:1
R e a l p r i v a t e c o n s u m p t i o n *
R e a l p r i v a t e c o n s u m p t i o n *
R e a l p r i v a t e c o n s u m
p t i o n *
R e a l t o t a l c o n s u m p t i o n *
R e a l G D P *
R e a l p r i v a t e c o n s u m p t i o n *
R e a l t o t a l c o n s u m p t i o n *
R e a l p r i v a t e c o n s u m
p t i o n *
R e a l p r i v a t e c o n s u m p t i o n *
R e a l p r i v a t e c o n s u m p t i o n *
* Growth in percent with respect to the same per iod of the previous year.
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FEDERAL RESERVE BANK OF DALLAS 25 ECONOMIC REVIEW FOURTH QUARTER 1997
volatility of consum ption relative to output is
substantially low er in the national accounts esti-
m ate at 1986 prices (released at the end of 1996)
than in the previous estim ates at 1970 prices.
The exam ple above em phasizes that in
dealing w ith countries such as A rgentina,
researchers should heed the usual w arning to
appropriately w eigh the quality of the data
before taking for puzzles anom alies that inreality m ay be m ere statistical artifacts. For that
reason, w e report the business-cycle regularities
obtained from using tw o different estim ates of
G D P and its com ponents. The com parison of
the results from each data set w ill eventually
give som e idea of the confidence one should
place on the business-cycle regularities of Argen-
tina reported here or elsew here (for exam ples,
see K aufm an and Sturzenegger 1996 and C arrera,
Féliz, and Panigo 1996).
O ne estim ate (the “old”estim ate), in con-
stant prices of 1970, covers the 1970:1–90:4
period and w as prepared by the Central Bank ofArgentina. W e obtained this estim ate from the
FIEL (Fundación de Investigaciones Económ icas
Latinoam ericanas) data bank. The other esti-
m ate (the “new ”estim ate), in constant prices of
1986, covers the 1980:1–95:4 period. The figures
for this estim ate w ere taken from the publica-
tion Oferta y Demanda Global es, 1980 –1995 ,
prepared by the D irección N acional de Cuentas
N acionales. N otice that these tw o estim ates
overlap only during the 1980:1–90:4 period.13
Methodology. W e characterize the busi-
ness-cycle regularities of A rgen tina u sing
K ydland and Prescott (1990) as a guide. Their
procedure is inspired by Lucas (1977), w ho
defines the business-cycle com ponent of a vari-
able as its deviation from trend. K ydland and
Prescott define the trend of a variable as that
w hich results from applying the H odrick–
Prescott filter (H P filter) to the raw data.
Inform ally, this filter produces trends that are
“close to the one that students of business
cycles and grow th w ould draw through a tim e
plot”(K ydland and Prescott 1990).14 Application
of the H P filter to Argentinean G D P, for ex-
am ple, produces the trend represented by thesm oother curves in Figure 2.15,16
Except for net exports, all variables in the
tables of this article are expressed in natural
logarithm s, as is standard in the business-cycle
literature.17 Since it is not possible to com pute
the logarithm of negative values, variables that
can take on such negative values, such as net
exports, w ere exp ressed instead as ratios to
G D P. All the variables w ere seasonally adjusted
using the X -11 procedure.
The tables rep ort statistics that m easure (1)the direction of the m ovem ents of a variable
com pared w ith that of real G D P (procyclical , in
the sam e direction;countercyclical , in the oppo-
site direction; acyclical , w hen there is no clear
pattern); (2) the degree to w hich the variable
follow s the m ovem ents of real G D P (contem po-
raneous correlation); (3) the am plitude of fluc-
tuations (volatility or relative volatility); and (4)
the phase shift—that is, w hether a variable
changes before or after real G D P does (leads or
lags the cycle, respectively.)
The statisticsvolatility
corresponds to the
standard deviation of the percentage by w hich
the cyclical com ponent of a variable deviates
from trend. The statistics r elati ve volatili ty is the
ratio betw een the volatility of the variable of ref-
erence and the volatility of real G D P.
Real f acts f or Argentina Output and its components: GDP. Table 1
rep orts statistics for real G D P and its m ajor
com ponents. The first striking feature of the
table is the high vo latility of real G D P.
According to the new national account esti-
m ates, the percentage standard deviation fromtrend of Argentina’s real G D P is roughly 2.5
tim es larger than for the U nited States. Real G D P
volatility is also high in the old national account
estim ates, but w ithin the range observed in
European countries such as G reece (2.85),
Portugal (3.05), and Luxem bourg (3.2).18,19
Total consumpti on. An im portant caveat in
interpreting the consum ption evidence is that in
Argentina’s national account, consum ption is
com puted as a residual, w hich casts consider-
Figure 2
Real GDP, Old and New EstimatesThousands of 1986 pesos (log scale)*
8.9
9.5
9.4
9.3
9.2
9.1
9
’94:1’92:1’90:1’88:1’86:1’84:1’82:1’80:1’78:1’76:1’74:1’72:1’70:1
Trend real GDP new
Real GDP new
Trend real GDP old
Real GDP old
* For visual effect, the old estimates have been rescaled so that
their level is the same as for the new estimates in 1980:1.
SOURCES: Dirección Nacional de Cuentas Nacionales for new
estimates; FIEL for old ones; authors’ calculations for
trends.
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able doubt on the nature of the anom alous
behavior of consum ption that w e discuss below .
The volatility of real G D P and the relative
one for consum ption im ply that the volatility of
this real G D P com ponent is higher than that for
the U nited States or European countries. But
this anom aly is not all that rem arkable becauseit results directly from the reported high vola-
tility of real G D P and the fact that consum ption
and G D P are highly correlated.
Perhaps w hat is rem arkable is that the
volatility of consum ption is larger than that of
output. A lthough theoretically the opposite
should hold, this excess relative consum ption
volatility is w ithin the ranges observed in Japan
and som e European countries.20 M ore specifi-
cally, according to the new national account
estim ates in Table 1, A rgentinean consum ption
is 19 p ercent m ore volatile than G D P. This is no
uncom m on by international standards. Backus
K ehoe, and K ydland (1995) report that the cor
responding figure is 14 percent for Austria and
15 p ercent for Japan. A ccording to Christo
doulakis, D im elis, and K ollintzas (1995), it is ashigh as 46 percent for the N etherlands.21
By contrast, relative consum ption volatility
does exceed international standards for the old
national account estim ates. A consum ption
volatility 70 percent larger than that of outpu
is indeed hard to explain. Som e studies have
attributed this excess volatility to the presence
of credit constraints.22 H ow ever, there are rea
sons to be skeptical about this explanation
because in m odels w ith credit constraints, con
Table 1
Cycl ic al Behavior of Real GDP and Its M ain Components in Argentina and Other Countries
Argentina Argentina OECD, G–7,
(new national (old national and other
account estimates) account estimates) United European
1980:1– 95:4 1970:1– 90:4 States1 countries2
Real GDP volatility3 4.59 3.06 1.71 .90 to 3.20
Total consumption Procyclical Procyclical Procyclical Procyclical
Contemporaneous correlation .96 .84 .82 .1 to .83
Relative volatility 4 1.19 1.69 .73 .66 to 1.46
Phase shift Coincidental Coincidental Coincidental Coincidental5
Gross fixed investment Procyclical Procyclical Procyclical Procyclical
Contemporaneous correlation .94 .71 .90 .15 to .90
Relative volatility 4 2.90 3.44 3.15 2.30 to 5.63
Phase shift Coincidental Coincidental Coincidental Coincidental
Government consumption indicator Acyclical6 Acyclical7 Acyclical Acyclical
Contemporaneous correlation .206 .247 .05 –.23 to .27
Relative volatility 4 3.196 4.437 1.21 .36 to 1.28
Phase shift Lagging6 Lagging7 Lagging —
Net exports8 Countercyclical Countercyclical Acyclical Acyclical/countercyclical
Contemporaneous correlation –.84 –.62 –.28 –.01 to –.68
Volatility3 2.28 3.27 .45 .5 to 1.33
Phase shift Coincidental Coincidental Leading —
Imports Procyclical Procyclical Procyclical —
Contemporaneous correlation .81 .71 .71 —
Relative volatility 4 4.05 5.61 2.88 —
Phase shift Coincidental Coincidental Coincidental —
Exports Countercyclical Countercyclical Procyclical —
Contemporaneous correlation –.61 –.21 .34 —
Relative volatility 4 1.68 3.21 3.23 —
Phase shift Coincidental Coincidental Lagging —
1 Statistics are from Kydland and Prescott (1990).2 Statistics are from Backus, Kehoe, and Kydland (1995) and Christodoulakis, Dimelis, and Kollintzas (1995).3 Percent standard deviation from trend.4 Ratio of volatility of the variable and the volatility of real GDP.5 Except in France, where, according to Christodoulakis, Dimelis, and Kollintzas (1995), it leads the cycle.6 For the period 1980:1–89:4.
7 For the period 1970:1–89:4.8 Trade balance as percentage of GDP.
NOTE: Seemingly anomalous statistics are in bold type.
SOURCES: Authors’ calculations, using the sources reported in the text.
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FEDERAL RESERVE BANK OF DALLAS 27 ECONOMIC REVIEW FOURTH QUARTER 1997
sum ption is not as sm ooth as it w ould be
otherw ise, but it is still typ ically sm oother than
incom e.23
In considering the correlation betw een
output and consum ption, it is the figure for the
old national account estim ates that is norm al
and the one for the new national account esti-
m ates that is abnorm al. The correlation of 0.84
for the old national account estim ates is aboutthe sam e as the 0.83 correlation reported for
Canada—the highest correlation am ong the
countries reported in B ackus, K ehoe, and
K ydland (1995) and Christodoulakis, D im elis,
and K ollintzas (1995). This m eans that the 0.96
correlation betw een deviations from trend of
consum ption and G D P reported for the new
national account estim ates is unusually high by
international standards. It seem s to be high even
by Latin Am erican standards, as that correlation
is 0.91 for M exico (our ow n estim ates for the
1980:1–95:4 period) and 0.88 for U ruguay (for
the 1976:1–93:4 period; see K am il Saúl 1997).Theory predicts that such correlation
should be higher the m ore perm anent the
shocks are to incom e. Therefore, the high cor-
relation observed for Argentina m ight be an
indication that its business cycle is indeed dif-
ferent in the sense that shocks are m ore perm a-
nent there than in other countries. W e suspect,
how ever, that m ost business-cycle m odels,
m onetarist or real, w ill have a hard tim e
accounting for this high correlation w ithout, at
the sam e tim e, failing to accom m odate other
key regularities of the Argentinean business
cycle. N onetheless, there are reasons to be
cautious about the m agnitude of the contem po-
raneous correlation betw een detrended con-
sum ption and G D P in A rgentina. O ne reason, of
course, is that the significant discrepancy
betw een the correlations obtained w ith the tw o
national account estim ates points to the possi-
bility of im portant m easurem ent errors. This
possibility becom es even m ore apparent w hen
w e recall that consum ption in Argentina, as in
m any developing countries, is calculated as
a residual. This residual includes governm ent
consum ption —for w hich Argentina producesno separate quarterly estim ates—and, in the
case of the new national accounts estim ate,
changes in inventories, for w hich there also is
no separate estim ate.
An additional m ethodological source of
spurious correlation betw een consum ption and
output is the w ay output in Argentina is allo-
cated betw een consum ption and investm ent.
M any goods—such as autom obiles, electronics,
furniture, com puters, and telecom m unications
equipm ent—m ay be used for consum ption or
investm ent purposes. U nfortunately, Argentina
does not have the inform ation necessary to
determ ine the categories in w hich these goods
are being applied. To circum vent this problem ,
the production of m any item s is im puted to
both consum ption and investm ent according to
fixed coefficients constructed w ith inform ation
available only for the base year. For exam ple, 80percent of autom obile production is alw ays
im puted to consum ption and 20 percent to
investm ent. The sam e p rocedure is applied to
im ports and to the output of m any other indus-
tries that produce goods that can be used for
both investm ent and consum ption purposes.24
O f course, the proportions in w hich m any
goods are purchased for consum ption or invest-
m ent purposes change over the cycle. As a
result, the fixed-proportion m ethodology used
for Argentina’s national account estim ates w ill
distort the true underlying features of the busi-
ness cycles. In particular, w ith this im putationm ethod, part of the investm ent boom s w ill
show up m isleadingly in the data as consum p-
tion boom s.25 Because investm ent is highly cor-
related w ith output, the fixed coefficients
m ethod of im putation can artificially increase
the m easured correlation betw een consum ption
and G D P. This problem could be especially
serious in the new national account estim ates
that include the unusual investm ent boom of
the 1990s (Figure 3 ).
In sum m ary, there are reasons to be cau-
tious about the interpretation of the high cor-
relation betw een consum ption and output for
Figure 3
Real Gross Fixed Investment,Old and New EstimatesThousands of 1986 pesos (log scale)*
Trend real investment new
Real investment new
Trend real investment old
Real investment old
6.5
8.1
7.9
7.7
7.5
7.3
7.1
6.9
6.7
’94:1’92:1’90:1’88:1’86:1’84:1’82:1’80:1’78:1’76:1’74:1’72:1’70:1
* For visual effect, the old estimates have been rescaled so that
their level is the same as for the new estimates in 1980:1.
SOURCES: Dirección Nacional de Cuentas Nacionales for new
estimates; FIEL for old ones; authors’ calculations for
trends.
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the new national account estim ates reported in
Table 1. Better data are needed before one can
confidently establish that this unusually high
correlation is indeed an anom aly by interna-
tional standards.
Gross fixed domestic investment. The m ag-
nitude and sign of the statistics for this com -
ponent (plotted in Figure 3) are in line w ith
those observed in other countries. It is particu-
larly notew orthy that the relative volatility of
this real G D P com ponent is close to that for the
U nited States.
Government consumption. As stated, Argen-tina does not have separate quarterly national
account estim ates for governm ent consum ption.
The disorganization of public accounts in com -
bination w ith the high inflation rates that pre-
vailed during the period have m ade estim ation
of such a series very difficult.
H ow ever, the sam e high inflation that pre-
vents the construction of reliable governm ent
consum ption estim ates also suggests that fiscal
policies m ay have played an im portant role in
the A rgentinean econom y. Therefore, w e believe
it is im portant to report statistics—albeit par
tial—for an indicator that show s the govern
m ent consum ption contribution to G D P a
quarterly frequencies. Figures for treasury pay
roll paym ents are available on a m onthly basis
for the 1970–89 period, so w e choose this vari
able as a potential indicator of fiscal policy. W e
m ust em phasize, how ever, that these disburse
m ents represent only a fraction of all such pay
m ents in the Argentinean public adm inistration
The statistics in Table 1 show that the
relative volatility of our real governm ent consum ption indicator is w ell above internationa
standards. It is also acyclical, a feature tha
characterizes governm ent purchases in the
U nited States as w ell. This acyclicality seem s
to be anom alous by Latin Am erican standards
(see Talvi and V égh 1996).
Trade balance. Som e of the statistics fo
Argentinean net exports (trade balance as a per
centage of G D P) are in line w ith the interna
tional evidence: net exports are countercyclical
Table 2
Cycl ic al Behavior of Argenti nean and U.S. Labor Inputs and Product ivi ty
Argentina Argentina
(new national (old national
account estimates) account estimates) United
1980:1– 90:4 1970:1– 90:4 States1
Industrial real GDP volatility 2 5.57 5.84 4.18
Real GDP Procyclical Procyclical Procyclical
Contemporaneous correlation .95 .90 .86
Relative volatility 3 .69 .52 .36
Phase shift Coincidental Coincidental Coincidental
Total hours Procyclical Procyclical Procyclical
Contemporaneous correlation .76 .77 .86
Relative volatility 3 .89 .70 .73
Phase shift Coincidental Coincidental Coincidental
Employment Procyclical Procyclical Procyclical
Contemporaneous correlation .55 .49 .79
Relative volatility 3 .66 .56 .60
Phase shift Lagging Lagging Lagging
Hours per worker Procyclical Procyclical Procyclical
Contemporaneous correlation .70 .68 .77
Relative volatility 3 .43 .38 .20
Phase shift Coincidental Coincidental Coincidental
Productivity Procyclical Procyclical ProcyclicalContemporaneous correlation .48 .72 .71
Relative volatility 3 .66 .65 .52
Phase shift Coincidental Coincidental Coincidental
1 The statistics correspond to the 1959:3–94:4 period and were constructed by the authors using a series of value added by the
manufacturing sector and a corresponding series of employment and hours worked in that sector published by the Bureau of Labor
Statistics (BLS) until 1994. The quarterly measure of industrial value added was taken from CITIBASE and corresponds to the
“fixed-weighted gross product originating” series for manufacturing produced by the BLS (see Gullickson 1995 for details).2 Percent standard deviation from trend.3 Ratio of volatility of the variable and the volatility of real industrial GDP.
NOTE: Seemingly anomalous statistics are in bold type.
SOURCES: Authors’ calculations based on sources in the text for national account estimates and on FIEL for labor market data.
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FEDERAL RESERVE BANK OF DALLAS 29 ECONOMIC REVIEW FOURTH QUARTER 1997
as in several O ECD countries, although the
Argentinean contem poraneous correlation w ith
output is on the high end of the range. By con-
trast, the volatility of this com ponent seem s to
be abnorm ally high by international standards.
A sim ilar situation arises w ith im ports: they are
procyclical, as in the U nited States, but exhibit a
m uch higher volatility relative to output. Finally,
alm ost all of the statistics for exports are out ofline w ith those for the U nited States.
O ne caveat in analyzing the trade balance
com ponents of G D P is that Argentinean im ports
and exports are subject to considerable m eas-
urem ent errors because A rgentina used open
or hidden form s of exchange rate controls dur-
ing substantial portions of the period under
analysis. D uring these p eriods, the private
sector had incentives to understate exports and
overstate im ports in order to exploit the differ-
ential (w hich eventually becam e large) betw een
the often m ultiple official exchange rates and
the higher exchange rate usually prevailing inthe black m arket.
Labor inputs. Table 2 presents facts on
aggregate p roduction and labor input for the
old and new national account estim ates.
Because w e are trying to follow the m ethod-
ological approach in K ydland and Prescott
(1990) as closely as possible, w e w ould like to
replicate in our Table 2 all the statistics those
authors rep ort in their Table 1. U nfortunately,
lack of data has prevented us from achieving
the sam e results so far: there are no reliable
quarterly estim ates of capital input. And infor-
m ation on em ploym ent and hours w orked is
available only for the m anufacturing sector,
w hose value added represents a 25 percent
average of total G D P in the 1980–95 period.
For these reasons, w e report in Table 2 the
correlation and relative volatility of labor inputs
w ith respect to real industrial G D P, rather than
aggregate overall real G D P, used in Tables 1 and
3. W e also construct sim ilar m easures for the
U nited States. To give som e idea of how w ell
these series eventually approxim ate the rela-
tionship betw een labor inputs and real G D P for
the w hole A rgentinean econom y, w e report thecorrelation and relative volatility of aggregate
and real industrial G D P.
Another serious lim itation of the data is
that there are no reliable estim ates of average
w orker com pensation. Also, the relevant series
for labor m arkets have not been updated since
1990. Thus, these series overlap the new G D P
estim ates only during the 1980:1–90:4 period.
W ith these caveats about the data in m ind,
Table 2 suggests that total hours w orked,
em ploym ent, and hours per w orker are strongly
procyclical. The statistics for those variables aresim ilar across the different national account esti-
m ates. Except for em ploym ent, this sim ilarity
extends also to the correlations for the U nited
States for both periods.
The correlation of em ploym ent in the
industrial sector w ith real industrial G D P is
low er in Argentina than in the U nited States.
This finding is not surprising given the m uch
m ore stringent labor m arket regulations in
Argentina. Because of high firing costs, firm s
w ill postpone hiring and firing decisions. So
changes in em ploym ent w ill not trace changes
in output as closely as they w ould in the
absence of labor m arket restrictions.
Relative volatilities are rem arkably sim ilar
across the countries, although volatility tends to
be higher in Argentina for the num ber of hours
per w orker. This finding, again, likely reflects
the labor m arket restrictions: w hen confronted
w ith the high costs of firing w orkers, firm s tend
to expand or contract the labor hours of those
already em ployed, rather than hire or lay off
m ore w orkers.
Finally, it is w orth noting that productivity
in the A rgentinean industrial sector is procycli-cal (Figure 4 ), w ith correlations and relative
volatilities on the sam e order of m agnitude as
those for the U nited States.
O verall, the business-cycle features of
Argentinean labor inputs are reasonably sim ilar
to those in the U nited States.
Nominal facts for Argentina Table 3 sum m arizes the statistical proper-
ties of the business-cycle com ponent of several
Figure 4
Argentina, Productivit y Is Procycli calPercent standard deviation from trend
–15
–12
–9
–6
–3
0
3
6
9
’90:1’89:1’88:1’87:1’86:1’85:1’84:1’83:1’82:1’81:1’80:1
Deviationsindustrial
productivity
Deviationsreal industrial GDP
SOURCES: Authors' calculations using new national account
estimates from Dirección Nacional de Cuentas
Nacionales and index of total hours worked in the
manufacturing sector from FIEL.
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30
nom inal and m onetary aggregate series. This
table presents inform ation analogous to that in
Table 4 of K ydland and Prescott (1990), w ith the
necessary m odifications to incorporate som e
idiosyncracies of the Argentinean econom y.
First, w e do not report statistics for the
m onetary base. Because of the frequent and
cum bersom e changes in financial regim e thatArgentina experienced in the period under
analysis, the concept of m onetary base does not
have the m eaning it has in the U nited States or
in the O EC D and European countries w e use for
com parison in this article.26
Second, the im plem entation of different
form s of price controls during the analysis
period m ay have distorted the true business-
cycle price features. Therefore, as proxy for the
true underlying nom inal price level, w e also
report statistics for the exchange rate in the
black m arket.
The intense inflationary process tha
Argentina experienced in the 1970s and 1980s
is responsible for the unusual high volatility o
all variables in Table 3. H ow ever, to correctly
interpret this volatility and other statistics in the
table, it is im portant to stress that m onetarypolicy in A rgentina during m ost of the 1970–95
period w as not m onetary policy in the sense
that it is in the U nited States, but rather a form
of im plem enting fiscal policies financed w ith
m oney creation.27
O ne striking sim ilarity w ith internationa
evidence stands out from the table: w hether
m easured by the consum er price index or the
black m arket exchange rate, the p rice level has
been countercyclical (Figure 5 ), as it is in the
Table 3
Cycl ic al Behavior of M onetary Aggregates and Pric e Level Indic es in Argentina and Other Countri es
Argentina Argentina
(new national (old national
account estimates) account estimates) United OECD, G–7, and other
1980:1– 95:3 1970:1– 90:4 States1 European countries2
M1 Countercyclical Acyclical Procyclical Acyclical/procyclical
Contemporaneous correlation –.36 –.09 .31 –.06 to .42
Relative volatility 3 15.13 15.68 1.00 .49 to 2.93
Phase shift Lagging No clear pattern Leading Leading (when countercyclical)
M2 Countercyclical Acyclical Procyclical Acyclical/procyclicalContemporaneous correlation –.40 –.07 .46 –.034 to .39
Relative volatility 3 12.51 13.08 .88 .59 to 5.56
Phase shift Lagging No clear pattern Leading No clear pattern
M2–M1 Acyclical Acyclical Procyclical —
Contemporaneous correlation –.23 .01 .40 —
Relative volatility 3 11.42 13.76 1.12 —
Phase shift No clear pattern Leading No clear pattern —
Velocity M1 Countercyclical Countercyclical Procyclical —
Contemporaneous correlation –.46 –.26 .31 —
Relative volatility 3 3.20 4.58 1.18 —
Phase shift Leading Leading Coincidental —
Velocity M2 Countercyclical Acyclical Acyclical —
Contemporaneous correlation –.37 –.24 .24 —Relative volatility 3 5.06 7.08 1.08 —
Phase shift Lagging Lagging Lagging —
CPI Countercyclical Acyclical Countercyclical Acyclical/countercyclical
Contemporaneous correlation –.47 –.20 –.57 –.55 to –.03
Relative volatility 3 16.92 17.54 .82 .18 to 1.82
Phase shift Lagging No clear pattern Leading Leading
ER Countercyclical Countercyclical — —
Contemporaneous correlation –.61 –.49 — —
Relative volatility 3 16.04 18.29 — —
Phase shift Lagging Lagging — —
1 From Kydland and Prescott (1990).2 From Christodoulakis, Dimelis, and Kollintzas (1995).3 Ratio of volatility of the variable and the volatility of real GDP reported in Table 1.
4 Only Spain exhibits a large negative correlation (–.30).
NOTE: Seemingly anomalous statistics are in bold type.
SOURCES: Authors’ calculations, based on sources reported in the text for national accounts and on FIEL for monetary aggregates and price level indices.
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FEDERAL RESERVE BANK OF DALLAS 31 ECONOMIC REVIEW FOURTH QUARTER 1997
U nited States and in m ost European countries
(Christodoulakis, D im elis, and K ollintzas 1995).
The countercyclicality of prices for the U nited
States w as pointed out in K ydland and Prescott
(1990) at a tim e w hen econom ists com m only
held the opposite view . N ot surprisingly, this
finding created considerable debate because it
w ent against the predictions of m ost K eynesian
or m onetarist-inspired theories of businesscycles.28
For nom inal M 1, how ever, the com parison
w ith other countries is not that clear cut. The
pattern of correlation for this m onetary aggre-
gate depends in an im portant w ay on the
national account estim ates used. For the old
estim ates, M 1 is acyclical and all correlations
are sim ilar in sign and m agnitude to those
reported for the N etherlands in Christodoulakis,
D im elis, and K ollintzas (1995). By contrast,
according to the new national account esti-
m ates, M 1 is countercyclical, w hereas in the
U nited States and the European countries inChristodoulakis, D im elis, and K ollintzas (1995),
it is acyclical or procyclical.
The differences betw een the tw o national
account estim ates should serve as a note of
caution to researchers w orking w ith nom inal
m onetary aggregates for Argentina. It is possible
that som e of the regularities taken for granted in
the p ast w ere derived using the old estim ates,
but now those regularities have disappeared or
becom e less obvious w ith the new national
account estim ates.
In any case, both national account esti-
m ates suggest that the m onetary aggregate of
savings accounts and tim e deposits (M 2–M 1) is
acyclical. This is in contrast w ith the U nited
States, w here, according to K ydland and
Prescott (1990), this m onetary aggregate is pro-
cyclical and leads the cycle. But it w ould be
w rong to conclude that this evidence suggests
that credit arrangem ents could play a m ore sig-
nificant role in U .S. business cycles than in
those of A rgentina, because during m ost of the
analysis period, there w as a considerabledegree of financial repression in the latter
country. As a result, part of the credit m arket
w as channeled through the inform al financial
sector, w hose transactions by its very nature are
not cap tured by the official m onetary statistics.
Finally, velocity of all m onetary aggre-
gates, w hether using the consum er price index
(reported in Table 3) or the exchange rate
(not reported) as a deflator, is countercyclical,
w hereas K ydland and Prescott (1990) reported it
is procyclical for the U nited States.
ConclusionIs the business cycle of Argentina really
different from that of other countries? W e hope
this article show s other researchers how difficult
it is to answ er this sim ple question. O ne reason
for this difficulty is that the business-cycle fea-
tures of Argentina can change substantially from
one national account estim ate to the next. A s
w e indicate, the com m only held view that
absolute volatility of output is abnorm ally high
in Argentina is a m yth by the old national
account estim ates but a fact by the new ones.
Sim ilarly, the correlation of the cyclical
com ponent of real total consum ption w ith that
of real G D P is w ithin the range observed in
other countries, according to the old national
account estim ates, but unusually high by the
new ones. W e have given reasons, how ever, to
consider this last feature as partly a figm ent of
the data.
The statistics related to production inputs
(labor and investm ent), w hich play a crucial
role in R BC m odels, display rem arkable sim i-
larities w ith the international evidence. In par-
ticular, except for absolute volatilities, all the
statistics for investm ent, labor inputs, and pro-ductivity are w ithin the range observed in the
U nited States or European countries.
Based on these statistics, the only chal-
lenge for an RBC m odel of A rgentina w ould be
to explain the larger volatility of output. But a
study by M endoza (1995) suggests that an RBC
m odel could accom plish that if properly
adapted to deal w ith the idiosyncracies of the
Argentinean econom ic environm ent. By that, w e
do not m ean a m odel that incorporates only
Figure 5
Argentina, Pr ices Are Countercyclic alPercent standard deviation from trend
–200
–150
–100
–50
0
50
100
150
200
’94:1’92:1’90:1’88:1’86:1’84:1’82:1’80:1 –12
–10
–8
–6
–4
–20
2
4
6
8
10
12Price index deviations
Real GDPdeviations
SOURCES: Authors' calculations using new national account
estimates from Dirección Nacional de Cuentas
Nacionales and the consumer price index from
Instituto Nacional de Estadísticas y Censos (INDEC)
as reported by FIEL.
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technology shocks, but one that uses other real
factors or econom ic policies w hose effects can
be captured through the aggregate p roduction
function of the econom y. M ore specifically,
M endoza’s study adds term s-of-trade shocks to
an RBC m odel w ith technology shocks and
show s that such a m odel can replicate about the
sam e proportion of G D P variability—50 percent
for G –7 and developing countries—even if theabsolute volatility of G D P is substantially larger
in the developing countries. Interestingly,
according to the M endoza study, the variability
of Argentina’s term s of trade is tw ice that for the
U nited States, w hich is the order of m agnitude
by w hich the variability of Argentina’s G D P
exceeds that of U .S. G D P (using the new
national account estim ates).29
A host of other em pirical studies confirm
the potential of RBC m odels to m im ic a large
fraction of the econom ic fluctuations observed
in Latin Am erican countries. For exam ple, using
a structural vector autoregression m odel (VAR),H offm aister and Roldós (1997) find that supply
shocks are, even in the short run, the m ain
source of the output fluctuations in these
countries. Sturzenegger (1989) also reports VA R
estim ates, according to w hich supply shocks
account for 90 percent of the A rgentinean out-
put fluctuations.
The results in Table 3 are unfavorable to
the hypothesis that nom inal factors play the
m ost im portant role in econom ic fluctuations. In
particular, the price level is countercyclical.
M onetary theories of business cycles have had a
hard tim e accom m odating this em pirical regu-
larity w ithin an em pirically successful (by som e
m easure) dynam ic stochastic general equilib-
rium m odel. Furtherm ore, the A rgentinean
m onetary aggregates display, in general, a very
different cyclical (countercyclical) pattern than
those of the U nited States and Europe (pro-
cyclical). Yet, these differences do not seem to
translate to the relative volatilities and other fea-
tures of real variables, w hich behave m ore sim i-
larly in Argentina and these other countries.30
In addition, our analysis of the business-
cycle debate in Latin Am erica suggests thatnom inal exchange rate shocks, even during
ERBS program s, do not seem to have had the
clear real effects the literature has alleged. In
fact, the evidence w e have presented—circum -
stantial as it m ay be—and the few available
studies that have attem pted to analyze it in a
m ore system atic w ay all point in the sam e direc-
tion: nom inal factors do not seem to be able
to account for any significant fraction of the
business cycles of Latin Am erican countries in
general, and of Argentina in particular. Perhaps
for this reason it is tim e to give real factors thei
fair chance to do the job. Therefore, it is essen
tial that a research agenda first specify the
em pirical regularities that real factors m us
account for.
To that end, w e have presented the facts
about the A rgentinean business cycle, follow ing
a w ell-defined, system atic ap proach that doesnot im pose on the data any strong a prior
belief on a particular theory of business cycles
W e hope that our atheoretical description o
em pirical regularities w ill m otivate further em
pirical and theoretical w ork that w ill ultim ately
lead to a better understanding of the econom ic
fluctuations and of the real effects of inflation
stabilization program s in Latin Am erican coun
tries in general, and in A rgentina in particular.
NotesThe authors are grateful to D avid G ould, C arlos Vég h,
and M ark W ynne for substantive and useful com -
m ents. W e are also thankful to A nne C oursey, w hose
editorial suggestions contributed to a clearer exp osi-
tion of our ideas.1 This distaste for econom ic fluctuations is im plied by
the assum ption that econom ic ag ents have concave
preferences. A n old joke illustrates the m eaning of this
econom ic jargon. A n econom ist is inform ed that a
fellow citizen, w ith one leg freezing in ice and the
other boiling in hot w ater, is in pain. “W hy?”the
econom ist asks. “O n average, he is O K .”A ctually,
this joke doesn’t do justice to the econom ics profes-
sion, w hose m em bers know very w ell that the citizen
has concave preferences: he w ould p refer to have
both feet in lukew arm w ater. Likew ise, econom ists
know that consum ers w ould prefer an econom y in
w hich output and consum ption g row at the sam e
steady rate, quarter after quarter, to one w hose grow th
is the sam e on average but varies from high (a hot
econom y) in som e quarters to slow (a cold econom y)
in others.2 So m uch so that a p rom inent m onetarist like Lucas
him self recently asserted, “M onetary shocks just aren’t
that im portant. That’s the view I’ve been driven to….
There’s no question, that’s a retreat in m y view s.”
(The N ew Yorker , D ecem ber 1996, 55.)3 For an excellent sum m ary, see Vég h (1992).4 For details, see K iguel and Liviatan (1992), Vég h
(1992), C alvo and Vég h (1993), and citations therein.5 The vertical line is d raw n on the tick corresponding
to the p eriod in w hich the p rog ram w as announced,
unless the announcem ent w as m ad e in the first third
of the period. In this case, the vertical line is draw n on
the tick corresponding to the im m ed iately preced ing
period. The im plicit assum ption is that the real effects
of ER B S p rog ram s did not have tim e to show up in the
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FEDERAL RESERVE BANK OF DALLAS 33 ECONOMIC REVIEW FOURTH QUARTER 1997
period of the announcem ent if it cam e too late in the
period .6 This p rediction arises from the intertem poral substitu-
tion effect originally em phasized by C alvo (1986): the
tem porary (by assum ption) red uction of the devalua-
tion rate translates into a tem porary reduction in the
nom inal interest rate that increases the dem and of cur-
rent trad able g oods relative to future tradab le g oods.
The em pirical relevance of this m echanism , how ever,has b een questioned by R einhart and Végh (1995a).
7 “W itty”analysis of plots is a valid and w idely used
m ethod of analyzing econom ic evidence, esp ecially in
the early stag es of a theoretical develop m ent. H ow -
ever, this casual em piricism presents serious prob lem s
(see E asterly 1996). To avoid am biguities and im pre-
cisions, plot analysis should b e com plem ented w ith
m ore form al quantitative m ethods w henever possible.
In our case, it w ould be im portant to construct m eas-
ures estab lishing w hether the consum ption grow th
rate im m ed iately after the announcem ent of ER B S
prog ram s w as significantly d ifferent (by som e criteria)
than im m ed iately before. The ER B S literature has yet
to p rovide such a m easure. The few form al quantitative
studies in that literature that have attem pted to go
beyond the plot analysis (R einhart and V ég h 1994,
1995b, and H offm aister and Végh 1996) are con-
cerned, instead, w ith the dynam ics of real variab les
w ithin different inflation stab ilization program s.8 It is true that nom inal factors d eliver im portant real
effects in the nom inal w age rigidity version of the
m onetarist-inspired m od els exam ined by R ebelo
and V ég h (1995). H ow ever, that success is achieved
at the expense of generating countercyclical real
w ag es, w hich goes ag ainst the availab le evidence.
For exam ple, C arrera, Féliz, and Panigo (1996) rep ort
that real w ages in A rgentina and B razil are procyclical.9 In fact, none of the stabilization program s reported in
the literature has been a “pure”m onetary experim ent.
They w ere alw ays associated w ith other policy m eas-
ures, such as financial liberalization, changes in taxes
and tariffs, and so on, all factors that w ould fall in
the category of “real”in the analytical fram ew ork of
real-business-cycle theory. The om ission of these
factors from the analysis m ay lead to serious m isinter-
pretations of the evidence on stab ilization prog ram s.
For exam ple, as pointed out by C alvo (1986), “…if
expected to be tem porary, a banking liberalizationpolicy w ill tend to have effects sim ilar to the typ e of
exchange rate policies analyzed above [in reference
to E R B S program s].”10 In this sense, w e enthusiastically agree w ith C alvo
and Végh (forthcom ing, 14) that “too little em pirical
w ork—relative to theoretical w ork—has been done
in the area.”11 This m ethodology is “theory free”in the sense that it
does not take any stand w ith resp ect to the causes of
econom ic fluctuations.
12 H eston (1994) provides a very thorough d iscussion of
all the m easurem ent prob lem s typical of the national
accounts of develop ing countries like A rgentina.13 The chang e in the base year is not the only difference
betw een the tw o series. There w ere also im portant
m ethodolog ical m od ifications and other ad justm ents in
the new estim ates. The m ag nitud e of the corrections
should be ap parent from the fact that the level of
annual real G D P for 1980 is 36 percent higher in thenew estim ates than in the old estim ates. Jum ps of this
size in the level of G D P betw een subseq uent national
account estim ates are not unusual in Europ ean coun-
tries as w ell (see M ad dison 1995, 124).14 A technical presentation of the H P filter can b e found
in H od rick and Prescott (1997).15 B ecause w e are d ealing w ith q uarterly data, w e follow
K ydland and Prescott (1990) in setting the “sm oothing
param eter”λ = 1600.16 W e acknow led ge that the statistical prop erties of the
detrended com ponents m easured w ith the H P filter
rem ain som ew hat controversial (see, for exam ple, King
and R eb elo 1993). B ut it is im portant to keep in m ind
that our m ain goal is to com pare the business-cycle
regularities of A rgentina w ith those of the U nited States
and Europe. Several recent stud ies for such countries
have ind eed detrended the data w ith the H P filter as
w ell. M oreover, no d etrending technique is free from
criticism .17 The reason for this transform ation of the data is that
the business-cycle literature is concerned w ith p er-
centage (rather than absolute) deviations from trend in
grow ing series.18 A s an exercise, w e extend ed the G D P series from
each national account estim ate to the entire 1970:1–95:4
period by applying to each estim ate the grow th rates
of the other during the nonoverlap ping period . The
cyclical volatility of G D P from the series constructed
this w ay is 3.9 for the new estim ates and 3.65 for the
old ones.19 See tab le A 2 in C hristod oulakis, D im elis, and
K ollintzas (1995).20 A ccording to the p erm anent incom e hypothesis, the
series for consum ption should be sm oother than that
for incom e (or G D P). H ow ever, this pred iction is valid
only for consum ption of nond urable goods, and the
series for consum ption typ ically includes d urable
goods.21 The conjecture that the excess volatility of consum p-
tion relative to that of output m ost likely reflects a m is-
m easurem ent prob lem , as hypothesized in note 20,
is reinforced by the find ing in B ackus, K ehoe, and
K yd land (1995) that consum ption volatility is indeed
low er than that of G D P in the U .K . once exp enditures
on consum ption d urab les are exclud ed from aggre-
gate consum ption.22 See, for exam ple, “O vercom ing Volatility,”Inter-A m erican
D evelopm ent B ank (1995, 191).
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34
23 Intuitively, in an econom y incapable of transferring
w ealth betw een period s, econom ic ag ents w ill use up
all they prod uce in every period —that is, consum ption
w ill be exactly equal to incom e period after period .
A lthoug h there is absolutely no credit in this econom y,
the volatility of consum ption cannot exceed that of
output (or incom e).24 H eston (1994, 43) discusses a concrete case in w hich
allocating im ports betw een consum ption and invest-m ent, w ith proced ures analog ous to the one outlined
ab ove, m ay lead to significant errors in consum ption.
The new national account estim ates used inform ation
from the N ational Econom ic C ensus of 1985 to im pute
im ports as consum ption or investm ent goods, and
data from the N ational Econom ic C ensus of 1973 for
the sam e im putation of dom estically prod uced good s.
For m ore d etails, see C EPA L/EC LA , final rep ort, 1991.
The p articular exam ple in the text about the allocation
of autom ob iles betw een consum ption and investm ent
w as p rovided in an interview w ith staff m em bers from
the S ub secretaría de Prog ram ación E conóm ica d el
M inisterio de Econom ía of A rgentina.25 This m ay have serious im plications for the prolific
literature inspired by rep orted consum ption boom s in
Latin A m erican countries: it m ay w ell be the case that
these boom s, or at least a part of them , are in reality
cap turing m ism easured investm ent boom s.26 For exam ple, in July 1982 all A rgentinean dep osits
w ere “nationalized”—that is, from that m onth on, all
deposits in financial institutions w ere considered
dep osits at the central bank. Since these dep osits are
by d efinition part of the m oney base, this b ase
becam e alm ost identical to M 2 and therefore experi-
enced an increase equal to the difference betw een
these tw o m onetary aggreg ates p revious to the reform .
A lm ost all of the resulting jum p in the m oney b ase that
m onth w as, then, an artifact of accounting proced ures
rather than the result of a chang e in m onetary policy.
For these and other details on the institutional features
of the A rgentinean financial system over the 1900 –95
period , see Zarazaga (1996).27 M onetary policy in the U nited States is closer to w hat
econom ists w ould reg ard as “pure”m onetary policy. In
particular, U .S. m onetary p olicy is carried out through
op en-m arket op erations that exchang e one form of
governm ent deb t (fiat m oney) for another (governm ent
bonds), leaving the overall level of outstand ing gov-ernm ent deb t unchanged . In A rgentina, by contrast,
the typical m onetary policy consisted of hand ing over
fiat m oney directly to the treasury, w hich used it to
finance its deficit and not to retire other form s of gov-
ernm ent deb t as in the U nited States. Thus, m onetary
policy in A rgentina has typ ically increased the overall
governm ent debt by expand ing the m oney base. It is
in this sense that A rgentina’s m onetary p olicy has
really b een a hidden form of fiscal policy.28 A bel and B ernanke (1992) provide an excellent, bal-
anced discussion of the business-cycle facts and their
consistency w ith R B C or K eynesian theories (see
especially S ections 11.2, 12.4, and 12.5).29 A recent paper by C rucini and K ahn (1996) show s
that tariffs can have a larger im pact on G D P than
generally b elieved. This is relevant in the light that
substantial im plicit or explicit changes in tariffs w ere
a usual ing red ient of the m any stab ilization prog ram s
im plem ented in A rgentina d uring the sam ple period .30 G avin and K ydland (1996) have recently rep orted a
related finding for the U nited States. They found that
real variab les in that country seem ed to have been
invariant to the changes in the cyclical behavior
ob served in the nom inal variab les after 1979. They
show ed that these ob servations can be g enerated by
a business-cycle m od el w ith im pulses to technolog y in
w hich m onetary policy affects the cyclical behavior of
nom inal variables b ut not that of real variables.
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