American Economic Association
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American Economic Association Exchange-Rate Regimes and International Trade: Evidence from the Classical Gold Standard Era Author(s): J. Ernesto López-Córdova and Christopher M. Meissner Source: The American Economic Review, Vol. 93, No. 1 (Mar., 2003), pp. 344-353 Published by: American Economic Association Stable URL: http://www.jstor.org/stable/3132179 . Accessed: 28/08/2014 14:27 Your use of the JSTOR archive indicates your acceptance of the Terms & Conditions of Use, available at . http://www.jstor.org/page/info/about/policies/terms.jsp .
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Exchange-RateRegimesand International Trade: Evidence fromthe Classical Gold Standard Era By J. ERNESTOLOPEZ-CORDOVA ANDCHRISTOPHER M. MEISSNER* The late nineteenth century experienced a profound rise in commercial integration(see Michael D. Bordo et al., 1999; Kevin H. O'Rourke and Jeffrey G. Williamson, 1999). Interestingly,this watershedin global history also witnessed a large amountof institutional change. For instance, a vast majorityof coun- triesadoptedthe gold standard after1870, anda numberof currencyunions consisting of eco- nomically significantcountriesappearedon the scene. Many observers in the late nineteenth century argued that disparate monetary re- gimes and separate national currencies were barriers to trade which stifled international commerce.
The question then arises: how did institutional arrangements such as currency unions and monetary regimes like the gold standardaffect globalization in the late nine- teenth century?
To findout we use a gravitymodel of trade, controllingfor geographic,economic, and po- litical factors, exchange-ratevolatility, mone- taryunionmembership,andcommoditymoney * L6pez-C6rdova:Inter-AmericanDevelopment Bank, INT/ITDStopW608, 1300 New YorkAvenue NW, Wash- ington, DC 20577 (e-mail: firstname.lastname@example.org); Meissner: King's College andFacultyof Economics,AustinRobinson Building, University of Cambridge, Sidgwick Avenue, CambridgeCB3 9DD, England (e-mail: chris.meissner@ econ.cam.ac.uk).The authorsthankPranabBardhan,Brad DeLong,BarryEichengreen,LarryKarp,MauriceObstfeld, ChristinaRomer, Andrew Rose, and Jeff Williamson for their comments and guidance.
Seminar participantsat Berkeley and the CliometricsConference,along with two anonymous referees, provided helpful suggestions. Andy Rose encouragedus to use a gravitymodel with this data. We thankhim for inspiringus. The Institutefor Business andEconomicResearch(IBER)atBerkeleyandtheJohnL. Simpson Fellowship from the Institute for International StudiesatBerkeleygraciouslyprovidedfinancialassistance for thisproject.We also thankRocio Aguilerafor excellent help with the data. Any errorsare our own. The opinions expressedhereinarethose of the authorsanddo not neces- sarilyreflectthe official position of the IDB or its member countries.
regime coordination.1In qualitativeterms we findstrongevidencethatcoordinationon a sim- ilarcommoditymoneyregimeis correlatedwith highertradeand some evidence that monetary unions are associated with large increases in trade. However, our results vary depending upon the exact specificationof our model. For instance,when we controlfor observablechar- acteristics and unobservable heterogeneity at the country level we find that gold standard countriestradeup to 30 percentmorewitheach otherthanwith nationsnot on gold. In Section I we highlight previous research along these lines and give a bit of historical background to theissues.
SectionIIpresentsour data.In Section III we exhibit our results.We concludeby notingthatglobaltradecouldhave been approximately20 percentlower between 1880 and 1910if no countryhaddecidedtojoin the gold standard.
I. HistoricalBackground andPreviousWork A. Previous Work Marc Flandreau(2000) was apparentlythe first to use a gravity approachon nineteenth- century trade data. He controls only for the productof totaltradefor each of the two coun- tries, distance, sharinga border,and member- ship in the Latin Monetary Union or the Scandinavian Monetary Union (the latter in 1880only). His resultssuggestthatmembership in the LatinMonetaryUnion or the Scandina- vianMonetaryUnioncouldnotexplainbilateral tradeflows in 1860, 1870, or 1880. To the best of ourknowledge,the only otherwork investi- ' Thegravityequationhasstrongtheoreticalsupport(see Alan V.
Deardorff,1998, as well as Simon J. Evenett and WolfgangKeller,1998).JamesE. Anderson(1979) also has an earliertheoreticalderivation.Ourempiricalapproachis similar to Andrew K. Rose's (2000) study of currency unions andtradein the late twentiethcentury. 344 This content downloaded from 188.8.131.52 on Thu, 28 Aug 2014 14:27:19 PM All use subject to JSTOR Terms and Conditions
VOL.93 NO. 1 L6PEZ-C6RDOVAAND MEISSNER:GLOBALTRADEAND THEGOLDSTANDARD TABLE1-MONETARY REGIMES OFTHECOUNTRIES INCLUDED IN THEBASELINE SAMPLE Year MonetaryUnion Country (MU) 1870 1875 1880 1885 1890 1895 1900 1905 1910 U.K. Australia New Zealand Canada U.S. France Belgium Switzerland Italy Denmark Norway Sweden Germany Netherlands Finland Austria Russia Spain Portugal Japan Brazil Mexico Chile Argentina Egypt India China Indonesia Philippines Total number of countries Total number of country-pairs Sterlingunion Sterlingunion Sterlingunion Sterling,U.S./Canada U.S./Canada LatinMU LatinMU LatinMU LatinMU ScandinavianMU ScandinavianMU ScandinavianMU Gold Gold Gold Gold Paper Bimetal Bimetal Paper Silver Silver Silver Silver Silver Silver Paper Paper Bimetal Gold Silver Paper Silver Silver Silver Gold Gold Gold Gold Paper Bimetal Bimetal Paper Gold Gold Gold Gold Gold Silver Gold Gold Gold Gold Gold Gold Gold Paper Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Paper Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Gold Paper Paper Gold Silver Paper Silver Paper Silver Silver Gold Gold Gold Gold Gold Gold Gold Paper Gold Gold Gold Gold Gold Silver Gold Gold Gold Gold Gold Gold Gold Gold Paper Gold Gold Gold Gold Gold Paper Gold Paper Paper Gold Paper Silver Paper Paper Gold Gold Silver Silver Silver Gold Gold Gold Gold Gold Gold Gold Gold Paper Gold Gold Gold Gold Gold Gold Paper Gold Paper Gold Paper Gold Gold Silver 23 14 14 14 22 14 28 23 90 56 59 70 139 81 274 189 Note " indicatesthatthe countrywas not includedin the sampleduringa given year.
Source: Based on Meissner(2002).
gating the effect of currency unions on trade in the 1800's is work by Flandreau and Mathilde Maurel (2001). This work is based on a limited Europeansample, and finds that monetary unions in Scandinavia and in Aus- tria-Hungary may have increased interna- tional trade twofold. Preliminaryunpublishedworkby AntoniEs- tevadeordalet al. (2001) finds that the gold standardis significantly associated with bilat- eral tradein 1913 and in the inter-warperiod. We believe our work is the first to use both currency unions and commodity regime data (i.e., not just gold standardadherence) in the same study.
Furthermore,no study to date has similar country coverage in the data as we have, and none have taken advantage of the time-series evidence or accounted for com- modity money regime coordinationin general as we do.
B. CommodityMoneyRegimes, Currency Unions,and Trade:A Brief History Table 1 presents the countries in our sam- ple and their monetary regime at any one time. Notably by 1905 most nations were de jure if not de facto gold standardcountries. This change was precipitatedby discussions focusing on the transaction-costsaving and Gold Gold Gold Gold Gold Gold Gold Gold Paper Gold Gold Gold Gold Gold Gold Paper Gold Gold Gold Paper Gold Gold Silver 23 182 345 This content downloaded from 184.108.40.206 on Thu, 28 Aug 2014 14:27:19 PM All use subject to JSTOR Terms and Conditions
THEAMERICAN ECONOMICREVIEW 50 40 / 0t / 20 _ - ._187- --1875 1 1 ----- 1870 1875 1880 1885 1890 1895 1900 1905 1910 Year FIGURE 1.
REGIME COORDINATION, 1870-1910 trade-creating benefitsof regime coordination.2 Figure 1 presents the percentage of country- pairsin our sample sharinga similarmonetary regime in the nineteenthcentury.Much of this increaseis accountedfor by fiatregimes in the peripherybecoming gold standardregimes,but the more developed countriesof the time also witnessed a rise in coordination.Importantly, the lags in the adoptionof the gold standard amongcountries,leading to varyingconfigura- tions of coordinationacross time and space, give us uniqueevidence on the impactof mon- etaryregimes on internationaltrade.
During the period we study, a number of principalcountriesof the world participatedin some form of a monetary union (see Table 1, column 2). If monetaryunions were formed becauseof existingtradepatternsoureconomet- ricresultsmaybe exposedto endogeneityprob- lems, but their establishmentmay have been drivenby otherfactors.
Historyis notdecisive ontheissue. American republicsconcludedthatthe only benefitsfrom a hypothetical American Monetary Union would accrueto tourists(GuillermoSubercase- aux, 1915).In ScandinaviaandwesternEurope, unions were formed in part to cooperatively coercenationsintocoiningcurrencyof a similar weight andfineness as theirneighborsandpar- 2 For modem analyses see Bordo and Anna Schwartz (1996), Eichengreen and Flandreau (1996), Flandreau (1996), andMeissner(2002). Forcontemporary debatesand positions, see, for example:United StatesMonetaryCom- mission (1879, p. 331); Henry BenajahRussell (1898, p.
100);CountMatsukata Masayoshi(1899, p. 191);andCom- mission on International Exchange(1904, pp. 94, 120). tially to "remedy the inconvenience to trade between their respective countries resulting from the diversity of their small silver coin" [LucaEinaudi,2001; see also IngridHenriksen and Niels Kaergard (1995) and Krim Talia (2001)]. Australiaand New Zealand used the poundsterling,which apparently hadto do with close colonial relationshipswith England(S. J. Butlin, 1986).
II. Data Ourbaseline regressionsuse an unbalanced panelconsistingof 1,140 country-pair observa- tions.3The datacover the period 1870 to 1910 at five-yearintervals.On averagewe have four observationsover time for each country-pair. The numberof countriesthatgenerateourpair- wise observationsis largertowardthe end of our sampleperiodas Table 1 illustrates. We complementeda dataset puttogetherby Katherine Barbieri (1996) with information from national statistical yearbooks and other publicationsfromtheperiod;a detaileddescrip- tionof oursourcesappearsinourworkingpaper (Lopez-Cordova and Meissner, 2000).
Trade figureswere transformed into 1990 U.S. dollars using a U.S. consumerprice index and annual averageexchangerates.
Informationon every country'smonetaryre- gime was used to createdummyvariablesindi- cating whether any pair of countries shareda common monetarystandardor a common cur- rency.InTable 1we reporttheregimesforeach country that enters our baseline regression basedon datafromMeissner(2002). Inourdata set we have just over 100 observations (i.e., roughly10percentof oursample)wherebothof the tradingpartnersare in a monetaryunion. We constructedour measure of exchange- rate volatility as the standarddeviation of the first difference of the naturallogarithmof the monthlybilateralexchangeratefortheprevious three years.
We also control for the effect of tradepolicy on bilateralexchangeby insertinga dummy if two countriesshareda tradeagree- mentwith a most-favorednation(MFN)clause. The standard distancevariable(thelogarithmof 3More informationabout the data and the sources are availablein the full workingpaper.
346 MARCH2003 (L) This content downloaded from 220.127.116.11 on Thu, 28 Aug 2014 14:27:19 PM All use subject to JSTOR Terms and Conditions
VOL 93 NO. 1 LOPEZ-C6RDOVA AND MEISSNER:GLOBALTRADEAND THEGOLDSTANDARD TABLE2-SAMPLE AVERAGES, POOLED ANDANNUALOLS REGRESSIONS Sample averages OLS regressions Variable (1) (2) (3) (4) (5) (6) (7) (8) (9) (10) (11) (12) (13) (14) Not on the On the Not in a In a Baseline 1870 1875 1880 1885 1890 1895 1900 1905 1910 gold gold monetary monetary standard standard union union Volatility 1.94 0.42 1.24 0 0.167 0.396 -1.102 1.995 -0.899 0.209 0.373 0.163 0.073 0.282 (1.79) (0.54) (1.51) - (0.044) (0.190) (0.246) (0.917) (0.971) (0.364) (0.340) (0.070) (0.172) (0.082) GDP 20.75 20.57 20.75 19.82 0.861 0.817 0.780 1.047 0.906 0.736 1.064 0.822 0.991 0.886 (1.55) (1.79) (1.64) (1.9) (0.028) (0.095) (0.097) (0.097) (0.091) (0.089) (0.112) (0.055) (0.076) (0.052) GDP per capita 15.05 15.84 15.4 15.92 0.656 1.617 1.519 1.120 1.268 0.991 0.559 0.825 0.298 0.291 (0.691) (0.7) (0.8) (0.63) (0.081) (0.280) (0.325) (0.404) (0.306) (0.412) (0.533) (0.157) (0.172) (0.140) Distance 7.93 7.6 7.82 7.14 -0.661 -0.349 -0.724 -0.977 -0.888 -0.912 -0.755 -0.607 -0.672 -0.520 (1.04) (1.25) (1.12) (1.4) (0.045) (0.210) (0.165) (0.156) (0.140) (0.141) (0.196) (0.099) (0.126) (0.090) Border 0.102 0.12 0.082 0.36 0.625 1.506 0.931 0.195 0.184 0.529 -0.023 0.645 0.503 0.697 (0.303) (0.32) (0.27) (0.48) (0.122) (0.366) (0.379) (0.377) (0.374) (0.371) (0.417) (0.275) (0.405) (0.274) Political union 0.02 0.09 0.02 0.45 0.927 0.143 0.401 0.970 0.439 0.514 0.917 1.088 0.816 0.465 (0.14) (0.3) (0.14) (0.45) (0.293) (0.615) (0.788) (0.778) (0.862) (1.026) (0.714) (0.709) (0.695) (0.907) Common 0.06 0.18 0.07 0.64 0.165 0.611 0.545 -0.076 0.046 0.799 -0.488 0.228 -0.214 0.078 language (0.24) (0.38) (0.25) (0.48) (0.167) (0.411) (0.506) (0.354) (0.303) (0.585) (0.350) (0.301) (0.555) (0.236) MFN 0.45 0.6 0.56 0.28 0.142 0.187 -0.346 0.030 0.234 0.370 -0.028 -0.102 0.139 0.287 (0.5) (0.49) (0.5) (0.45) (0.095) (0.344) (0.287) (0.308) (0.252) (0.306) (0.350) (0.205) (0.246) (0.190) Constant -18.438 -34.649 -28.406 -29.905 -26.777 -18.849 -22.036 -21.114 -15.682 -15.030 (1.392) (5.121) (5.335) (6.502) (5.044) (6.281) (7.781) (3.134) (3.105) (2.456) Silver 0.765 1.479 - -0.307 - (0.394) (0.410) (1.230) Bimetal -0.303 -0.366 -0.987 - (0.269) (0.282) (0.529) Gold 0.479 1.583 0.894 2.603 0.191 -0.465 1.993 0.449 0.161 0.662 (0.124) (0.490) (0.320) (0.733) (0.681) (0.371) (0.460) (0.256) (0.306) (0.250) Monetaryunion 0.716 0.129 -0.138 2.558 0.737 0.363 1.448 0.380 0.933 0.778 (0.186) (0.404) (0.662) (0.817) (0.784) (0.646) (0.477) (0.459) (0.575) (0.408) Numberof observations 518 622 1,022 118 1,140 90 56 59 70 139 81 274 189 182 R2 0.595 0.673 0.852 0.835 0.794 0.486 0.753 0.567 0.568 0.650 Root MSE 1.453 1.134 0.877 0.963 1.172 1.755 1.310 1.550 1.643 1.215 Notes: Dependentvariableis In(trade).Standarderrorsare reportedin parentheses.Forregressionsthese areheteroskedasticityrobust.
great circle distance between capitals) is the literature'sproxy for transportation costs and was takenfromRose (2000). We includedcom- mon language, common border,and year-spe- cific indicators. We also created a "political union" dummy encompassing a colonial rela- tionship-colony-colonizer and colonies with the same colonizer-as well as countriesthat formed a single political entity. Table2, columns(1)-(4), shows thatthe lev- els of the regressorsare reasonablysimilarfor countries on and off the gold standard.After adoptingthe gold standardor a currencyunion, nations had slightly higher levels of GDP per person than those that did not.
Additionally, these countries were more likely to be geo- graphically closer, as indicated by the mean distance coefficient and the common border variable. Countriesin a monetaryunion were also morelikely to havea politicalunionandto sharea common language.4 4 Formal statistical tests reject the equality of means between both groups for all variables except borderand the productof GDP for gold and non-gold countries.The issue raises Torsten Persson's (2001) critique on Rose. Namely, our findings may be spuriousbecause of selec- tion on observables or nonlinear associations. By inter- acting our institutional variables with other explanatory variables, we found no evidence that nonlinearitiescan explain our findingson the gold standardcoefficient while statisticalsignificanceof the currencyunion estimatesde- cline when we include such effects.
The results of those 347 This content downloaded from 18.104.22.168 on Thu, 28 Aug 2014 14:27:19 PM All use subject to JSTOR Terms and Conditions
THEAMERICAN ECONOMICREVIEW III. BaselineResults In column (5) of Table 2 we reportpooled ordinaryleast-squares(OLS) estimates of the regressionof the naturallogarithmof realtrade levels on our set of covariates.Throughoutwe present White heteroskedasticity-robuststan- darderrors.This specificationexplains nearly 60 percent of the variation in bilateral trade flows. Our estimates show that monetaryre- gimes may have hada nonnegligibleimpacton international trade.Otherexplanatoryvariables seem in line with our predictionsalthoughin some instances they are statisticallyinsignifi- cant. Annual cross-section regression results also appearin Table 2.
The limited size of our annualsamplesfor some yearsresultedin poor regressionresultsin termsof statisticalsignifi- cance on ourmonetaryvariables,but, in quali- tativeterms,annualpointestimatessupportour conclusionsfrom the pooled regression. A. MonetaryVariables Ourbaselineregressionrejectsthehypothesis that regime coordinationhad an insignificant associationwithbilateraltradeflows. The coef- ficienton "gold,""silver," and"monetary union" are positive and statistically significant. Our baselineresultsshow thattwo countrieson the gold standardtraded62 percentmore with one anotherthan with countries under a different monetaryregime.
Tradebetween countrieson silver may have received an even bigger boost from the common monetaryregime of approx- imately 115percent.But the numberof pairsin which "silver"is equalto 1 is small, andthese observationstend to appearat early stages of our period of analysis. Bimetallism does not seem to be a significantforce encouragingbi- lateraltradeflows either because of the small numberof observationsor becauseof its inher- ent instability.
Countriesin amonetaryunionappearto trade morethantwo times morewith each otherthan they would with countries outside the union. Furthermore, the associationbetweentradeand a monetaryunionis likely understated by look- longer specifications are available in our longer working paperversion of this paper. ing atthecoefficientson thatvariable.Joininga monetaryunioneffectively impliedbeingon the samecommodityregimestandard. Forexample, in ourbaselinesample96 outof 118pairswhich sharea currencyarealsoon thegold standard. It is reasonableto assertthatbilateraltradewould be about3.30 times largerwhen both countries belonged to a monetaryunion.5 Nominal exchange-rate volatility is posi- tively associatedwith international trade,with a statisticallysignificantcoefficientof 0.17.
This findingcontradictsour expectationsof a nega- tive effect and is in contrast to Jeffrey A. Frankeland Shang-JinWei (1998) and Rose's (2000) findings. In our defense, Philippe Bac- chetta and Eric van Wincoop (2000) show that the theoretical effect of exchange-rate volatility on commerce is ambiguous, histor- ical actors seem not to have paid too much attention to such oscillations and most mod- ern researcherslike Maurice Obstfeld (1997) and CharlesWyplosz (1997) have all but dis- counted the intuitive negative effect of vola- tility on trade.6 B. GravityEquationand ControlVariables The estimatedcoefficient on the productof thecountry-pair's GDPis 0.86.
A literalreading of our estimate suggests that trade openness duringthe nineteenthcenturywas affectedto a lesserextentthantodayby the size of a country and thatcommercialintegrationhad reacheda level at least as high as today's level. Ouresti- matefor theproductof GDPpercapita,0.66, is identicalto thatin FrankelandRose (2002). In our regression,a 1-percentincreasein the dis- tance between two countries reduces bilateral 5 We consideredCanadato be in a monetaryunionwith the United Kingdomand some of the Britishcolonies and dominions, as well as with the United States. This is be- causebothBritishsovereignsandthe U.S.
dollarwere legal tender in Canada. Our econometric results are identical whetherornotwe considerCanadaandthe UnitedStatesas having a currencyunion.
6 In the longerworkingpaperwe devote an appendixto this finding.Unexpectedlylargetradein BrazilandChile- despite high exchange-ratevolatility in both countries- may explain the positive coefficient. We also found a quadraticrelationshipbetweenexchange-ratevolatilityand trade.This suggests thatat high levels of volatility(e.g., in the aftermath of crisesand/orrapiddepreciations),tradecan be spurredwhile in normaltimes volatilityreducestrade. 348 MARCH2003 This content downloaded from 22.214.171.124 on Thu, 28 Aug 2014 14:27:19 PM All use subject to JSTOR Terms and Conditions
1 L6PEZ-C6RDOVAAND MEISSNER:GLOBALTRADEAND THEGOLDSTANDARD trade by only 0.66 percent-compared to a 1-percentdecline in the late twentiethcentury, according to Frankel and Rose (2002).7 The distance coefficients may be capturingdiffer- ences in the degree of relative tradeopenness thatexisted in each period. Estimateson therestof ourexplanatoryvari- ables have the expected sign, althoughwe did not find statisticallysignificantcoefficients for the common language and MFN dummies. If product differentiation during the nineteenth centurywere limited, then culturalsimilarities captured in the common language dummy would have been a less importantdeterminant of trade,explainingthe lack of significanceof the former variable.
We attributethe statisti- cally insignificantestimateof the MFN dummy to thedearthof easily accessiblesourcesregard- ing nineteenth-century tradetreaties.Both con- tiguity and close political ties between two countries (or colonies) are highly correlated with trade.
C. OtherSpecifications The mainresultsaresurprisinggiven thatour data has overcome the theoreticalobservation thatlargecountrieswhich tradea lot shouldsee relatively small effects from droppingbilateral barriers to trade (i.e., joining a monetary union).8Neverthelesstherearea host of factors that might be influencingour results. We also present a few alternativespecificationsof our 7In our working paper, we allow for a time-varying distancecoefficientand,as Jeff Williamsonsuggested,fora more direct measure of transportation costs. We also al- lowed for a bilateralmeasureof transportation costs devel- oped by Nuno Limao and Anthony J.
Venables (1999) which takesinto accountrailroadsandtelegraphiccommu- nications.This does not affect ourqualitativeresultson the monetaryvariables.
8In a theoreticalanalysis of these issues, Rose and van Wincoop (2001) and Anderson and van Wincoop (2003) note that in situations where pre-union trade is low or currencyunionsaresmall,theeffect of joining a unionmay be largeand,whentradeamongcountriesis extensiveorthe (potential)unionis economicallylarge,the effect of joining a unionwould be small. The logic is that,if tradecosts are reduced among a group of countries that already trade substantiallywith each other,multilateraltradebarriersfall considerablywhile bilateralresistance falls only slightly. Trade increases the most when bilateral resistance falls relativelymore thanmultilateralresistance.
40 - 20 - 0D c00 5) c. 0 - -20 - 5-9 10-14 15-19 20-24 25-29 30-34 35-39 Yearsaftermovingtocoordination onthegoldstandard FIGURE 2. ESTIMATED LAGGED IMPACT ON TRADE OF A MOVE TO COORDINATION model to see if they validate the conclusions from Table 2.
We illustratethefact thatit took a numberof years to achieve the impact on trade that the baseline regressions imply by allowing for lagged effects of moves to coordinationin our baseline pooled regression.That is we add to our gravity equation indicator variables five years,tenyears,etc., aftera country-pair moved to coordinationon the gold standard.9Figure 2 shows thattradebetween countriesrises rel- ativeto noncoordinating countriesforatleast30 years, reachingan impactof nearly50 percent after about 15 years. However, the few obser- vations we have suggest that after 30 to 35 years,tradeis slightlylower comparedto coun- tries thatnever moved to coordinationby 5 to 20 percent.
Next, we consider the possibility thatunob- served country-pairor country characteristics are drivingour baseline results.Column(1) of Table 3 presents a country-pairfixed-effects specificationof the gravityequation.Here co- ordinationon the gold standardstill has a pos- itive and statistically significant association with trade.Ourestimatesays thatcoordination 9The effect we refer to in the text is the sum of the coordinationdummy(0.90 with a standarderrorof 0.14 in this specification)plus the coefficienton the move to coor- dination t years later. The coefficients and their standard errorsbeginningwith thaton five yearsafterthe move and finishing 35 years after are -0.85 (0.14), -0.78 (0.18), -0.49 (0.21), -0.49 (0.23), -0.65 (0.22), -0.96 (0.19), and -1.13 (0.34).
I I I I I 349 This content downloaded from 126.96.36.199 on Thu, 28 Aug 2014 14:27:19 PM All use subject to JSTOR Terms and Conditions
THEAMERICAN ECONOMIC REVIEW TABLE 3-HETEROGENEITY AND ENDOGENEITY REGRESSIONS (1) (2) (3) Regressors Country-pair FE CountryFE IV regression Gold 0.154 0.283 0.973 (0.077) (0.125) (1.318) Silver 0.177 1.104 (0.267) (0.396) Bimetallism 0.189 -0.308 (0.312) (0.352) Monetaryunion 0.258 1.335 1.305 (0.54) (0.389) (1.358) Volatility 0.017 0.054 0.110 (0.024) (0.034) (0.369) GDP 0.550 0.360 0.905 (0.148) (0.265) (0.091) GDP per capita 0.312 0.434 0.806 (0.092) (0.113) (0.519) Distance X cost index -0.33 - (0.121) Distance -0.487 -0.74 (0.136) (0.103) Border 0.567 0.27 (0.298) (0.389) Political union 0.155 0.906 0.215 (0.690) (0.533) (1.137) MFN -0.031 -0.160 0.067 (0.105) (0.133) (0.223) Country-pair controls yes no no Countrycontrols no yes no Constant -9.530 -7.43 -21.264 (3.290) (6.584) (6.210) Numberof observations 1140 1140 681 R2 0.502 0.735 0.670 Notes: Robuststandarderrorsarereported.Yeardummiesarenot reported.
IV regression-Variable instrumentedfor: gold and monetaryunion. Instruments:Ratio of gold reserves to domestic liabilities outstandingand common languageindicator.
on the gold standardraises tradeby 15 percent relative to the country-pairsample average.10 Thecoefficienton themonetaryunionindicator shrinksin magnitude.Itis no longerstatistically significant,possibly becausethereis little vari- ationin this regressorover time. Reassuringly, the pointestimateis still positive. Regression2 uses country-specificfixedeffects.Thiscontrols for unobserved multilateral barriers to trade specific to each country as theoreticaldiscus- sions of gravitymodels suggest (see Anderson and van Wincoop, 2003). Country-pairson a gold standardtradeabout30 percentmorewith 10In our data there are 79 moves to coordinationon commoditymoney regimeswhile thereareonly two obser- vations thatinvolve a currencyunion regime switch.
each otherwhile countriesin a currencyunion tradenearly2.8 times more thanthey might if not in a currencyunion, and both coefficients are statisticallysignificant.
Onecouldalso arguethatanendogeneitybias may be affecting our results. Countries that traded disproportionatelymay have found it more lucrativeto coordinateon the gold stan- dardor to form a currencyunion.11In orderto address this concern, we estimate our model usinginstrumental variables.We instrumentfor the gold standarddummy with the productof each country's ratio of gold reserves to bank "1We focus on the gold standardbecause most of our observationsarefor gold standardcountries. 350 MARCH2003 This content downloaded from 188.8.131.52 on Thu, 28 Aug 2014 14:27:19 PM All use subject to JSTOR Terms and Conditions
VOL.93 NO. 1 L6PEZ-C6RDOVAAND MEISSNER:GLOBALTRADEAND THEGOLDSTANDARD notes in circulation.A countrynecessarilypos- sessed gold reservesto be on the gold standard. However,it is unlikelythatthis gold coverratio would be affected by the level of integration between countries.Moreover,the ratiomay be reflecting fundamentalfinancialcapabilitiesor exogenous legal stipulations on the level of requiredreservesratherthanunaccountablede- terminantsof trade.We instrumentforthemon- etary union variablewith a common language indicator. Countries in our sample that had monetary unions often shared a similar lan- guage.
Yet we find no reason why language might be correlatedwith the errorterm espe- cially since we are controlling for so many factorsalreadyandexplainingnearly60 percent of the variationin trade,because of our earlier argumentthatproductsmay not have been too differentiatedin the nineteenth century, and since the common language dummynever en- ters our earlierspecificationwith a statistically significantcoefficient.12Using two-stage least squares,the magnitudeof both point estimates increases but neithercoefficient is statistically significant(Table 3, regression3). However, a Hausmantest cannotreject the null hypothesis of exogeneity of the regressors (X2 = 0.07, p-value 1.00).
Therefore,we findno conclusive evidence thatan endogeneitybias explains our baseline parameterestimates.
In our working paper we ran a numberof otherchecks. We allowed for the possibility of a sample selection bias using a Heckmantwo- stage estimatorand we found that the coeffi- cients on the monetary variables of interest increased in magnitude and were still highly statisticallysignificant.We also estimateda To- bit model to account for the potential sample selectionandobtainedresultssimilarto thosein the baselineregression.Moreover,we explored whether other variables we omitted from the baseline were biasing our coefficients or if the endogeneity of GDP was an issue. Last, we testedfor theinfluenceof outliersandcorrected for autocorrelation in theerrorterm.Theresults 12 The coefficientsandtheirstandarderrorsin parenthe- ses in first-stagelinearregressionspredictinggold standard coordination and monetary union adherence were 0.0000177 (0.000000193) and 0.088 (0.004) respectively; the F-statistics for zero-slopes were 83.87 and 460.77; the R-squaredvalues were 0.07 and0.04.
fromourbaseline specificationarequiterobust to potential specification problems. Virtually none of the checks we undertookin this sub- section radicallyalteredthe qualitativeconclu- sions of ourbasic model:monetaryregimesare significantlyassociatedwith highertrade. IV. Concluding Remarks Inthispaperwe findstrongevidenceconsistent with the idea thatmonetaryregimechoicehad a largeimpactonpatterns of tradeinthefirstperiod of globalization.Trade flows may have been nearly 30 percent larger when two countries adoptedthe gold standard.
Some evidence sug- gests that monetaryunions are associatedwith levels of tradenearlytwo timeshigher.Combin- ing these two effects, which was the case more often thannot, suggestsa very largeassociation betweentradeandmonetary regimecoordination.
Withtheseresultsit is also possible to gauge the contributionof the gold standardto global integrationbefore 1913. How might tradehave evolved without the rise of the classical gold standard(i.e., no regime switching having oc- curredafter1870)orwithno commoditymoney regimecoordination?13 Firstwe predictedtrade in such a counterfactualworld using our base- line pointestimates.We thencomparedit to the level of predictedtradegiven actualcommodity regime adherence.This reveals thatthe rise of the classical gold standardaccountsfor perhaps 20 percentof the rise in global tradebetween 1880 and 1910.If eachcountryhadoperatedits own fiat money throughoutthe period, trade might have been even slightly lower than this figure.All of this stronglysupportstheideathat commodity money regime coordination and currencyunions were an importantcatalystfor nineteenth-century globalization.
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