European Economic Review 56 (2012) 54–71
Contents lists available at ScienceDirect
European Economic Review
journal homepage: www.elsevier.com/locate/eer
Trade, conflict, and political integration: Explaining the
heterogeneity of regional trade agreements
Vincent Vicard
Banque de France, France
a r t i c l e in f o
abstract
Article history:
Received 7 December 2009
Accepted 29 June 2011
Available online 21 July 2011
Many historians argue that the main goal of European trade integration was the
preservation of peace. This paper investigates whether this reasoning is relevant for the
EU and other regional trade agreements (RTAs). I provide empirical evidence that customs
unions and common markets (deep RTAs) do reduce the probability of war between
members. Partial scope and free trade agreements (shallow RTAs) however have no effect
on war probabilities. Accordingly, international insecurity has a differential impact on
incentives to create RTAs. Deep RTAs are signed between countries that are involved in
many interstate disputes and that have low trade costs with the rest of the world, whereas
the opposite is true for shallow RTAs.
& 2011 Elsevier B.V. All rights reserved.
JEL classification:
D74
F15
F51
F52
Keywords:
International relations
Regionalism
Trade
War
1. Introduction
The European Union (EU) is unquestionably the most integrated regional trade agreement (RTA) in the world, and a
yardstick for other regions of the world. Many historians argue that the main goal of the European integration process was
the preservation of peace after three increasingly destructive wars in Europe in less than a century. This view is illustrated
by Robert Schuman’s proposal for the creation of the European Coal and Steel Community, the forerunner to EU: ‘‘by
pooling basic production and by instituting a new High Authority, whose decisions will bind France, Germany and other member
countries, this proposal will lead to the realization of the first concrete foundation of a European federation indispensable to the
preservation of peace’’.1 This paper investigates whether the reasoning linking security and regional trade integration is
relevant for Europe and other regions of the world by asking two questions: do RTAs prevent the outbreak of war and is
international security a motive for RTA creation?
The literature in international relations identifies two channels through which regional trade integration is likely to affect
international insecurity (Bearce, 2003).2 First, since war disrupts bilateral trade (Martin et al., 2008; Glick and Taylor, 2010), an
RTA increases the opportunity cost of war by increasing intra-regional trade (Martin et al., 2008; Polachek, 1980; Oneal and
Russett, 1999). Second, supranational institutions created in relation to regional integration promote the exchange of
information on military capabilities and resolve and patience in conflicts, through formal security/military substructures, joint
military exercises and forums of defence ministers. Moreover, regular meetings of head of states and high level officials or the
E-mail address: [email protected]
Declaration of 9 May 1950, http://europa.eu/abc/symbols/9-may/decl_en.htm.
Schiff and Winter (2003) identify a third channel related to access to raw materials.
1
2
0014-2921/$ - see front matter & 2011 Elsevier B.V. All rights reserved.
doi:10.1016/j.euroecorev.2011.06.003
V. Vicard / European Economic Review 56 (2012) 54–71
55
existence of an executive secretariat create habits of negotiation and build trust between political leaders.3 International
institutions are thus likely to reduce asymmetries of information in conflicts and to mitigate problems of credible
commitment in interstate negotiations, which reduces the probability that a dispute escalates into war (Fearon, 1995;
Grossman, 2004).4
Supranational institutional frameworks however differ greatly depending on the form of regional trade integration.
Creating a customs union requires agreement on a common external tariff and revenue distribution between member
states. A common market requires more comprehensive political institutions to agree on a broader set of issues
(harmonization of regulation and standards, free movement of goods and factorsy),5 whereas a free trade agreement
or partial scope arrangement involves little or no political or institutional integration.6 According to the political
integration criterion, two categories of RTAs can be distinguished: deep (customs unions and common markets) and
shallow (partial scope and free trade agreements) RTAs. Only deep RTAs require a significant common institutional
framework likely to promote the negotiated settlement of conflicts and support peace between members.
If countries can design regional trade agreements to pacify interstate relations, international security should affect decisions
on trade policy. In a purely economic framework in which RTAs have no effect on war probabilities, two partners that have more
issues of dispute would have fewer incentives to create an RTA since disputes may escalate into war and disrupt bilateral trade.7
Conversely, if an RTA reduces the probability of war between members, then having more disputes may increase the likelihood
of an RTA being set up. To the extent that only deep RTAs pacify relations between members, international insecurity should
have a differential effect on incentives to create RTAs (and should affect both the choice of form of trade integration and the
choice of partner). In a nutshell, a history of conflicts should enhance the creation of deep RTAs, but not shallow ones.
In this paper, I set out an empirical analysis of the relevance of international security in the creation of shallow and
deep RTAs. The motives for choosing different strategies of integration are based on the premise that a deep RTA reduces
the probability of war between members while a shallow RTA does not. I first test this proposition using data on
militarized interstate disputes from the Correlates of War (COW) project covering the 1950–1991 period. I address the
selection issue due to the heterogeneity of dispute occurrence across country pairs using a bivariate probit model
accounting for selection and event data from Kinsella and Russett (2002) to measure interstate dispute occurrence. I find a
sizeable impact of membership in a deep RTA: it reduces the probability of a dispute escalating into war by two-thirds.
Membership of a partial scope or free trade agreement has no significant effect on war probabilities.
Second, I test the determinants of the likelihood of deep and shallow RTAs on a cross-section of 2814 unique country
pairs in 2005. The endogeneity bias related to past membership of RTAs and omitted variables is addressed by controlling
for several co-determinants of regionalism and conflict and by implementing an instrumental variable strategy. I find that
deep RTAs are signed between countries that have many interstate disputes and that have low trade costs with the rest of
the world, whereas the opposite is true with respect to shallow RTAs. These empirical results provide strong support for
the differential effect of international insecurity on incentives to create deep and shallow RTAs. Besides the reduction of
tariffs, this paper explicitly emphasizes the role of RTAs as a regulating mechanism for interstate relations. By offering
empirical evidence on the choice of RTA partners as well as the form of regional integration, this paper complements Baier
and Bergstrand’s (2004) analysis of the economic determinants of RTAs.
This paper is related to the theoretical literature on the endogenous formation of RTAs that emphasizes non-traditional
gains from regional integration.8 This strand of the literature explicitly recognizes various motives for regional integration
and identifies distinct problems that a trade agreement may solve (Maggi and Rodriguez-Clare, 1998; Mitra, 2002; Limao,
2007). Regional trade integration may indeed provide non-traditional gains and help solving problems of time
inconsistency, signaling, insurance, cooperation and security (Fernandez and Portes, 1998; Whalley, 1998; Schiff and
Winters, 1998). In such a framework, RTAs are not only regarded as engines of preferential liberalization but also as
international institutions providing public goods to their members (Limao, 2007; Alesina et al., 2005). These papers
however only consider the case of free trade agreements or customs unions, or do not distinguish between RTAs according
to the form they take.9 The usual classification of RTAs, derived from Balassa (1961), considers regional trade integration as
a step-by-step process leading to economic union, through free trade area, customs union and common market. The
underlying assumption is that more integrated arrangements provide deeper trade integration.10 Vicard (2009) however
3
For instance, Manzetti (1993/94) reports that discussions of sensitive policy issues such as nuclear proliferation concerns have taken place within
the MERCOSUR institutions.
4
Jackson and Massimo (2007) also show, in a setting where countries are at war because of the political biases of their leaders that when state
leaders lack the ability to credibly commit to a negotiated deal, the scope for negotiated settlement of disputes is reduced.
5
See, for instance, Alesina and Wacziarg (1999) for a detailed mapping of policy areas carried out at the EU level, and Bouzas and Soltz (2001)
concerning the institutional framework of MERCOSUR.
6
The ASEAN free trade agreement is an illustrative example, with weak regional institutions in order to limit any supranationalism (Best, 2005).
Pomfret (1997) also emphasizes how the will to limit political integration has been fundamental to the creation of NAFTA.
7
This reasoning assumes either that negotiating or implementing an RTA involves costs or that increasing trade integration increases the number of
disputes and thus the probability of trade disruption in the future.
8
The optimal tariff strand of the literature focuses on traditional trade gains (Riezman, 1985; Yi, 1996; Ornelas, 2005).
9
From an empirical point of view, Mansfield and Pevehouse (2000) and Martin et al. (2010) also investigate the impact of RTAs on security without
differentiating between RTAs.
10
In his seminal paper, Balassa (1961) also mentions social integration, but he dismisses this second criterion.
56
V. Vicard / European Economic Review 56 (2012) 54–71
shows that different kinds of RTAs have, on average, a similar effect on intra-regional trade. Empirical evidence of gradual
regional integration processes is also lacking: out of the 18 customs unions created worldwide since 1948, 14 have been
created directly as such. This paper offers an alternative explanation for the choice of different strategies of regional
integration. The form of RTAs reflects different institutional arrangements that provide different non-traditional gains to
their members.
The paper proceeds as follows. The next section presents data on interstate dispute and war. Section 3 presents the
analysis of the impact of RTAs on war probabilities. Section 4 investigates the determinants of the creation of deep and
shallow RTAs, and Section 5 concludes.
2. Data on interstate dispute and war
The probability of war between two countries i and j at date t is the probability of occurrence of a dispute between i and
j multiplied by the conditional probability that the dispute escalates into war:
Prðwar ijt Þ ¼ Prðdisputeijt Þ Prðescalationijt 9disputeijt Þ:
ð1Þ
The probability of dispute varies by country pair and over time. Disagreements over territorial issues or religious and
ethnic minorities are more likely to occur between neighboring countries. Policies fostering international trade integration
are also likely to increase dispute occurrence. In the same manner, many factors affect the probability of a dispute
escalating into war. For instance, one of the few accepted regularities in international relations is that democracies are less
prone to war with each other because their state leaders are accountable to citizens (Oneal and Russett, 1997; Conconi
et al., 2008). Accordingly, the history of wars does not accurately reflect the extent of dispute issues at stake between two
countries. The question addressed in this paper is to understand what motivates the choice to settle disputes through
negotiation rather than through war and how international institutions affect this choice. Distinguishing the presence of
dispute issues between two countries from their most severe outcome, i.e. war, is therefore particularly relevant to this
study. In this paper, I use two different sources of data to measure the occurrence of interstate disputes and wars.
The Correlates of War (COW) project provides detailed data on Militarized Interstate Disputes (MIDs) over the period
1816–2001 (Faten et al., 2004). War is restrictively defined as a MID involving at least 1000 deaths of military personnel,
which reduces the number of events considered as war to less than 100 cases since 1815. Adopting such a restrictive
definition prevents robust empirical analysis. I follow the literature and use a broader definition of war that includes
armed conflicts involving the display or the use of military force, i.e. a MID of hostility level 3 (display of force), 4 (use of
force) or 5 (war) in the COW database.11 Robustness analysis is conducted on a stricter definition of war, i.e. MID of
hostility levels 4 and 5.
Such qualitative data on military conflicts require that the actors, duration, geographical location and intensity of each
conflict have been coded by researchers. Such a process prevents similar exercises on conflicts of lower intensity. In order
to measure dispute occurrence, I use an alternative type of data: event data. Event data are reported, by trained students or
automatically by computers, on a day-by-day basis from newspapers or wire services and coded by actor, target, form of
action and date. Data on daily events have the great advantage of providing information on interstate interactions
whatever the intensity of the underlying event. Event data do not allow us to assess the evolution of a given conflict over
time. They nevertheless make it possible to measure the occurrence of disputes that clear a minimum threshold between
two countries in a given year and that are resolved through negotiation or using military force. I use the event data
compiled by Kinsella and Russett (2002) to measure disputes exceeding a threshold defined as strong verbal hostility.12
Kinsella and Russett (2002) overlap data from three event databases, the Conflict and Peace Data Bank (COPDAB), the
World Event/Interaction Survey (WEIS) and the Protocol for the Assessment of Nonviolent Direct Action (PANDA), to
construct a dummy variable coded 1 if a dispute occurs for any dyad-year over the 1950–1991 period.13 Table 1 provides
event categories coded as disputes and their level of severity according to the Goldstein (1992) scale. Only events classified
at least as conflictual as categories ‘‘Cancel or postpone planned events’’ and ‘‘Charge; criticize; blame; disapprove’’ are
coded as a dispute.
Table 2 shows that the proportion of MIDs and RTA members remains similar when the sample is restricted due to the
availability of event and trade data. MIDs are nevertheless slightly biased towards less severe MIDs. Out of the 128 368
dyad-years of our restricted sample, 7937 experience a dispute, of which 586 spillover into MIDs.
Appendix A provides details on the source of other data. The dataset includes 103 RTAs, comprising 17 partial scope
agreements, 70 free trade agreements, 14 customs unions and two common markets.
11
MID level 2 (threat to use force) is not deemed to be a military conflict. See the COW website (http://www.correlatesofwar.org/) for more
information and records of MIDs.
12
See Kinsella and Russett (2002, pp.1054–1055) for more details on databases used and the operationalizing of the minimum conflict intensity
threshold. Schrodt and Gerner (2000) present limitations related to the use of event data. Using events exceeding a certain intensity reduces the biases
they identify.
13
One hundred and eighty nine cases exhibit a MID but no dispute in the restricted sample, of which 170 exhibit a dispute in t 1. I follow Kinsella
and Russett (2002) and treat them as measurement errors, due to the fact that event databases rely on major news media and do not cover all regions of
the world equally. The dummy variable is thus recoded as if a dispute occurred.
V. Vicard / European Economic Review 56 (2012) 54–71
57
Table 1
Events and Goldstein scale.
Event category
Goldstein
Request action; call for
Explicit decline to comment
Urge or suggest action or policy
Comment on situation
Deny an accusation
Deny an attributed policy, action, role or position
Grant asylum
Make complaint (not formal)
Cancel or postpone planned events
Charge; criticize; blame; disapprove
Issue formal complaint or protest
Give warning
Denounce; denigrate; abuse
Halt negotiation
Turn down proposal; reject protest, demand, threat
Refuse; oppose; refuse to allow
Reduce routine international activity; recall officials
Detain or arrest person(s)
Threat without specific negative sanction stated
Issue order or command, insist, demand compliance
Expel organization or group
Order person or personnel out of country
Nonmilitary demonstration, walk out on
Reduce or cut off aid or assistance; act to punish/deprive
Threat with specific negative nonmilitary sanction
Ultimatum; threat with negative sanction and time limit
Threat with force specified
Break diplomatic relations
Armed force mobilization, exercise, display; military buildup
Noninjury destructive action
Nonmilitary destruction/injury
Seize position or possessions
Military attack; clash; assault
0.1
0.1
0.1
0.2
0.9
1.1
1.1
1.9
2.2
2.2
2.4
3
3.4
3.8
4
4
4.1
4.4
4.4
4.9
4.9
5
5.2
5.6
5.8
6.9
7
7
7.6
8.3
8.7
9.2
10
Source: Goldstein (1992).
Table 2
Descriptive statistics 1950–1991.
Sample
Full
Restricteda
Observations
MIDs of which:
Display of force (3)
Use of force (4)
War (5)
Dispute
366 546
1969
14%
66%
20%
–
128 368
586
20%
75%
5%
7937
Deep RTAs
Shallow RTAs
Mean
0.007
0.028
0.008
0.042
a
Sample conditioning on the explanatory variables in column 4 of Table 3.
3. The impact of regionalism on the likelihood of war
3.1. Empirical strategy
The aim of this section is to test the proposition that RTAs requiring significant political integration do reduce
significantly the probability that disputes between members escalate into war, whereas other RTAs do not. As emphasized
above, using a simple probit or logit model to estimate the conditional probability of war would severely bias results
because heterogeneity in dispute occurrence creates a selection bias. The literature on the determinants of war have dealt
with this issue either by limiting the sample to ‘politically relevant dyads’ (i.e. pairs of countries sharing a common border
58
V. Vicard / European Economic Review 56 (2012) 54–71
Table 3
Impact of RTAs on war: bivariate censored probit model.
Dependent variable
Deep RTA membership dum.
Shallow RTA membership dum.
No. of peaceful years
Log distance
Contiguity dum.
(1)
(2)
MID
MID
0.21
(0.20)
0.07
(0.11)
0.02a
(0.00)
0.23a
(0.04)
0.78a
(0.10)
(3)
MID
b
0.48
(0.24)
0.09
(0.15)
0.03a
(0.01)
0.10
(0.09)
0.66a
(0.16)
Dispute
a
0.67
(0.23)
0.20c
(0.11)
0.01a
(0.00)
0.04
(0.06)
0.47a
(0.15)
No. of landlocked countries
(4)
a
0.26
(0.09)
0.00
(0.05)
0.00a
(0.00)
0.28a
(0.02)
0.58a
(0.07)
0.36a
(0.03)
Bil. trade dependence (t 4)
MID
Dispute
b
0.59
(0.29)
0.06
(0.12)
0.01a
(0.00)
0.08
(0.11)
0.60a
(0.17)
0.51
(0.55)
0.26b
(0.13)
0.05
(0.19)
0.28a
(0.10)
0.17
(0.24)
0.09
(0.13)
0.00
(0.05)
0.38a
(0.09)
0.27
(0.21)
0.25
(0.37)
Multil. trade dependence (t 4)
Zero trade dum. (t 4)
Common language dum.
Colonial relationship dum.
Common colonizer dum.
Sum log area
Sum of democracy indexes
Common defense alliance dum.
UN voting correlation (t 4)
Observations
Uncensored observations
Log likelihood
Rho (Wald test of independent eqn.)
201 627
–
4518.0
–
6503
–
924.8
–
Estimation method
Sample
Time dummies
Probit
Full
Yes
Probit
dist o 1000 km
Yes
201 627
12 047
43 452.2
0.32
Full
Yes
0.11
(0.10)
0.03
(0.06)
0.00a
(0.00)
0.32a
(0.02)
0.29a
(0.07)
0.22a
(0.03)
1.66a
(0.23)
0.30a
(0.05)
0.21a
(0.03)
0.16a
(0.04)
0.55a
(0.09)
0.11b
(0.06)
0.13a
(0.01)
0.23a
(0.03)
0.48a
(0.06)
0.99a
(0.05)
128 368
7937
2358.3
0.18
Bivariate probit with censoring
Full
Full
Full
Yes
Yes
Yes
Note: Robust standard errors in parentheses. Intercept and time dummies not reported. Standard errors clustered by country pair.
a
Significance at the 1% level.
b
Significance at the 5% level.
c
Significance at the 10% level.
or involving a major power) or by focusing on one cross-sectional determinant of disputes, i.e. bilateral distance. I go one
step further and use event data to directly measure disputes (Kinsella and Russett, 2002).
Determinants of dispute occurrence are indeed numerous. Focusing on some as a proxy for dispute probability ignores
much heterogeneity in disputes between country pairs and over time. Indeed, the restricted sample for which data on
disputes and other explanatory variables are available (column 4 of Table 3) contains 584 MIDs at the country pair level,
of which 404 (69%) occur between ‘‘politically relevant dyads’’ and 199 (34%) between dyads separated by less than
1000 km.14
Since the escalation into war process can only be observed when a dispute has occurred, using a bivariate probit
with censoring is a natural empirical strategy to estimate the conditional probability of war for each dyad-year.
The log-likelihood function is based on the unconditional probabilities associated with the three possible outcomes
(Greene, 2003, p. 713): no dispute (dispute¼0), a dispute emerges but does not escalate into war (dispute¼1 and war¼0),
and the dispute escalates into war (dispute¼1 and war¼ 1). Two equations are jointly estimated, one explaining the
14
The corresponding figures for the full sample are 1969 MIDs, of which 1358 (69%) are between ‘‘politically relevant dyads’’, and 595 (30%) are
between countries separated by less than 1000 km.
V. Vicard / European Economic Review 56 (2012) 54–71
59
initiation of the dispute and the second the dispute’s escalation into war. Consider y1 and y2, two latent (unobserved)
variables, representing the difference in utility levels from the initiation of the dispute and the dispute’s escalation into
war respectively. The model estimated is derived from a standard bivariate probit model:
(
1 if y1 4 0
y1 ¼ b1 X1 þ e1 and dispute ¼
,
0 if y1 r 0
(
y2 ¼ b2 X2 þ e2
and war ¼
1
if y2 40
0
if y2 r0
,
ð2Þ
where X1,2 are vectors of explanatory variables, b1,2 vectors of parameters, and error terms e1 and e2 are assumed to be
independent from X1,2 and to follow Eðe1 Þ ¼ Eðe2 Þ ¼ 0, Varðe1 Þ ¼ Varðe2 Þ ¼ 1, and Cov½e1 , e2 ¼ R.
Wooldridge (2002, p. 564) emphasizes that, technically, the coefficients can be identified due only to the nonlinearity of
the two equations in the bivariate probit. Hence, it is not necessary for X2 to be a strict subset of X1 for the outcome
equation to be identified. However, the identification of the model’s parameters is better achieved when X1 contains at
least one variable that is not in X2, so that we have an exclusion restriction, i.e. a variable that influences the selection
equation but not the outcome equation. The number of landlocked countries in a dyad is a good candidate as an
identification variable. Having no access to the ocean indeed reduces a country’s geographical scope of interest and
reduces the opportunity for contacts with countries other than neighboring countries. Landlocked countries should
therefore be less likely to interact with distant countries, through cooperative as well as conflictual actions, and should
have less interstate disputes on average. There is however no reason that being landlocked affects the way conflicts are
settled.15
All specifications control for autocorrelation by clustering standard errors by country pair.
3.2. Results
The results are presented in Table 3. All specifications include the basic determinants of war put forward by the
literature in political science: contiguity, distance and the number of peaceful years between the two countries (Beck et al.,
1998). I first present the results from a simple probit estimator. Column 1 shows that RTA membership has no significant
effect on war probabilities on the full sample of dyad-years. When only countries separated by less than 1000 km are
considered, belonging to a deep RTA, but not a shallow one, does significantly reduce war probabilities (specification (2)).
These crude results emphasize the need to account for heterogeneity in dispute occurrence. Specification (3) reports
estimates using the bivariate probit accounting for selection. The first and second columns present the results of the
escalation into war and dispute initiation equations respectively. As expected, landlocked countries face a lower
probability of dispute. The coefficient on the exclusion variable is significant at the 1% level in all specifications. The
results confirm that membership of a deep RTA does significantly reduce the probability that a dispute escalates into war.
In this specification, membership of a shallow RTA is found to have a lower, albeit significant, effect on escalation into war
probabilities. It is however not significant when additional controls are included.
Specification (4) in Table 3 includes several potential co-determinants of regionalism and war (see Appendix A for data
source). First, I include year dummies to control for any global shock affecting war probabilities and regionalism over time.
Other controls can be divided into two sets: trade and political variables. Regarding trade, we control for both bilateral (the
log of the mean of bilateral imports as a percentage of GDP) and multilateral (the log of the mean of multilateral (excluding
bilateral) imports as a percentage of GDP) trade openness.16 Martin et al. (2008) indeed argue that trade has an ambiguous
effect on peace. Bilateral trade does reduce war probabilities but multilateral trade openness, by reducing the dependence
on a particular partner, increases war probabilities. The results confirm that multilateral trade openness increases the
probability that a dispute escalates into war, but show no significant effect of bilateral trade.17 A dummy variable for zero
trade flows is included as well as dummies indicating countries sharing a common language, countries that have had a
colonial relationship or a common colonizer. Countries sharing a colonial history are indeed more likely to have unresolved
disputes. They also share historical, cultural and institutional traits that make them more likely to create a RTA.
The set of political variables control for political regime, the size of countries, alliances and diplomatic relations. The
level of democracy is included as a control since democracies have been found to be less likely to wage wars (see Levy and
Razin (2004) and Jackson and Massimo (2007) for a theoretical treatment, and Oneal and Russett (1997) among others for
empirical evidence). Democratic status has also been argued to affect the choice to create an RTA (Mansfield et al., 2002).
The (sum of the log of the) area of the two countries is included since a bigger territory is exposed to more opponents.
Larger countries also depend less on foreign trade, which could affect their incentives to create RTAs. Finally, I control for
diplomatic affinity between countries using a dummy variable for common membership of a defense alliance and the
correlation of voting in the UN General Assembly (lagged four years).
15
16
17
When introduced into a probit model of the second stage equation, the number of landlocked countries is not statistically significant.
Following Martin et al. (2008), trade variables are lagged 4 years to remove any contemporaneous reverse effect of war on trade.
More precisely, the peaceful effect of bilateral trade vanishes when other controls are added. Martin et al. (2008) find a similar pattern.
60
V. Vicard / European Economic Review 56 (2012) 54–71
Table 4
Impact of RTAs on war: robustness.
Dependent variable
(1)
MID 4 & 5
(2)
MID
(3)
MID
(4)
MID
(5)
MID
(6)
MID
Deep RTA membership dum.
0.52c
(0.29)
0.07
(0.11)
1.19b
(0.59)
0.23
(0.26)
0.48c
(0.28)
0.02
(0.12)
0.36
(0.24)
0.46b
(0.22)
1.45a
(0.40)
0.51
(0.32)
0.03
(0.12)
0.51c
(0.27)
0.05
(0.11)
0.57b
(0.29)
0.06
(0.12)
Shallow RTA membership dum.
No. of major powers
One communist country dum.
Two communist countries dum.
(7)
MID
1.04a
(0.28)
0.18
(0.39)
0.21a
(0.08)
GATT membership dum.
0.07c
(0.04)
0.03
(0.02)
Sum log military
expenditure (t 4)
Abs. diff. log military expenditure (t 4)
First stage IV
Sum deep RTAs with third countries (t 5)
Shallow
0.06a
(0.02)
0.18a
(0.01)
Sum shallow RTAs with third countries (t 5)
Observations
Uncensored observations
Log likelihood
Rho (Wald test of independent eqn.)
128 368
7937
23 245.2
0.40
55 409
5013
12 998.7
0.18
128 368
7937
22 659
0.22
128 368
7937
23 447.7
0.02
Estimation method
Sample
Time dummies
Full
Yes
Bivariate probit with censoring
OECD
Full
Full
Yes
Yes
Yes
123 405
7776
21 642.1
0.42
7937
–
1543.4
–
Full
Yes
Probit
dispute ¼ 1
Yes
Deep
0.24a
(0.03)
0.21a
(0.05)
7678
7678
–
–
1462.6
–
–
IV Probit
dispute ¼1
Yes
Note: Robust standard errors in parentheses. Intercept and time dummies not reported. All estimates include the same control variables as in our
preferred specification (specification (4) in Table 3). In specifications (6) and (7), only the second step of the bivariate censored probit is reported.
Standard errors clustered by country pair.
a
Significance at the 1% level.
b
Significance at the 5% level.
c
Significance at the 10% level.
Controlling for all these potential co-determinants of war and regionalism, specification (4) confirms that deep RTAs
promote the peaceful settlement of disputes between members. Membership of shallow RTAs has no effect on war
probabilities. Moreover, the insignificant impact of bilateral trade on the likelihood of a dispute escalating into war
confirms that the institutional channel dominates in regional integration. It is worth noting that when all these controls
are added, the Wald test of independent equations is no longer significant, meaning that the two equations estimated
simultaneously are independent.
3.3. Robustness
Tables 4 and 5 present a number of robustness tests.18 The first column tests the sensitivity of the results to a more
restrictive definition of war including only MIDs of hostility levels 4 and 5, i.e. implying the actual use of military force and
wars resulting in the death of at least 1000 military personnel. This more restrictive definition of war reduces slightly the
significance of the coefficient on deep RTAs but the results remain qualitatively similar. I also restrict the sample to
country pairs including at least one OECD country in 1991. Event data are indeed coded from international newspapers
which do not cover all the world’s countries evenly. Disputes involving richer countries may be more likely to be reported
by international newspaper than disputes involving poor countries. On this restricted sample, deep RTAs—but not shallow
ones—are still negatively and strongly significantly associated with the likelihood of disputes escalating into war. The
coefficient on deep RTAs is however twice as large as in the benchmark estimation, suggesting measurement errors.
Finally, columns 3–5 control for additional potential co-determinants of war and regionalism. Controlling for the number
of major powers, communist regimes, membership of GATT and the level and difference of (log of) military expenditure
18
To save space, only the coefficients of the variables of interest and of the second stage equation are reported. All specifications include the control
variables included in specification (4) of Table 3.
V. Vicard / European Economic Review 56 (2012) 54–71
61
Table 5
Impact of RTAs on war: robustness, continued.
Dependent variable
(8)
MID
(9)
MID
(10)
MID
(11)
MID
(12)
MID
(13)
MID
(14)
MID
Deep RTA membership dum.
0.54c
(0.31)
0.04
(0.13)
0.17c
(0.09)
0.05
(0.22)
0.53c
(0.29)
0.02
(0.12)
0.18b
(0.09)
0.03
(0.21)
0.03
(0.05)
0.11c
(0.06)
0.14a
( 0.03)
0.05
( 0.04)
0.12a
(0.03)
0.03
(0.04)
0.14a
(0.04)
0.00
(0.04)
0.10a
(0.03)
0.06
(0.04)
0.58b
(0.29)
0.06
(0.12)
0.87a
(0.2)
0.57c
(0.3)
0.06
(0.06)
0.19
(0.12)
0.35
(0.53)
0.34
(0.29)
0.76
(0.61)
0.05
(0.29)
0.43
(0.92)
0.55
(0.42)
Shallow RTA membership dum.
One oil exporter dum.
Two oil exporters dum.
Abs. diff in log GDP per capita (t 4)
log sum GDP per capita (t 4)
Interaction deep RTA
Interaction shallow RTA
Abs. diff in share agriculture in value added
Avg share of agriculture in value added
Abs. diff in share industry in value added
0.04
(0.06)
0.19b
(0.09)
Avg share of industry in value added
0.17b
(0.07)
0.08
(0.12)
Abs. diff in share manufacturing in value added
Avg share of manufacturing in value added
Abs. Diff in of share services in value added
0.14
(0.15)
0.13
(0.26)
Avg share of services in value added
Lagged deep RTA membership dum.
0.56
(0.50)
Observations
Uncensored observations
Log likelihood
Rho (Wald test of independent eqn.)
128 368
7937
23 456.3
0.01
128 368
7937
23 318.9
0.15
Estimation method
Sample
Time dummies
Heckprob
Full
Yes
Full
Yes
4204
–
–
–
4190
–
–
–
4190
–
–
–
Yes
LPM
dispute ¼ 1
Yes
Yes
1867
–
–
–
128 368
7937
23457.7
0.19
Yes
Heckprob
Full
Yes
Note: Robust standard errors in parentheses. Intercept and time dummies not reported. All estimates include the same control variables as in our
preferred specification (specification (4) in Table 3). Standard errors clustered by country pair.
a
Significance at the 1% level.
b
Significance at the 5% level.
c
Significance at the 10% level.
does not affect the results. Nonetheless, omitted (unobservable) variables could bias the results. The next section
implements an instrumental variable (IV) strategy to deal with this issue.
3.3.1. Instrumental variables
The domino theory of regionalism suggests that the creation or enlargement of an RTA increases incentives of nonmembers to apply for membership (Baldwin, 1997). Using spatial econometrics, Egger and Larch (2008) show that the
probability that two countries create an RTA increases with the creation of RTAs by third countries. The number of RTAs
signed by each country with third countries could therefore qualify as a strong instrument correlated with the existence of
an RTA between two countries in a given year. Since we control for bilateral and multilateral trade dependence in our
regression, there is no reason to believe that the number of RTAs signed with third countries is correlated with war
probabilities. The number of deep and shallow RTAs signed with third countries in t 5 by the two countries are used
separately as instrumental variables for RTA membership. The enlargement of an existing RTA to include a new country
indeed extends the same form of RTA; a country that is already a member of a deep RTA would be more likely to create a
deep RTA with a new country and less likely to create a shallow RTA.
62
V. Vicard / European Economic Review 56 (2012) 54–71
Since our endogenous variables are dummy variables, using the traditional two stage methodology would yield
inconsistent estimates unless the first-stage model is exactly right. The fitted value from a probit model may nevertheless
be used as an instrument, together with other exogenous covariates, to generate first-stage estimates using an ordinary
least square (OLS) model (Angrist and Krueger, 2001). We thus apply a three stage methodology. In the first stage, fitted
values for deep and shallow RTA membership are generated using a probit model with our two instruments. The fitted
values are used, together with other exogenous covariates, to predict the two endogenous RTA membership dummies
using OLS. These predictions are then used in the estimation of war probabilities. In the third stage, we estimate directly
the probability of war between countries involved in a dispute, i.e. the second stage of the bivariate probit with censoring.
The Wald test of independent equations reported in Table 3 indeed shows that the second equation, i.e. dispute escalating
into war, can be estimated independently when we control for enough covariates. For the sake of comparison, specification
(6) in Table 4 presents the estimate of the dispute escalating into war equation using a probit model without IV. Results on
deep and shallow RTA membership are similar to specification (4) in Table 3.
The coefficient on the IVs in the first stage equation are reported at the bottom of column (7) in Table 4. Both are
significant at the 1% level, confirming that they are strong instruments. As expected from the domino theory of
regionalism, countries that have already signed RTAs with third countries are more likely to conclude similar agreements
with new partners. The results of the final stage of the IV empirical strategy confirm that members of deep - but not
shallow - RTAs are less likely to escalate a dispute into war. The coefficient is strongly significant and is almost twice as
large as in the benchmark regression. It suggests that unobservable variables that make country pairs more likely to belong
to the same deep RTA simultaneously increase the probability that a dispute escalates into war.
3.3.2. Asymmetric members
Schiff and Winter (2003) argue that a deep RTA may have negative security effects when countries are very different,
because tariff preferences under a common external tariff may lead to large income transfers and a concentration of
industry in a given country/region. Asymmetric gains from regional integration may therefore increase the risk of conflicts
between unbalanced members. Table 5 deals with this issue. First, columns 8 and 9 include additional control variables for
differences in export specialization: dummy variables indicating whether one or two countries in the pair are oil exporters,
as defined by the IMF in the DoTS, and the difference and average level of GDP per capita. The results show that poor
countries and country pairs involving one oil exporter are more likely to escalate disputes into war, but the deep RTA
variable remains significant. Differences in GDP per capita are found to have no impact on war probabilities.
A direct test of Schiff and Winter’s argument requires to account for the interaction between asymmetries between
countries and RTA membership. The non-linearity of our estimator however makes the interpretation of coefficients on
interaction terms potentially misleading (Ai and Norton, 2003). I therefore estimate a linear probability model (LPM) on
the dispute escalating into war stage. This alternative methodology yields the marginal effects on the interaction terms
(Angrist and Pischke, 2009). Schiff and Winter’s argument involves a positive coefficient on the interaction between
countries asymmetries and deep RTA membership. Asymmetries between countries’ structure of production are measured
by the difference in the share of agriculture, industry, manufacturing and services in value added. Columns 10–13 of
Table 5 present the results using, in turn, each measure and its interaction with deep and shallow RTA membership
dummies. The average share of the sector considered (agriculture, industry, manufacturing and services) in value added is
also controlled for. We find weak support for the argument of Schiff and Winter (2003). As expected, the coefficient on the
interaction term with the deep RTA variable is positive, but significant only when using the difference in the share of
agriculture. Note that the impact of deep RTA membership is negative for the median level of differences in production
structure.
Finally, column 14 shows that deep RTAs that fall apart do not increase conflicts in the following 5 years.
3.4. Quantification
This section aims to quantify the impact of RTA membership on war probabilities. Since the estimator used is nonlinear,
coefficients cannot be interpreted immediately. Fig. 1 presents marginal effects (or discrete change for dummy variables) on
the conditional (on dispute occurrence) predicted probability @Prðwar ¼ 19dispute ¼ 1Þ=@X , computed from specification (4)
in Table 3. The marginal effects are computed at two different levels: at the mean value of all variables and for contiguous
countries separated by 1000 km and exhibiting positive trade flows (all other variables being held at their mean).
For contiguous countries separated by 1000 km, being members of a customs union or common market reduces the
probability that a dispute escalates into war by 9.1 percentage points. This effect is sizeable since the mean predicted
conditional probability is 13.8%. The impact of belonging to a deep RTA is comparatively similar when all variables are held
at their mean: it reduces the probability of a dispute escalating into war by 1.9 percentage points, while the predicted
conditional probability is 2.5%.
In comparison with other determinants of war, this effect is also sizeable. Belonging to a deep RTA is equivalent to
increasing the number of peaceful years between two countries by 28 years from its mean value. Moreover, the peaceful
effect of sharing a common language or a common defence alliance is two times smaller. Finally, when compared to
the democratic peace channel, the impact of deep RTA membership is shown to be of the same order of magnitude as
an increase from the mean to the top level of democracy indexes for the two countries.
V. Vicard / European Economic Review 56 (2012) 54–71
63
0.16
Predicted probability
Deep RTA membership dum.
10 additional peaceful year
0.12
Common language
Sum democracy indexes (1 std.dev.)
Common defence alliance dum.
0.08
0.04
0
Contiguous and trading countries, separated by
Mean country pair
1000km
Fig. 1. Impact of RTAs on war: quantification.
4. Determinants of deep and shallow regional trade agreements
Having established that deep and shallow RTAs have differential effects on the probability that a dispute escalates into
war, this section tests whether security issues are a motive for creating RTAs.
4.1. Empirical specification
In line with Baier and Bergstrand (2004), I estimate the following specification on a cross-section of country pairs
in 2005:
PrðRTAhij ¼ 1Þ ¼ b0 þ b1 DPij þ b2 OPEN ij þ b2 Controlsij þ eij ,
h ¼ fdeep,shallowg
ð3Þ
where Controlsij includes economic and political determinants of regionalism, DPij is the propensity to dispute and OPENij is a
proxy for multilateral trade openness. Baier and Bergstrand (2004) show that, in a model of RTA formation in a secure world
(i.e. without interstate conflict), two countries facing low trade costs with the rest of the world should have lower welfare
gains from joining an RTA compared to country pairs facing high trade costs. Martin et al. (2008) however show that countries
more open to multilateral trade are more likely to escalate disputes into war. In an insecure world, a country facing low trade
costs should therefore need more mechanisms to prevent dispute from escalating into war and secure gains from trade with
their partners. They are more likely to create deep rather than shallow RTAs. Trade costs with the rest of the world should
unambiguously decrease the likelihood of shallow RTAs but are likely to have an ambiguous impact on the likelihood of deep
RTAs. Baier and Bergstrand (2004) measure the remoteness of countries by their mean distance to their trading partners.
Anderson and van Wincoop (2003) nevertheless show that remoteness indexes are a bad proxy for trade openness and
propose a strategy for estimating structurally what they call multilateral resistance indexes from a gravity model of trade. The
methodology used to compute the multilateral resistance indexes is presented in Appendix B.
Other economic determinants of RTAs are those put forward by Baier and Bergstrand (2004). Welfare gains from
creating an RTA should increase with the size and similarity of GDPs of the two countries. Pairs of countries with dissimilar
relative factor endowment and whose relative factor endowments are similar to the rest of the world should also receive
higher welfare gains from creating an RTA. Relative factor endowment is measured by GDP per capita.
The set of political controls includes a number of potential co-determinants of disputes and regionalism. First, the fact
of sharing a common border, a common language or a common colonial history is likely to affect the probability of bilateral
disputes as well as the incentives to create a RTA. Landlocked countries experience fewer disputes on average and face, at
the same time, higher trade costs. The level of democracy of the two countries is also included as a control. Finally,
diplomatic affinity between states is likely to influence decisions to create RTAs as well as the set of disputes between
countries. I use two variables to control for diplomatic affinity: a dummy variable indicating countries sharing a common
defense alliance and the correlation of voting in the UN General Assembly (see Appendix A for details on data).
Finally, dispute probability is measured using the event data presented above, as the propensity to bilateral disputes
between 1950 and 1969. I use lagged dispute data because past membership of RTAs is likely to affect variables in 2005.
64
V. Vicard / European Economic Review 56 (2012) 54–71
An important issue when estimating Eq. (3) is the endogeneity related to past membership of RTAs (Baier and Bergstrand,
2004). Lagged data may suffer from an endogeneity bias as well since some regional integration processes began shortly
after World War II (the Benelux, the EU,y) and several current RTAs have been preceded by earlier attempts at integration
in the 1950s and 1960s.19 Moreover, endogeneity may also arise because of omitted variables affecting the likelihood of
RTAs and disputes.
To deal with endogeneity, I implement an IV strategy.20 Since the dispute propensity variable is not a continuous
variable, there is no straightforward nonlinear IV model available. As suggested by Angrist and Krueger (2001) and
Wooldridge (2002), I use a three-stage least square (3SLS) model to estimate Eq. (3) for deep and shallow RTAs
simultaneously taking dispute propensity as endogenous. The two exogenous instruments for dispute propensity are
derived from international relations theory; they are the number of major powers in the country pair and the difference in
national military expenditure. Major powers are countries that have military, diplomatic, economic and cultural power
that allow them to exert a global influence (Waltz, 1979). Unlike other countries whose geographical scope of interest is
regional or local, major powers have a global geographical scope of interest. They are thus likely to experience more
interstate disputes with any country around the globe. Major power status is however not likely to be correlated with
incentives to create RTAs since it is not defined solely by economic power—Japan and Germany are not considered to be
major powers, even if they are major economic powers. Moreover, controls for economic size are included in the main
specification, which should remove the economic component of major power status. The second instrument is the
difference in (log) military expenditure. Countries that have unbalanced military capabilities are indeed more likely to be
involved in disputes (Kinsella and Russett, 2002). While the level of military expenditure is likely to be affected by
decisions on RTAs, the difference in military expenditure is not, since both countries should decrease their military
spending proportionally in the face of decreasing military threats.
Since no exogenous instruments for multilateral resistance terms and GDP are available, these variables are lagged in
1960. This strategy reduces the sample of countries, because several countries were not independent in 1960. We are left
with a sample of 2814 unique country pairs.
4.2. Results
Table 6 presents estimates of Eq. (3). I start with a simple univariate regression.21 Dispute propensity is positively and
significantly associated with the likelihood of deep RTAs, but has no significant effect on shallow RTAs. These crude results
may suffer from an endogeneity bias. Specification (2) of Table 6 controls for endogeneity using an IV model. Regarding the
first stage equation, the number of major powers and the difference in military expenditure strongly and significantly increase
dispute propensity. The partial R2 of IVs is 0.29 and the F-statistic of weak identification on IVs greatly exceeds the threshold
of ten recommended by Staiger and Stock (1997), confirming the relevance of the two instruments. In this specification, the
Hansen–Sargan overidentification test however rejects the exogeneity of the instruments. This simple univariate specification
does not control for the economic size of countries, which should affect welfare gains from joining a RTA. Since economic
power is also one of the attributes of major powers, the number of major powers is not exogenous when economic size is not
controlled for. In specification (3), when we control for economic and political determinants of regionalism, the Hansen–
Sargan overidentification test does not reject the exogeneity of IVs, thus confirming their validity. As expected, the partial R2
and F-statistic on IVs are greatly reduced, but still confirm that the instruments can be regarded as strong.
Controlling for endogeneity confirms that dispute propensity has a differential impact on incentives to create deep and
shallow RTAs. The coefficients on dispute propensity are significant at the 1% level. Country pairs that have many interstate
disputes are more likely to create a deep RTA and less likely to create a shallow one.
The relevance of security motives for decisions on trade policy is further illustrated by the differential impact of trade
openness on the likelihood of deep and shallow RTAs. Pairs of countries naturally more open to trade (having larger
multilateral resistance indexes) tend to create deep RTAs. Countries naturally more open to trade depend less on one
partner in particular, which increases the probability that disputes spillover into war (Martin et al., 2008) and increases
accordingly the need for international institutions supporting the peaceful resolution of disputes. By contrast, shallow
RTAs are driven by the economic motives put forward by Baier and Bergstrand (2004) and are created between country
pairs facing large trade costs with the rest of the world.
Geographically close countries and countries sharing a common border are more likely to join all kinds of RTAs. Other
economic determinants of regionalism are in line with Baier amd Bergstrand’s (2004) results concerning shallow RTAs. Country
pairs with similar endowments and with large and similar markets tend to create shallow RTAs. As in Egger and Larch (2008),
countries that have endowments similar to the rest of the world are more likely to create shallow RTAs. Economic determinants
are however less relevant to the creation of deep RTAs. Differences in endowments have no significant impact on the creation of
19
For instance, the East African Community was disbanded in 1977 and relaunched in 2000 as a partial scope agreement and the Central American
Common Market collapsed in 1975 and relaunched as a customs union in 1993.
20
Another advantage of using an IV model is that it also deals with measurement error in the endogenous explanatory variable, which is, as
explained above, also valuable in our case. For instance, institutions in deep RTAs are likely to publicize disputes, creating a downward bias on the
coefficient of dispute propensity in the case of deep RTAs.
21
Using a bivariate probit estimator yields similar results.
V. Vicard / European Economic Review 56 (2012) 54–71
65
Table 6
Probability of deep and shallow RTAs between two countries.
Estimator
(1)
(2)
(3)
Deep RTA
Shallow RTA
3SLS
Deep RTA
Shallow RTA
3SLS
Deep RTA
Shallow RTA
3SLS
sample
Propensity to dispute
Full
0.08a
(0.02)
Full
0.02
(0.05)
0.11a
(0.04)
Full
0.38a
(0.08)
Sum log multilateral resistance indexes
Log distance
Continent dum.
Abs. Diff in log GDP per capita (1960)
Squared abs. diff in log GDP per capita (1960)
Abs. Diff of (log) GDP per capita with RoW (1960)
Log sum GDP (1960)
Log diff. GDP (1960)
Contiguity dum.
Common language dum.
Common colonizer dum.
No. of landlocked countries
Sum of democracy indexes
Common defence alliance dum.
UN voting correlation
0.29a
(0.01)
0.00c
(0.00)
No. of major powers
Abs. diff. log military expenditure
Observations
Hansen–Sargan overidentification statistic
Partial R2d
F-test on IV
0.27a
(0.07)
0.09a
(0.01)
0.09a
(0.01)
0.01
(0.01)
0.01
(0.01)
0.01c
(0.00)
0.01c
(0.01)
0.08a
(0.01)
0.02a
(0.01)
0.11a
(0.03)
0.04a
(0.01)
0.10a
(0.02)
0.02a
(0.01)
0.10a
(0.01)
0.06a
(0.01)
0.18a
(0.02)
2814
2814
2814
0.23a
(0.01)
0.00c
(0.00)
2814
60.94a
0.29
871.7
1.06a
(0.14)
0.16a
(0.03)
0.07a
(0.01)
0.08a
(0.03)
0.09a
(0.02)
0.01
(0.01)
0.05a
(0.01)
0.27a
(0.03)
0.14a
(0.01)
0.15a
(0.05)
0.02
(0.02)
0.14a
(0.04)
0.10a
(0.02)
0.16a
(0.03)
0.04
(0.03)
0.04
(0.04)
2814
2814
0.94
0.16
273.4
Note: Standard errors in parentheses.
a
Significance at the 1% level.
b
Significance at the 5 % level.
c
Significance at the 10% level.
d
Computed using 2SLS.
deep RTAs, and deep RTAs are more likely to cover smaller and more dissimilar country pairs in terms of GDP.22 Moreover, the
economic significance of GDP size is much smaller for deep than for shallow RTAs.
Conversely, political determinants are more relevant to the creation of deep RTAs than to shallow ones. Having a common
colonizer increases the likelihood of deep RTAs, but decreases the likelihood of shallow RTAs. Former colonies from the same
region are indeed more likely to have unresolved territorial disputes related to the decolonization process. They also share
institutional features from their past colonizer that reduce the cost of creating common supranational institutions. Sharing a
common language is surprisingly negatively and significantly associated with deep RTAs.23 In addition, diplomatic affinity only
22
Note that the negative impact of GDP similarity is difficult to reconcile with the theory of economic determinants of RTAs put forward by Baier and
Bergstrand (2004). The coefficient on GDP similarity however becomes insignificant when EU country pairs are excluded (see Table 7).
23
This result however disappears when EU countries are excluded (see Table 7). The fact that most direct communication languages originate from
Europe, because of its history of colonialism, and that these languages are those coded as common languages in other regions of the world may explain
this result.
66
V. Vicard / European Economic Review 56 (2012) 54–71
Table 7
Probability of deep and shallow RTAs between two countries: robustness.
1
2
Deep RTA
Estimator
sample
Propensity to dispute
Sum log multilateral resistance indexes
Log distance
Continent dum.
Abs. diff in log GDP per capita (1960)
Squared abs. diff in log GDP per capita (1960)
Abs. diff of (log) GDP per capita with RoW (1960)
Log sum GDP (1960)
Log diff. GDP (1960)
Contiguity dum.
Common language dum.
Common colonizer dum.
No. of landlocked countries
Sum of democracy indexes
Common defence alliance dum.
UN voting correlation
Shallow RTA
EU pairs excl.
Full
0.10b
(0.04)
0.01
(0.01)
0.04a
(0.01)
0.09a
(0.01)
0.01
(0.01)
0.00
(0.00)
0.01
(0.00)
0.02a
(0.01)
0.00
(0.00)
0.25a
(0.02)
0.01
(0.01)
0.14a
(0.01)
0.01c
(0.01)
0.05a
(0.01)
0.12a
(0.01)
0.06a
(0.01)
0.98a
(0.14)
0.11a
(0.03)
0.11a
(0.02)
0.13a
(0.03)
0.11a
(0.02)
0.01b
(0.01)
0.05a
(0.01)
0.23a
(0.03)
0.13a
(0.01)
0.08
(0.05)
0.01
(0.02)
0.18a
(0.04)
0.09a
(0.02)
0.12a
(0.03)
0.08a
(0.03)
0.04
(0.04)
0.38a
(0.07)
0.09a
(0.01)
0.07a
(0.01)
0.02
(0.01)
0.00
(0.01)
0.00
(0.00)
0.02a
(0.01)
0.10a
(0.01)
0.04a
(0.01)
0.08a
(0.03)
0.04a
(0.01)
0.11a
(0.02)
0.03a
(0.01)
0.11a
(0.01)
0.06a
(0.01)
0.16a
(0.02)
0.01a
(0.00)
0.24a
(0.01)
0.00b
(0.00)
No. of major powers
Abs. diff. log military expenditure
Shallow RTA
3SLS
Total GDP of RTA partners
Observations
Hansen–Sargan overidentification statistic
Partial R2d
F-test on IV
Deep RTA
3SLS
2736
0.23a
(0.01)
0.00b
(0.00)
2736
3.79
0.17
289.2
0.40a
(0.08)
0.11a
(0.02)
0.06a
(0.01)
0.02
(0.02)
0.00
(0.01)
0.00
(0.00)
0.01
(0.01)
0.11a
(0.02)
0.04a
(0.01)
0.03
(0.03)
0.05a
(0.01)
0.08a
(0.02)
0.04a
(0.01)
0.09a
(0.02)
0.03c
(0.02)
0.17a
(0.02)
0.05a
(0.00)
2814
2814
2.37
0.16
260.5
Note: Standard errors in parentheses.
a
Significance at the 1% level.
b
Significance at the 5% level.
c
Significance at the 10% level.
d
Computed using 2SLS.
impacts the decision to create a deep RTA: having a common defense alliance or similar voting patterns in the UN General
Assembly increases the likelihood of deep RTAs between two countries. Finally, states’ democratic status also have a differential
impact on deep and shallow RTAs. Democratic countries tend to create deep RTAs, whereas they are less likely to create shallow
RTAs. Joining a deep RTA indeed involves sharing common supranational institutions and providing some public goods in
common. Giving up some national sovereignty is possible only between countries that are similar in terms of political system,
type of government and origin of legitimacy. This constraint is less binding regarding shallow RTAs, in which more autocratic
regimes can retain more independent power while benefiting from gains from trade.
In Table 7, I perform two robustness tests. First, the argument put forward in this paper draws on the EU experience.
The EU is undoubtedly the most integrated RTA worldwide and accounts for a large share of country pairs in deep RTAs in
our sample. In the first two columns of Table 7, I re-estimate the baseline specification excluding EU country pairs.
V. Vicard / European Economic Review 56 (2012) 54–71
67
The results on dispute propensity remain qualitatively similar.24 The coefficient on natural trade openness becomes
insignificant for deep RTAs, but remains negative and significant for shallow RTAs. These results are in line with the
expected ambiguous impact of trade openness on incentives to create deep RTAs. The main results of the paper are
therefore robust to the exclusion of EU countries.
Finally, so far we have considered that the decision to create an RTA takes place at the country pair level. Many RTAs
have more than two members, and a country could well join an agreement because it expects gains with some members
but not all of them. Moreover, individual members of deep RTAs cannot sign separate agreements with third countries
because the common external tariffs are negotiated by all members. Not controlling for the characteristics of trade blocs
could bias the results. I thus re-estimate the baseline specification including a proxy for the economic size of the country
pair’s RTA partners if one or both countries are member of an RTA: (the log of) the sum of the GDPs of each country’s RTA
partners. The results show that RTAs’ characteristics matter. Members of large trading blocs are more likely to join deep
and shallow RTAs, but the impact is larger for shallow RTAs. Controlling for the economic size of RTAs does not alter
quantitatively the main results of the paper.
5. Conclusion
This paper asks whether international security is a motive for creating RTAs. I provide robust empirical evidence that
international security does matter, but that its impact depends on the form of trade integration. In this paper, RTAs are
classified according to their level of political integration. A deep RTA (customs union or common market) involves the
creation of common supranational institutions whereas a shallow agreement (partial scope or free trade agreement)
requires only weak institutional integration. The empirical analysis shows that deep RTAs promote the negotiated
settlement of disputes between members and prevent war. Shallow RTAs have no impact on war probabilities. I then show
that security issues are an important motive for the creation of deep RTAs. A pair of countries involved in many interstate
disputes and naturally more open to multilateral trade is more likely to create a deep RTA. Supranational institutions
created on the basis of a deep RTA help to manage interstate conflicts and secure gains from bilateral trade. Conversely,
interstate disputes reduce incentives to create shallow RTAs. Shallow RTAs are mainly driven by economic determinants.
This paper provides evidence of non-traditional gains from RTAs that help to explain the different strategies of regional
integration worldwide. The analysis focuses on wars because of the availability of data on such extreme form of
conflicts—essentially a regional issue. More broadly, by analyzing RTAs as regulating institutions in a world where no
supranational institution enforces property rights properly, this paper raises the question of the relationship between
trade integration and other areas of interstate cooperation. By providing an institutional framework or making available
sources of gains, RTAs may facilitate the internalization of international externalities and negotiations on other, more
complex, issues (Schiff and Winter, 2003). Such interactions could explain the differential impact of regionalism on
multilateralism depending on the form of RTAs found in the literature (Karacaovali and Lima~ o, 2008; Estevadeordal et al.,
2008). More work is however needed to understand the link between (regional) trade integration and deeper integration
on non-trade issues.
Acknowledgements
I would like to thank Thierry Mayer for providing data and guidance on this chapter of my Ph.D. dissertion. I also thank
James Anderson, Jeffrey Bergstrand, Antoine Berthou, Marius Bruhlart, Paola Conconi, Arnaud Costinot, Mathieu
Couttenier, Peter Egger, Philippe Martin, Vincent Rebeyrol, Mathias Thoenig, Julien Vauday and Thierry Verdier. I gratefully
acknowledge the financial support from ANR. This paper represents the views of the author and should not be regarded as
reflecting those of the Banque de France.
Appendix A. Data
Data on RTAs have been assembled from notifications to the WTO under article XXIV of GATT or the Enabling Clause for
developing countries, Frankel (1997), Foroutan (1993, 1998), Langhammer and Hiemenz (1990), Machlup (1977) and other
public sources. I look at all regional trade agreements which take the form of partial scope trade agreements, free trade
areas, customs unions, or common markets.25 Membership of an RTA is defined by dummy variables coded 1 when two
countries belong to the same RTA in year t (Table 8).
Trade data come from the database assembled by Katherine Barbieri, who mainly uses information from the IMF and
the League of Nations international trade statistics, and are supplemented by Martin et al. (2008) using the IMF DOTS
24
The results are also robust to the exclusion of country pairs including at least one EU country.
A partial scope trade agreement is defined as an agreement in which reciprocal preferences are exchanged to cover a limited range of the parties’
trade in goods (partial in scope); a free trade area is defined as an agreement in which reciprocal preferences are exchanged to cover substantially all
trade in goods; a customs union is defined as an RTA with the exchange of trade preferences and a common external tariff; and a common market is
defined as an RTA allowing free movements of factors (goods, capital and workers).
25
68
V. Vicard / European Economic Review 56 (2012) 54–71
Table 8
Name
Date
Common markets
Benelux
European Union (EU)a
1961
1992
Customs Unions
Benelux
European Communities (EC)
Equatorial Customs Union
Customs Union of West African States
East African Community
Mano River Union
Customs Union EU–Cyprus
Caribbean Community and Common Market (CARICOM)
Southern Common Market (MERCOSUR)a
Central American Common Marketa
Economic and Monetary Community of Central Africa
Andean Customs Uniona
Customs Union EU–Turkeya
West African Economic and Monetary Uniona
1947–1960
1958–1991
1959–1965
1960–1966
1967–1977
1973–1988
1973
1973
1991
1993
1994
1995
1996
1998
Free Trade Agreements
European Free Trade Agreement (EFTA)a
Central American Common Market
United Kingdom–Ireland
Caribbean Free Trade Area
EU–Norway
EU–Switzerland
EU–Egypta
Papua New Guinea and Australia Trade and Commercial Relation Agreement
Closer Trade Relations Trade Agreementa
United States of America–Israela
EU–Hungarya
EU–Polanda
EFTA–Turkeya
EFTA–Bulgariaa
EFTA–Hungarya
EFTA–Israela
EFTA–Polanda
Central European Free Trade Agreementa
European Economic Areaa
North American Free Trade Agreement (NAFTA)a
EU–Bulgariaa
Group of Threea
Mexico–Boliviaa
Mexico–Costa Ricaa
MERCOSUR–Chilea
MERCOSUR–Boliviaa
Canada–Chilea
Canada–Israela
Israel–Turkeya
Poland–Israela
EU–Tunisiaa
Hungary–Israela
Hungary–Turkeya
India–Sri Lankaa
Bulgaria–Turkeya
Chile–Mexicoa
EFTA–Moroccoa
EU–Israela
EU–Moroccoa
EU–Mexicoa
EU–South Africaa
Mexico–Israela
Poland–Turkeya
EU–Israela
EU–Mexicoa
EU–Moroccoa
EU–South Africaa
Dominican Republic–El Salvadora
Dominican Republic–Guatemalaa
Dominican Republic–Hondurasa
1960
1961–1975
1966–1972
1968–1972
1973
1973
1977
1977
1983
1985
1992
1992
1992
1993
1993
1993
1993
1993
1994
1994
1994
1995
1995
1995
1996
1996
1997
1997
1997
1998
1998
1998
1998
1998
1999
1999
1999
2000
2000
2000
2000
2000
2000
2000
2000
2000
2000
2001
2001
2001
V. Vicard / European Economic Review 56 (2012) 54–71
69
Table 8 (continued )
Name
Date
a
EFTA–Mexico
Mexico–El Salvadora
Mexico–Guatemalaa
Mexico–Hondurasa
United States–Jordana
Canada–Costa Ricaa
Chile–Costa Ricaa
Chile–El Salvadora
Dominican Republic–Costa Ricaa
EU–Jordana
EFTA–Jordana
El Salvador–Panamaa
EU–Chilea
CARICOM–Costa Ricaa
EFTA–Chilea
United States–Chilea
United States–Moroccoa
Thailand–Australiaa
Thailand–New Zealanda
Mexico–Japana
2001
2001
2001
2001
2001
2002
2002
2002
2002
2002
2002
2002
2003
2004
2004
2004
2005
2005
2005
2005
Partial scope agreements
Council for Mutual Economic Assistance (CMEA)
Latin American Free Trade Association (LAFTA)
Tripartite Agreementa
Protocol relating to Trade Negotiations among Developing Countriesa
West African Economic Community
Bangkok Agreementa
South Pacific Regional Trade and Economic Cooperation Agreement
Andean Community
General System of Trade Preferences among Developing Countries (GSTP)a
Economic Cooperation Organization
ASEAN Free Trade Agreement
Melanesian Spearhead Group
Latin American Integration Association (LAIA)a
Chile–Venezuelaa
Chile–Colombiaa
Common Market for Eastern and Southern Africa (COMSESA)a
South Asian Preferential Trade Agreementa
1949–1990
1961–1980
1968
1973
1973–1997
1976
1981
1988–1997
1989
1992
1992
1993
1993
1993
1994
1994
1995
Source: WTO (http://www.wto.org/english/tratop_e/region_e/region_e.htm), Foroutan (1993, 1998),
Langhammer and Hiemenz (1990), Frankel (1997), Machlup (1977) and other public sources.
a
RTAs included in the second part of the empirical analysis.
database. Income data also come from Martin et al. (2008), and are assembled from the Penn World Table (version 6.2),
Katherine Barbieri’s database and the World Bank WDI database. Geographical and colonial data are from the CEPII. Data
on formal defence alliances and military spending are taken from the COW project. The composite democracy indicator is
taken from Polity IV. It measures the openness/closedness of political institutions on a 10 / þ10 scale (10 indicates a
high level of democracy). Finally, UN voting correlation is taken from ‘‘The Affinity of Nations: Similarity of State Voting
Positions in the UN General Assembly’’ computed by Erik Gartzke. Data on shares of agriculture, industry services and
manufacturing in value added are from the World Bank WDI database.
Appendix B. Multilateral resistance terms
Let tij be the variable trade costs of exporting from i to j. Assuming symmetrical trade costs (tij ¼ tji ), a gravity-like
model of trade yields (Anderson and van Wincoop, 2003):
yi yj tij 1s
xij ¼
ð4Þ
yw Pi Pj
and
Pi1s
0
11s
X yj tij
@
A
¼
,
yw Pj
j
ð5Þ
where xij is the value of exports, s is the elasticity of substitution between all goods, and yj is country j’s nominal income.
70
V. Vicard / European Economic Review 56 (2012) 54–71
Table 9
Gravity equation with country fixed effects (1960).
Dependent variable
Imports
Log distance
0.54a
(0.04)
0.54a
(0.09)
0.12
(0.09)
0.53a
(0.16)
1.30a
(0.20)
0.27
(0.37)
0.20b
(0.09)
1.20a
(0.18)
0.15
(0.10)
1.61a
(0.13)
Contiguity dum.
Common language dum.
Ever in colonial relationship dum.
Colonial relationship since 1945 dum.
Common colonizer dum.
RTA dum.
Common currency dum.
GATT membership dum.
Communist country dum.
Observations
R2
8467
0.89
Note: Robust standard errors in parentheses. Constant and country
dummies not reported.
a
Significance at the 1% level.
b
Significance at the 5% level.
c
Significance at the 10% level.
We assume the following trade cost function:
X
ah zh :
ln tij ¼ ln dist ij þ
ð6Þ
h
The vector of observable bilateral linkages affecting trade costs includes variables measuring geographical, cultural and
historical proximity as well as trade policy variables:
Zij ¼ ½Contig ij ,Lang ij ,Colonyij ,ComColij ,Comcur ij ,RTAij ,GATT ij ,Cocoij ,
ð7Þ
where Contigij, Langij and Comcurij are dummies for countries sharing a common border, a common official language and a
common currency respectively. RTAij measures the existence of a regional trade agreement between countries i and j, and
GATTij equals 1 if both countries are members of the General Agreement on Tariffs and Trade (GATT). Colonyij and ComColij
are dummies equal to 1 for countries that have ever been in a colonial relationship and that share a common colonizer,
respectively. Finally, Cocoij equals 1 if i and j are communist countries.
Combining (5) and (6), we have:
P
X
yj
Pjs1 w expb1 ln distij þ h bh zh :
ð8Þ
Pi1s ¼
y
j
The b’s coefficients in (8) can be consistently estimated from (4) with country fixed effects (Feenstra, 2004; Anderson and
van Wincoop, 2003). We can then solve the vector of Pi1s using the system of N goods market equilibrium condition (8),
estimated coefficients from (4) and GDP shares.
An important issue here is to measure domestic trade costs, p
since
(8) ffiincludes all trading partners including the country
ffiffiffiffiffiffiffiffiffiffiffiffiffiffiffiffi
itself. Internal distance of country i is calculated as dii ¼ 0:67n area=P. Contigij, Langij and Colonyij are set at 1 for intranational trade. Eq. (5) is estimated on a cross-section of country pairs in 1960 using a Poisson quasi maximum estimator
(Santos Silva and Tenreyro, 2006). Table 9 reports the results.
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