Power-Sharing or Power-Dividing?
Walking out of the Maze.
Valentin Gold
Department of Politics and Management
University of Konstanz, Germany
[email protected]
Paper prepared for presentation at the ECPR Joint Sessions of Workshops,
St. Gallen, Switzerland, April 12-17, 2011
[-Paper Version RC1∗-]
10,801 words
Abstract
Post-conflict societies are constantly threatened by the danger of recurring violence.
Therefore, constitutional engineers propose specific institutional settings which help to
keep peace. Following Roeder and Rothchild (2005), the transition to a stable and peaceful democracy is divided into two phases: the initiation and consolidation phase. Whereas
in the first phase, power-sharing institutions are highly recommended, the second phase
of transition to democracy is characterized by a change to power-dividing institutions.
However, there are almost no empirical studies analyzing the effect of recurrent civil war
for both power-sharing and power-dividing institutions. In this paper, the effect of various institutions in combination with ethnic diversity on the risk of civil war onset is
investigated. The semi-proportional Cox-models yield support that several institutional
settings can indeed reduce the likelihood of recurring civil war – however, neither exclusively power-sharing nor power-dividing institutions. Additionally, a close examination of
the country’s ethnic structure is necessary.
Introduction
Post-conflict societies face a high risk of conflict renewal. Compared to countries that have
never experienced an internal armed conflict, the risk is almost double (Collier, Hoeffler, and
∗
An earlier version of this paper was presented at the 5th ECPR General Conference, Potsdam, Germany,
September 10-12, 2009
1
Soderbom 2008; Collier, Laporte, and Seawright 2008; see also Suhrke and Samset 2007). Between 1975 and 2005, 91 civil war onsets occurred in post-conflict societies whereas only 50
armed conflicts were counted in countries without a civil war background. The costs of these
conflicts were and are still high, not only in economic terms but also and in particular, in terms
of injury, disease and mortality rate. Having realized that these countries seem to need help,
more and more scholars and politicians propose specific institutional arrangements which – in
their point of view – will help to provide and keep peace.
However, scholars and politicians disagree about the best institutional setting (see e.g. the
debate for designing the South African electoral system between Lijphart 1991 and Horowitz
1991). Overall, two competing theories of constitutional engineering have emerged: PowerSharing and Power-Dividing. Supporters of the power-sharing theory favor either consociational
(Lijphart 1977, 1989, 1991, 1999, 2004, 2008; McGarry and O’Leary 2006a, 2006b; O’Leary
2005) or integrative (Horowitz 1985, 1990, 1991, 2004, 2006, 2008; Sisk 1996; Reilly 2001;
Wimmer et al. 2004) institutions, whereas supporters of the power-dividing theory (Roeder and
Rothchild 2005) in the long run favor the American model of democracy. In particular, Roeder
and Rothchild stress the sequential effect of a transition to democracy: While in the initiation
phase confidence-building measures, i.e. power-sharing institutions, are strongly required, the
consolidation phase is dominated by stability aspects, i.e. power-dividing institutions have to
be implemented. Following their argument, there should be a trade-off between the short-term
benefits and the long-term costs of having included all political relevant groups.
Whether the transition to democracy succeeds is not only a factor of the institutions to
be implemented but also a factor of the characteristics of the groups. In particular, ethnicity
is seen as one of the most decisive factors (Byman 2002; Horowitz 1985; Wimmer et al. 2004;
Wolff 2006). Constitutional changes in ethnically diverse societies can be observed twice as
often as in ethnically homogeneous countries (Elkins, Ginsburg, and Melton 2008) – on average
every eight years (Elkins and Sides 2007). Even if other social divisions are included in the
analyses, ethnicity is still an important factor (see Selway 2011)1 . In ethnically diverse societies,
the risk of conflict renewal even enforces. In some instances, political parties even create
terrorist organization (Danzell 2011). As the forms of ethnic (or other) division vary widely
across failed (or failing) states, there are no one-size-fits-all ready-made solutions. Hence,
the institutional engineering procedures should take the particular situation in a country into
1
For a more general critique towards the factor ethnicity, see Chandra (2006) and Gilley (2004); for a more
detailed critique, see Dunning and Harrison (2010).
2
account. Additionally, “the conditions that promote instability are [. . . ] particularly acute in
post-conflict situations in ethnically divided societies” (Roeder and Rothchild 2005, 127).
There are, however, few empirical studies concentrating exclusively on post-conflict societies
(e.g. Call and Cousens 2008; Collier, Hoeffler, and Soderbom 2008; Walter 2004). Several studies
evaluate the effect of institutions for delimited geographic regions (e.g. Lijphart 1996; Weidmann
and Ward 2010; Wilkinson 2004)2 or for specific institutions only, e.g. party system or electoral
system (see Anderson and Guillory 1997; Kuenzi and Lambright 2005). Few concentrate on
recurrent civil war (see e.g. Collier, Hoeffler, and Soderbom 2008; Senese and Quackenbush
2003; Walter 2004). And, finally, most studies do not include the ethnic structure of these
countries in their analyses (see e.g. Bratton and Chang 2006).
In this study, these factors are combined. The analysis of this study concentrates exclusively
on post-conflict societies with a focus on the following questions: Is conflict recurrence based
on ethnic diversity? Are there specific institutions which mitigate the effect of ethnic diversity?
What is the influence of these institutions on the effect of ethnic diversity?
The paper is organized as follows: After introducing the theories of constitutional engineering in detail, I determine the challenges to peace in post-conflict societies. The third section
includes the research design. The results of the analyses are given in section four. Section five
concludes.
Engineering Political Systems
The influence of consociationalism in the OECD countries demonstrates the impact of the theory. In Arend Lijpharts3 various publications he predicts and demonstrates the advantageous
effect of power-sharing institutions. If a state is characterized by broad-based or “grand” coalitions, minority vetoes, proportionality in allocation of civil service positions as well as public
funds and group autonomy, it is assumed that all major ethnopolitical groups can participate in
the policy process and therefore abstain from using violence (see Lijphart 2004). Even though
the theory emerged from empirical observations of the OECD countries, Lijphart explicitly
“advocate[s] consociational or power-sharing democracy [. . . ] for deeply divided societies” (Li2
3
Since Lijphart’s theoretical and analytical basis are the OECD-countries, most literature concentrates on this
region, see e.g. Lijphart 1981; 1989; 1990; 1994; 1994 or Vatter 2009.
Since Lijphart is definitely “the key pioneer of contemporary consociational thinking” (O’Leary 2003, 697),
I will refer to him as the founder of the theory. However, I am aware that other scholars have a major share in
developing the theory, e.g. as Lijpharts major critic Bogaards (1998, 2000, 2002, 2006) and Andeweg (2000).
See also refinements by O’Leary (2003, 2005) and McGarry and O’Leary (2005).
3
jphart 2004, 3; see also Lijphart 1977, 47–52). The basic characteristic is that “elites cooperate
after elections to form multiethnic coalitions and manage conflict” (Sisk 1996, 35, emph. in
original). Consequently, the procedures of interethnic cooperation have to be implemented before the election. Institutions which promote these effects are a parliamentary governmental
system, a proportional electoral system and a proportional representation in legislative and
executive bodies (compare Sisk 1996, 35).
On the other hand, there is the integrative power-sharing approach, which is also labeled
Centripetalism (see Reilly 2001). Contrary to the consociational approach, “parties [are] encouraged to create coalitions before elections [to create] [. . . ] broadly inclusive, but majoritarian
governments” (Sisk 1996, 35, emph. in original). The integrative approach varies basically with
respect to the design of the electoral system. Both Horowitz (2003, 2004, 2006) and Reilly
(2001, 2002a, 2006) prefer the Alternative Voting (AV) system. Due to the fact that countries,
in which the AV system is implemented – Australia, Fiji, and Papua New Guinea –, have been
peaceful for a long time, they are positive examples for the advantageous effect of the AV system
(see e.g. Reilly 1997, 1999, 2002b). Other institutions, which promote integrative power-sharing
are federalism and presidential systems in which the president is elected by “supermajority”
(Sisk 1996, 35).
Roeder and Rothchild (2005) criticize both approaches. The initiation phase of democratization is dominated by – often necessarily – power-sharing concessions to different ethnic
groups “to get leaders of different ethnic groups to come to the bargaining table [. . . ] simply
to initiate the transition to sustainable peace” (Roeder and Rothchild 2005, 2). Furthermore,
a “modicum of political stability [is] necessary to conduct elections to a constitutional assembly [. . . ]” (ibid, 2). However, implementing power-sharing institutions in the first phase may
cause – in the long run – gridlock, because every ethnopolitical group has the right to veto
important political decisions. Therefore, Roeder and Rothchild advocate in the second phase a
transition to the Madisonian model of democracy: strong civil liberties, multiple majorities and
checks and balances. The basic characteristic of the power-dividing approach is an “allocation
of power between government and civil society such that ’strong civil liberties [. . . ] take many
responsibilities out of the hand of government’, while those that are left there are distributed
’among separate, independent organs that represent alternative, cross-cutting majorities’, thus
’balanc[ing] one decisionmaking center against another so as to check each majority [. . . ] [f]or
the most important issues that divide ethnic groups, but must be decided by a government
common to all ethnic groups’” (Wolff 2007, 380, citing Roeder and Rothchild 2005, 15). But
4
still there is a selection problem, because some short-term designed institutions may not be
capable of being transformed into long-term consolidating institutions (Roeder and Rothchild
2005, 6). Rather, the two phases are linked, because the institutions of the initiation phase are
built to empower political actors which will later decide on the institutional long-term setting
and, secondly, often ethnopolitical groups only agree to new institutional rules in the first stage
if these rules endure “forever” (ibid, 14), and as Roeder and Rothchild further point out, even
if third parties ensure peace.
Each of the three principles of power-dividing theory – strong civil liberties, multiple majorities, checks and balances – has an advantageous effect on the avoidance of recurring violence:
expanding civil liberties and rights at the expense of government keeps key decisions in the
hand of individuals. This lowers the stakes in politics and “create[s] the most favorable institutional conditions for interests to become fluid and ’self-determining’ rather than fixed and
’pre-determining’ by government” (ibid, 16). Because the majority will divide into separate
interests, interethnic competition is being replaced by competition within the majority group
(ibid, 16). By empowering majorities in different organs of government, multiple majorities are
created, which will further “lead to identities that are multidimensional, situation- or issuespecific, and crosscutting rather than unidimensional, recurring, and cumulative” (ibid, 17).
Multiple majorities lead to bargaining situations in which “members of ethnic minorities will
be parts of political majorities on some issues and many members of any ethnic minority will be
members of political minorities on some issues” (ibid, 17). Hence, the ethnic dimension looses
relevance. Checks and balances – the third principle – controls for attempts to gain power at the
expense of competing agencies (ibid, 17; see also Choi 2010). Roeder and Rothchild conclude
that presidential systems are preferable to parliamentary systems, bicameral legislatives to unicameral legislatives, and independent judiciaries to weak judiciaries. Rather than designing
the institutions in a top-down perspective, the power-dividing approach “leads statesmen and
-women to begin at the bottom with elections to such diverse organs as hospital administration, school districts, and village or borough governments and build upward [. . . ] to construct
city governments, various crosscutting regional administrations, and diverse national governing
organs” (ibid, 18).
The influence of power-sharing and power-dividing per transition phase leads to two hypotheses:
H1 Power-sharing institutions reduce the probability of recurring violence during the initiation
phase of that country.
5
H2 Power-dividing institutions lower the risk of recurring civil war during that countries consolidation phase.
These two hypotheses are derived directly from the theory. There is, however, some empirical evidence of which institutional setting – power-sharing or power-dividing – is best in
post-conflict societies.
Institutional engineering in divided societies
In their 2008 published study, Schneider and Wiesehomeier evaluate the effect of different
institutions for various ethnic settings. In a first step, their analysis included all countries, in a
second step, only democratic countries. Their twofold analysis lends support to the proposition
that, first of all, democratic countries are more peaceful than autocratic ones. However, “[a]
fractionalized society seems to be the biggest challenge to a democratic country” (Schneider
and Wiesehomeier 2008, 194). Neither do polarized nor ethnic dominated societies face such
a challenge. There are, however, certain institutional settings which decrease the likelihood
for civil war onset, e.g. a parliamentary system is best in ethnically fractionalized societies.
Schneider and Wiesehomeier conclude that their results “lend support to the hopes of those
’constitutional engineers’ who advance certain institutions as a tool to alleviate social conflict”
(Ibid.). It is important to point out, however, that their analysis measures civil war onsets in
general and not only for post-conflict societies.
Most studies, however, do not differentiate between power-sharing and power-dividing institutions; e.g. Walter (2004) concludes that the greater the access to political participation in
general, the less likely recurring civil war (371). However, her measure of the dependent variable
– civil war onset – is based on the definition by the Correlates of War project, i.e. measuring an
onset if at least 1,000 battle-related deaths were counted. Albeit Hegre et al. (2001) and Collier,
Hoeffler, and Soderbom (2008) also do not concentrate exclusively on power-sharing or powerdividing institutions, their analysis reveals that for democracy, the effect on the risk of internal
armed conflict is inverted u-shaped. Both, completely autocratic and democratic countries face
less challenges in keeping peace; in transitional societies, the risk increases. Therefore, “international pressure for democracy should be justified by criteria other than peace-strengthening”
(470). Another major finding is the time-stationary effect: None of the variables that they
had included in their models have proven to be non-stationary – the effect over time remains
the same. By including fractionalization as a major independent variable in their analysis, the
6
effect on the (exponential) hazard rate is negative, i.e. the higher the degree of fractionalization,
the less is the likelihood of civil war recurrence.
Following a different approach of measuring power-sharing4 , Hartzell and Hoddie (2003,
2007), Jarstad and Sisk (2008) and Jarstad and Nilsson (2008) determine that the more powersharing dimensions, the lower the risk of recurring internal armed conflict. In total, four powersharing dimensions are included in their analyses: political, economic, territorial and military
power-sharing. Implementing one more power-sharing dimension, e.g. not only political but
also military power-sharing, the risk of civil war onset reduces by 53% (Hartzell and Hoddie
2003, 327). Territorial and military power-sharing have the greatest effects on the likelihood of
civil war onset. However, their measure of power-sharing is not sophisticated. Single changes
in the institutional setting which might have a significant effect on the risk of armed conflict are
not captured – the dummy variable remains the same. Additionally, their approach assumes
an equal effect of the power-sharing dimensions for autocratic and democratic countries.
Finally, a third empirical approach analyzes the effect for ethnopolitical dyads, including
ethnic diversity through ethnopolitical groups indirectly as a proxy for ethnic diversity. Roeder
(2005, 68) demonstrates the effect of power-dividing institutions by analyzing ethnopolitical
dyads consisting of leaders of governments versus ethnopoliticians from an ethnic group within
the state. The results yield support for the power-dividing theory, as “the signs of 23 of
28 coefficient estimates are as expected [. . . ]” (Roeder 2005, 74), versus five coefficients for
power-sharing. However, by using the Minorities at Risk dataset (see Gurr 2000), Roeder
“do[es] not include all minorities at risk; excluded are non-ethnic regional groupings [. . . ] or
religious subgroups within ethnic groups [. . . ]” (68, emph. in original). Additionally, time
is incorporated inappropriately: the independent variables and also the dependent variable
can only vary at the beginning of each 5-year time interval, changes during that interval are
integrated in the next quinquennium. This can lead to inappropriate results since the outcome –
violence – would not have taken place till the covariate had changed. Therefore, in my analysis,
time and ethnic diversity are incorporated in a more direct manner.
The most refined method of measurement of a country’s ethnic structure is based on three
manifestations of diversity – polarization, fractionalization and ethnic dominance. The effect
on the risk of civil war onset varies not only per manifestation, but also on the combinations
of these three forms of ethnic diversity.
4
Within this tradition, Linder and Bächtiger (2005) differentiate between vertical and horizontal power-sharing.
Though their measure comes closest to the original Lijphartian theory of consociationalism, they analyze the
effect on democratization.
7
Polarization is largest when two equally powerful groups πi compete for power. “The
original purpose of this index was to capture how far the distribution of the ethnic groups is
from the (1/2, 0, 0, . . . 0, 1/2) distribution (bipolar)” (Montalvo and Reynal-Querol 2005, 798).
The index was proposed by Reynal-Querol (2002) and is calculated as follows:
RQ = 4
N
X
πi2 (1 − πi )
i=1
The higher the index, the more polarized the country. Based on a formal model, it has been
shown that polarization decreases the risk of recurrent civil war (Esteban and Ray 2008). If
power-sharing or power-dividing institutions are implemented adequately, both groups5 are able
to participate in the policy process, either through multiethnic coalitions and cooperations at
the executive level or through changing majorities - therefore, reducing the risk.
Fractionalization is largest if every individual belongs to his/her own ethnic group. Hence,
the measure counts the number of ethnic groups within one country by asking the hypothetical
question to which ethnic group an individual belongs (compare Fearon 2003, 199), and secondly,
calculates the probability that two individuals selected by chance belong to different ethnic
groups. For a population with p1 , p2 , p3 , . . . , pn ethnic group shares, the index is (compare
Fearon 2003, 208)6
F ≡
n
X
p2i
I−1
Contrary to polarization, there is some evidence that the effect of fractionalization on the risk
of civil war recurrence is non-linear and inverted u-shaped (see e.g. Ellingsen 2000). The less
groups there are, the easier it is to incorporate all of them in the power-sharing or power-dividing
institutional setting; the more groups there are, the greater the collective-action problem is
when organizing violence. For an intermediate number of ethnic groups, however, civil war risk
is lowest. Yet, the risk for power-sharing institutions should be lower than for power-dividing
institutions since power-sharing is – at least in a short-term view – more concrete to ethnic
groups. Contrary to power-dividing, governmental power is shared directly between ethnic
groups.
Ethnic dominance is a dummy variable, which equals one if the largest group has an
5
6
The maximum of polarization is obtained for two groups, i.e. RQ = 1.
Most fractionalization indices are based on the Atlas Narodov Mira (for details see Posner 2004).
8
group share either between 45-90% (see Collier and Hoeffler 2004) or between 60-90% (see
Schneider and Wiesehomeier 2008). Defining ethnic dominance as a share of 45-90% is, however,
“unfortunate, as it contains cases of polarized societies” (Ibid., 186). For this reason, similar to
Schneider and Wiesehomeier, the more restricted measure is integrated in the analysis below.
Since majoritarian power-dividing institutions manifest the rule by an ethnic dominant group,
the risk of conflict recurrence should decrease compared to power-sharing regimes.
Ethnic Diversity, Political Institutions and Civil War
The effect of ethnic diversity may vary depending on the institutional structure of a country.
For example, it is reasonable to argue that polarization in majoritarian electoral systems has a
different effect than polarization in proportional representation electoral systems. If one party
wins everything and the other nothing – as is the case in pure majoritarian voting systems –
the incentives for the loosing party to use violence increase. Therefore, this section includes
the expected effects for different institutions7 .
Democracy Similar to Schneider and Wiesehomeier (2008), a negative effect on the risk of
civil war recurrence for democracy is assumed. The effect of all ethnic diversity variables should
decrease, compared to the risk in autocratic regimes. Both power-sharing and power-dividing
institutions should perform better in democratic systems. Additionally, ethnic dominance is assumed to have the strongest negative effect on the probability of recurring violence in autocratic
and democratic systems. The higher the dominant group’s share, the easier the repression of
the other group, even in democratic societies.
Governmental systems In parliamentary systems, the leaders of the government are dependent on forming coalitions, more than in presidential systems. Through sharing power
between more than one group, the effect of fractionalization and ethnic dominance in parliamentary governments on the risk of conflict renewal should decrease. The increasing effect of
polarization can be attributed to the vast stress of a bipolar society. The higher the polarization, the lower the common denominator, the higher the probability that conflicts are solved
outside the range of the legal institutional setting.
Electoral system Pure single-winner voting systems are assumed to have a manifold effect
on ethnic diversity: Whereas the effect for ethnic dominance should decrease the likelihood of
violence, polarization is assumed to have an increasing effect. A majoritarian voting system can
be used by an ethnic dominant group to repress the minority, not only within the institutions
7
These expectations are printet in table 4.
9
but also in other areas. For the ethnic minority, the costs of organizing a civil war are extremely
high. In polarized societies, however, the struggle for power should increase the risk of civil
war onset through a winner-takes-all system. Even though power may change from one ethnic
group to the other (because of regular elections), one ethnic group is always in power. The effect
of fractionalization is more complex and depends on other covariates, e.g. number of voting
districts, district magnitudes, electoral threshold and spatial concentration of ethnic groups.
For proportional representation (PR) electoral systems, whereas the expected influence of
fractionalization and ethnic dominance is negative, the expected effect of polarization is positive.
Due to the fact that votes proportionally represent each group’s vote share in the population,
all ethnopolitical groups are incorporated in the policy process. In highly polarized societies,
however, stakes in politics are still high in power-sharing regimes. Since the institutional voting
system is not able to compensate, conflict emerges. Additionally, for highly fractionalized countries, formation of government is more complicated due to the number of parties in parliament.
As a consequence, the probability of conflict recurrence increases.
Federalism Federalism is both, a measure for power-sharing and for power-dividing institutions. It is assumed that the effect of federalism combined with polarization, fractionalization
or ethnic dominance decreases the risk of conflict. If all ethnopolitical groups – regardless of
the ethnic division – have the right to decide on their own issues, the stakes in politics decrease.
However, forming federalistic states is limited to a medium number of ethnic groups. Therefore,
fractionalization decreases the risk only for societies with small and medium number of groups.
The higher fractionalization, the higher the risk of civil war onset. Overall, the effect should
be lowest for completely homogeneous societies.
Legislative design Broad based or “grand” coalitions – one of the four principles of consociationalism – are expected to lower the risk of recurring civil war (compare Lijphart 2004). The
higher the number of parties in the coalition, the more possibilities for each party to implement
their own policies. Each implemented policy is the result of a bargaining process between the
coalition parties. Therefore, ethnic diversity in combination with coalitions should lower the
likelihood of violence. Reserved legislative seats can hinder the executive repression able to
be exercised by an ethnic majority. In particular in combination with polarization or ethnic
dominance, the risk of civil war onset should decrease – given seats are reserved for opposition
groups. For polarization, reserved seats absorb the negative effect of other institutions, e.g.
majoritarian voting systems. Contrary to polarization and ethnic dominance, the effect of
fractionalization is assumed to increase the risk. The more ethnic groups there are, the higher
10
the percentage of reserved seats – if a proportional allocation of seats to groups is assumed.
Yet, this leads to an institutional structure in which elections are eclipsed.
Checks and Balances The division of a state in separate and independent branches controls the power within the state. Gaining power is only possible if other branches lose power
instantly (compare Roeder and Rothchild 2005, 17). In ethnically heterogenous countries, checks
and balances should decrease the effect of recurring civil war8 .
Initiation and Consolidation Phase Following Roeder and Rothchild (2005), powersharing institutions should in the long run increase the likelihood of civil war onset – whereas
power-dividing institutions decrease the likelihood. Empirically, the initiation phase ends after
5 years (Hartzell, Hoddie, and Rothchild 2001, 199).
Operationalization of dependent variable
The dependent variable, recurrent civil war onset, is taken from UCDP/PRIO Armed Conflict
Dataset (Gleditsch et al. 2002). Because only internal conflicts are of interest, intra- and internationalized minor and major conflicts (UCDP/PRIO coding 3 and 4) are coded. The event
variable equals 1 if a conflict resulted in at least 25 battle-related deaths within a given year,
0 otherwise. Subsequent years of conflict are excluded from the analysis.
This definition encompasses all conflicts within one country within the observation window (1975-2005). Therefore, a new conflict is not necessarily a continuation of the preceding
one – it also might be a new conflict with other parties entering combat. The underlying
general assumption, however, is that each armed conflict broke out due to imperfect political
institutions.
Operationalization of independent variables
Four out of eight independent variables are taken from the Institutions and Elections Project
(IAEP) Dataset (Regan and Clark 2009). The IAEP dataset contains information on constitutions and their change over time. As the IAEP dataset is a good starting point, to evaluate
the effect for various institutions, other datasets had to be merged, e.g. Database of Political
Institutions (Beck et al. 2001), Arthur Banks Cross-National Time-Series Data Archive (CNTS)
(Banks 1996) and Cheibub, Gandhi, and Vreeland (2010) revised dataset on democracy and
dictatorship.
8
Not only in heterogenous countries, also in homogenous countries.
11
Ethnic Division
For this study, the ethnic composition data is based on calculations by Alesina et al. (2003),
who used the Encyclopedia Britannica (2001) as their primary source. One of the reasons for
using Alesina et al. (2003) is the high data availability. Other datasets’ country coverage is
lower, e.g. Fearon (2003); Easterly and Levine (1997); Roeder (2001).
• Ethnic polarization Although a continuous measure of ethnic polarization is best, due
to collinearity problems with fractionalization (see Schneider and Wiesehomeier 2006),
a binary variable was taken. The data used in this analysis originates from Schneider
and Wiesehomeier (2008), who based their calculation on the formula of Reynal-Querol
(2002). Polarized societies are coded 1, 0 otherwise.
• Ethnic fractionalization Even though it would be more appropriate to use non-timestationary data on ethnic fractionalization9 , due to data availability, the measure by
Alesina et al. (2003) is taken for analysis. Ethnic fractionalization is measured continuously with the lower bound 0 and upper bound 1.
• Ethnic dominance Taken from Schneider and Wiesehomeier (2008), this dummy variable equals 1 if the majority’s group share is between 60 and 90%, 0 otherwise.
Institutional Variables
• Democracy To measure democratic systems, the coding of Cheibub, Gandhi, and Vreeland (2010) is taken. An autocratic regime (coding 0) is given, if the chief executive is not
elected, there are no legislative elections taking place, there is no more than one party in
the executive or if there is no alternation in power. Democratic regimes are coded 1.
• Parliamentarism This variable is 1 for parliamentary systems, 0 for presidential systems. The variable is taken from the Database of Political Institutions (Beck et al. 2001).
Parliamentary systems are characterized by countries in which legislatures elect the chief
executive. Transition years are coded as the regime that emerges in that year.
• Proportional representation systems The variable equals 1 if there is a proportional
electoral system, 0 otherwise. It is taken from Regan and Clark (2009).
9
For a discussion see Posner (2004).
12
• Majoritarian electoral system Regan and Clark (2009) code majoritarian systems
versus proportional representation and mixed electoral systems. The variable takes the
value 1 if the country-year is characterized by a majoritarian system, 0 otherwise.
• Federalism Federalism is 1 if the state structure is federalistic, 0 otherwise. It is based
on the measure by Regan and Clark (2009).
• Coalitions The coalition variable is taken from Banks (1996) and equals 1 if there was
a coalition government in the year under observation.
• Reserved seats If seats within the legislature are reserved for domestic ethnic or other
identity groups, the variables equals 1, 0 otherwise. The variable originates from Regan
and Clark (2009).
• Checks and Balances This variable reflects the number of different branches struggling
for power within a state. Basically, this measure counts the number of veto players. It
differentiates between formal and informal institutions. If, for example, multiple candidate elections are held but only one candidate wins, the actual incident is coded, rather
than the formal one. The variable is taken from the Database of Political Institutions
(Beck et al. 2001).
Table 1 demonstrates the changes of the institutional covariates for either all, only democratic, only post-conflict, and finally, post-conflict democratic country-years. It is shown that
these institutional covariates for which the constitution has to be changed, vary the less. For
the subset of country-years used in this analysis – all post-conflict democratic societies, table 5
in the appendix gives a more detailed overview of which variables had changed.
Table 1: Number of changes of independent variables
Total Democratic Post-conflict Post-Democratic
Parliamentarism
51
26
24
11
Prop.Voting System
31
19
12
7
Maj.Voting System
50
32
20
13
Federalism
72
29
41
19
Coalition
324
152
110
48
Reserved Seats
41
17
29
11
C&B
567
429
232
161
13
Control Variables
To evaluate the effect of various institutional structures, three control variables, which have
been proven in the previous study by Schneider and Wiesehomeier (2008) to have a significant
effect, are included in the analysis.
• GDP This variable measures the economic development of a country. The higher the
economic development measured in GDP, the lower the risk of recurring civil war (see
Collier and Hoeffler 1998). The variable originates from the United Nations Statistics
Division (2009). For adjusting the skewed distribution, the logarithm of GDP per capita
is used.
• Size of population To control for population size, the log of population is included in
the analyses. The variable originates from Banks (1996).
• Regime durability It is assumed that the longer a regime endures, the lower the risk
of recurring civil war. This variable is taken from the Polity IV project (Marshall and
Jaggers 2007) and counts the number of years since the last major regime transition.
Research Design
To evaluate the risk of civil war onset for various institutional structures in combination with
different measures of ethnic heterogeneity, a survival analysis is applied. The advantage of this
method is the explicit modeling of the factor “time”. By defining the dependent variable as
duration of a social phenomenon, it is possible to evaluate the history, present and the unknown
future of this phenomenon. The event-history analysis tries to capture the process responsible
for the duration: “In particular, we want to understand why units under consideration remain
for some period of time in one state and then change to another” (Bennett 1999, 257, emph.
in original). In reference to this analysis, all post-conflict countries between 1975 and 2005 are
analyzed. One problem in investigating only post-conflict nations, however, is the sparse data
availability. The event-history analysis determines the instantaneous transition rate from one
discrete state – peace – to another – civil war – by specifying the function of continuous time
and covariates. Using Cox (1972), the hazard rate for the ith individual is
hi (t) = h0 (t)exp(β 0 x)
14
where h0 (t) is the baseline hazard function and β 0 x = β1 x1i + β2 x2i + . . . + βk xki the effect
of the covariates (compare Box-Steffensmeier and Jones 2004, 48). Similar to Schneider and
Wiesehomeier (2008) the conditional risk-set model by Prentice, Williams, and Peterson (1981,
see also , ) is used. This allows ordered failure times. “This means [. . . ] that the risk set
at time t for the kth onset contains only those countries that have already experienced k − 1
onsets, [therefore the] model is stratified on the failure order, allowing for strata-specific baseline
hazards” (Schneider and Wiesehomeier 2008, 190).
To measure the effect of institutions over time, each institutional covariate is tested for its
non-proportional effects. If the effect is significant, the null hypotheses of zero slope has to be
rejected, i.e. these covariates effect changes over time (see Stata Corp LP 2007, 180-181).
As the inclusion of right-censored countries, i.e. countries which have not failed before 2005,
has an unbiasing effect on the equation, integrating left-censored countries is more complicated.
In most studies (see e.g. Schneider and Wiesehomeier 2008) left-censored countries are ignored.
Left-censored countries are those that have experienced an internal armed conflict but data of
political institutions is not available immediately right after the conflict has ended, i.e. the
conflict ended before 1974. However, ignoring these cases in my analysis would reduce the unit
of analysis – country-years – significantly, and, additionally, information would be lost. Some
countries either experience their first conflict at the right end of the observation window, i.e.
being peaceful for a long time before a civil war emerged, or experience no conflict at all. For
example, within the observation window (1975-2005), Cape Verde never experienced organized
violence, and Guinea-Bissau experienced the first civil war onset in 1999. If left-censored
country-years are excluded, the institutional covariates describing Cape Verdes peaceful history
will get lost. However, the information that, for any given time, a country was peaceful because
of its specific setting of covariates, is not missing at random. Therefore, left-censored countries
are included; country-years are only excluded in the analysis if, and only if, the first civil war
onset is within the first five years of the observation window. If civil war onsets occur later
than 1980, and data is available, then the country is included. Left-censored countries, however,
bias the semi-proportional Cox model. To correct that, the strata variable “left-censored” is
integrated into the analysis that allows the modeling of separate baseline hazards for each strata.
The stratifying variable is significant, as not only the test of equality of the survivor functions
across groups (compare Stata Corp LP 2007, 386), but also the observed versus predicted plot,
shows that for left-censored countries the effect of civil war onset is slightly underestimated
in the first half of the study (up to year 15). For non-left-censored countries, the effect is
15
overestimated in the first half of the study. Overall, these statistical tests demonstrate the
need for stratifying the analyses.
To determine the effect of institutions in heterogenous societies, interaction models are
included. Tables 2 and 3 show the results for the semi-proportional interaction Cox-models. All
models are estimated by using the Efron-method for ties10 . Robust standard errors are used.
Furthermore, only statistically significant time-varying covariates are kept. The regressions
are stratified by left-censored countries. β-Coefficients are reported; for interpretation of the
multiplicatory interaction model, the log-relative hazard is calculated, compare Cleves et al.
(2010, 186) and more general Brambor, Clark, and Golder (2006). Depending on the derivative
to be analyzed, the effect of the interactive log-relative hazard term can be interpreted in two
ways. Firstly, the marginal effect with respect to the constitutional variables can be evaluated.
As a consequence, the effect of these variables then depend on the values of the ethnic diversity
variables. Secondly, the derivative with respect to the three manifestations of ethnic diversity
can be analyzed. This implies that the marginal effect is not constant but changes with the
institutional covariate.
Results
Model 1 evaluates the hypothesis that the effect of ethnic diversity is reduced in democratic
countries. The probability of recurring civil war for presidential and parliamentary systems is
analyzed in model 2. This is one of the major differentiating characteristics of power-sharing and
power-dividing: power-sharing favors parliamentarism, power-dividing favors presidentialism.
For a transition from power-sharing to power-dividing institutions, the time-varying effect for
parliamentarism (and the appropriate interaction terms) should be negative, i.e. the effect
of parliamentarism on the hazard rate should increase with time and, analogous, decrease in
presidential systems. Models 3 and 4 calculate the effect for either proportional representation
or majoritarian electoral systems. In Model 5, the effect of federalism is analyzed. The 6th
model tests one of the major consociationalist principles: the effect of multi-party coalitions
on civil war onset. Measuring if a coalition government was in power, however, differs from
the principle formulated by the theory of consociationalism (see Lijphart 2008): broad-based or
“grand” coalitions have to be guaranteed by the constitution. The variable used in the analysis
is not a necessary, but sufficient proxy of guaranteed constitutional rights to form coalitions.
10
One advantage of handling tied failures with the Efron method is that the Efron method is a better approximation than the Breslow method (for details see Stata Corp LP 2007, 153).
16
In the 7th model, a proxy for proportional reservation of seats is taken to evaluate the effect
of ethnic diversity in combination with reserved seats. Contrary to the former variable, this
measure is a necessary, but not sufficient proxy for a reservation of seats to political groups,
because the variable coding is based on the specification of the formal constitution (see Regan
and Clark 2009). In model 8 the influence of the power-dividing principle “checks and balances”
is determined. Here, attention should be made on the time-varying influence of this variable: if
there is a positive relationship with time, then the hazard rate increases with time, i.e. checks
and balances are more important in later stages of transition.
First of all, in table 1, the results of Schneider and Wiesehomeier (2008, 193) are replicated.
Using partly different datasets and – more important – stratifying the analysis for left-censored
countries, overall, the results of model 1 confirm Schneider and Wiesehomeier’s analysis: Democratic regimes face less challenges with respect to civil war onset than autocratic regimes. In
completely homogeneous countries, the risk is reduced by 83%. As long as the degree of fractionalization is below 0.6, the effect remains significant – if polarization and ethnic dominance
are set to their mode values of zero.
Table 2: Effect of ethnic diversity and democracy
Variable
Coefficient
(Std. Err.)
Main effects
Polarization
-0.059
(0.301)
Fractionalization
0.560
(0.726)
Dominance
-0.011
(0.385)
Democracy
-1.809∗
(0.974)
Pol. × Democracy
0.558
(0.507)
∗
Frac. × Democracy
2.392
(1.308)
Dom. × Democracy 0.559
(0.453)
Population
0.450∗∗∗
(0.125)
∗∗∗
GDP
-0.277
(0.104)
Regime Dur
-0.044
(0.053)
Interaction with time
Regime Dur
0.014
(0.018)
N
Log-likelihood
χ2(11)
3500
-368.439
34.454
Significance levels:
∗ : 10%
∗∗ : 5%
∗ ∗ ∗ : 1%
Cell entries report coefficients. For the interaction-terms, cell entries report marginal effects. Robust standard errors in parentheses. The regression is stratified by left-censored countries, see text
for details.
17
In autocratic regimes, the effect of ethnic diversity is not significant. However, in democratic
regimes, fractionalization increases the risk of civil war onset. The marginal effect of both
polarization and democracy as well as ethnic dominance and democracy remain insignificant.
This, however, demonstrates empirically the importance of analyzing only democratic (postconflict) countries: The effect of ethnic diversity differs in autocratic and democratic countries.
Also, the test of equality of the survivor functions for democratic and autocratic countries
is significant. Therefore, the next models concentrate exclusively on democratic post-conflict
societies.
The effect of the three control variables is as expected: the more people live within a country,
the more challenging it is to have a peaceful constitutional building process. Furthermore, the
higher the level of economic development is, the lower the risk of recurring civil war onset.
And, finally, the effect of time since major regime transition is insignificant.
For the first model, I also checked for non-linear quadratic terms of fractionalization and
its interaction terms. Here, the results of the first model are confirmed: democracy reduces
the risk of recurrent armed conflict in homogeneous societies11 . Even though, for almost all
combinations of ethnic diversity the effect is insignificant, an inverted u-shaped relationship can
be seen: both, low and high fractionalization reduces the risk of civil war onset; fractionalization
at a medium level increases the likelihood of civil war. Since it has been shown that – for this
and all other models – the quadratic term of fractionalization is only significant in rare situations
and, additionally and more important, the effects of the non-linear model equal the effects of
the linear model, I refrain from presenting the results.
Table 3 shows that almost no non-stationary time effect exists. The only significant time
effect is the size of the population and GDP in model 7 (see discussion below). The effect
of institutions, therefore, remain constant over time. Neither do power-sharing institutions
increase the risk in later stages of transition nor do power-dividing institutions lower the risk
of civil war onset during the consolidation phase of a country. Therefore, both hypotheses 1
and 2 have proven statistically wrong.
As it can be seen from model 2, for completely homogeneous countries, the effect of parliamentary democracies on the risk of civil war onset does not differentiate statistically from the
effect of presidential democracies. However, as fractionalization increases, countries characterized by a polarized society and a dominant ethnic group are more likely to experience internal
war onset – the marginal effect for these countries is significant above the fractionalization value
11
Up to a degree of 0.2 for fractionalization.
18
19
Model 3
# PR
[rh]
1.251 ∗
(0.643)
0.686
(1.373)
1.064
(0.823)
3.846 ∗∗
(1.704)
7.071 ∗∗∗
(2.457)
-12.069 ∗∗
(5.317)
-6.190 ∗∗∗
(1.594)
0.947 ∗∗
(0.429)
-0.695 ∗∗
(0.321)
0.020 ∗∗
(0.008)
542
-41.547
40.302
Model 4 Model 5
Model 6
# Maj.
# Fed.
# Coalition
[rh]
[rh]
[rh]
[t]
∗
∗∗∗
∗∗∗
2.372
2.690
23.223
(1.255)
(0.983)
(1.027)
-2.011
-2.823
-2.134
(3.117)
(2.401)
(4.140)
-0.622
-1.318
-0.822 ∗∗∗
(0.767)
(0.934)
(1.461)
∗∗
-0.913
-3.316
20.805 ∗∗∗
(1.560)
(1.657)
(2.981)
1.210
-2.254
1.924
(0.806)
(1.440)
(.)
∗
0.491
6.235
-2.571
(1.398)
(3.456)
(2.571)
∗∗
-0.132
2.832
-0.175
(1.294)
(1.231)
(0.757)
0.720 ∗∗
0.401
0.566
(0.351)
(0.385)
(0.406)
-0.450
-0.248
0.820
-0.448
(0.292)
(0.290)
(0.881)
(0.315)
0.018 ∗
0.014 ∗
0.007
(0.010)
(0.008)
(0.14)
542
542
508
-45.763
-43.51
-36.043
28.749
33.177
3600.325
Model 7
# res. Seats
[rh]
[t]
∗∗
2.240
(0.931)
-1.939
(2.433)
-0.316
(0.689)
-2.073
(1.987)
3.121 ∗∗
(1.226)
-0.252
(2.800)
-0.383
(1.533)
-3.707 ∗∗ 1.735 ∗∗
(1.781)
(0.803)
2.561 ∗∗ -1.172 ∗∗
(1.045)
(0.480)
0.025 ∗∗
(0.011)
538
-41.29
38.77
Model 8
# Checks & B.
[rh]
[t]
0.767
-0.293
(5.402)
(1.259)
-5.675
(4.178)
-1.987
(1.533)
-1.591 ∗∗
(0.628)
0.524
(0.846)
1.617
(1.210)
0.575
(0.384)
0.793 ∗∗
(0.387)
-0.495
(0.313)
0.016
(0.008)
539
-43.851
32.738
Cell entries report coefficients. For the interaction-terms, cell entries report marginal effects (exception: model 8, see below). Robust standard
errors in parentheses. Row [t] denotes the results of the interaction with time ln(t). The regressions are stratified by left-censored countries, see
text for details.
The number sign (#) is a placeholder for the appropriate independent variable which can be found in the heading. For the second model, #
indicates parliamentarism (and its interactions), for the third model # has to be replaced by PR voting systems, same for models 4 - 8. The
variables are coded binary with the only exception in model 8: checks and balance; see text for details.
Significance levels: ∗ : 10%
∗∗ : 5%
∗ ∗ ∗ : 1%
N
Log-Likelihood
χ2(#)
Regime Dur
GDP
Population
Dom. × #
Frac. × #
Pol. × #
#
Dominance
Fractionalization
Variables
Polarization
Model 2
# Parl.
[rh]
1.566 ∗∗∗
(0.607)
-1.094
(1.919)
-0.603
(0.764)
-1.515
(1.549)
-1.567
(2.702)
5.781
(4.384)
2.570 ∗∗
(1.293)
0.854 ∗∗
(0.347)
-0.557 ∗∗
(0.248)
0.020 ∗∗∗
(0.007)
547
-45.034
31.495
Table 3: Results of the semi-proportional Cox-models with the dependent variable recurrent civil war onset
of 0.3. In a polarized country with a moderate number of ethnic groups (fractionalization of 0.4)
and one of them dominant, e.g. Sri Lanka, the risk of civil war onset increases by the factor
2.5.
Ethnic diversity has two significant effects: in presidential systems, the hazard rate of
civil war onset increases for polarized countries. Additionally, the marginal effect of ethnic
dominance and parliamentarism demonstrates an increase in the risk of civil war onset: a
dominant ethnic group is a major challenge to parliamentary systems.
The implications of model 2 are, firstly, to avoid implementing presidential systems in
polarized societies, and secondly, to avoid parliamentarism in countries, in which one dominant
ethnic group is existent. However, the marginal effect for parliamentarism and polarization
does not significantly reduce the hazard rate; similar to presidentialism and ethnic dominance.
In model 3, if ethnic diversity is absent, a proportional representation voting system increases the likelihood of civil war onset by a factor of about 2.5. This factor even increases
for polarized societies. However, this effect is significantly mitigated by the other two ethnic
variables: both, a higher degree of fractionalization and the existence of a dominant ethnic
group decreases the likelihood of civil war onset. Figure 1 demonstrates these effects.
Figure 1: Marginal effects of PR voting systems on the risk of civil war onset
as ethnic diversity change
6
4
2
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
*
0
-2
-4
-6
-16 -14 -12 -10 -8
Marginal Effect of PR voting system
8
10
Dependent Variable: Risk of Civil War Onset
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
*
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
* *
*
0
.1
.2
.3
.4
.5
.6
.7
.8
.9
pol.
dom.
X
–
–
X
–
X
–
X
1
Degree of Fractionalization
* indicate significance at the 10%-level
These effects are confirmed if one analyzes the marginal effects for the diversity variables:
20
Polarization represents the greatest threat to democratic countries. This effect is even increased in proportional representational voting systems. Conversely, the effect of proportional
representational systems reduces for fractionalization and ethnic dominance.
Therefore, proportional representation voting systems should be implemented in fractionalized and ethnic dominated societies. Polarized societies, however, can’t be pacified by this
voting system. The questions remains, however, if majoritarian voting systems are able to
mitigate the effect of the hazard ratio of recurrent civil war in polarized societies.
The analysis of model 4 confirms the results of the proportional representation model.
Majoritarian voting systems – contrary to PR systems – mitigate the likelihood of civil war
onset in polarized countries; compared to model 3, the effect of the other two interaction effects
has a reversed effect. However, most coefficients and marginal effects are insignificant. The
same interpretation applies for the marginal effect of majoritarian voting systems: The effect
itself is negative and decreases for polarized societies, whereas fractionalization and ethnic
dominance have an increasing effect – though these effects are insignificant.
Therefore, the implications can be drawn that, given a polarized society, majoritarian voting
systems are best; given a highly fractionalized society which is additionally characterized by a
dominant ethnic group, proportional representation systems are best.
In a combined model of proportional representation and majoritarian voting systems the
results of model 3 and 4 are confirmed12 . The effect of PR voting systems increase with polarization and decrease with fractionalization and ethnic dominance. If fractionalization exceeds
a degree of 0.2, the marginal effects are significant for various combinations of ethnic diversity. On the other hand, the effect of majoritarian voting systems in combination with ethnic
diversity is reversed: polarization decreases the likelihood of civil war onset whereas fractionalization increases the effect. Contrary to model 4, ethnic dominance has a reducing effect. In
comparison to model 4, the same insignificance pattern is found in this model.
The effect of federalism is shown in model 5 and figure 2: it is lowest for completely
homogeneous countries. Both, more ethnic groups and a dominant ethnic group increase the
effect of federalism; for polarization, the likelihood of recurrent armed conflict reduces. These
results are as expected: the more groups there are, the higher the hazard rate of civil war
onset. Additionally, time is stationary, i.e. the effect of federalism is not changing over time.
Consequently, unitaristic states should be preferred in highly fractionalized and dominated
societies, and federalism in polarized societies.
12
The model can be found in the webappendix.
21
Figure 2: Marginal effects of federalism on the risk of civil war onset as ethnic
diversity change
6
Dependent Variable: Risk of Civil War Onset
–
X
X
–
X
–
X
–
-2
0
2
dom.
* *
* *
* *
*
* *
* *
* *
*
* *
* *
* *
*
* *
* *
* *
*
* *
* *
* *
* *
*
*
* *
-6
-4
Marginal Effect of Federalism
4
* *
* *
* *
*
* *
* *
* *
*
*
* *
* *
* *
* *
* *
* *
*
* *
pol.
0
.1
.2
.3
.4
.5
.6
.7
.8
.9
1
Degree of Fractionalization
* indicate significance at the 10%-level
As it is demonstrated in model 6, coalition governments represent a major threat to postconflict societies. However, this effect is mitigated by polarization and fractionalization. Due
to collinearity with the dead and censor variable, the standard error of the interaction term
polarization and coalition government cannot be calculated. Nevertheless, the effects of ethnic
diversity remain the same. For a non-polarized society, the effect of the first-difference term
of coalition governments is significant positive for all values of fractionalization – no matter if
an ethnic dominant group is existent – even though the effect is mitigated by higher degrees of
fractionalization. Additionally I checked, whether parliamentarism acts like a trigger, and indeed, the hazard rate decreases for parliamentary countries. However, this effect is insignificant
and varies per ethnic constellation13 .
It is important to note that coalition governments are only a necessary condition for a
constitutional provision to build multi-ethnic coalitions. As a consequence, this explains the
unexpected low effect of coalition governments versus single-party governments in polarized
societies: The raw effect of coalition governments is negative, i.e. a coalition government reduces the risk of civil war onset. The high increasing value for polarization in single-party
governments exceeds the decreasing effect of coalition-governments. However, no implications
13
Due to collinearity with the failure and censoring variable (year 2005), the standard errors of most triple
interaction terms could not be calculated.
22
Figure 3: Marginal effects of Checks & Balances on the risk of civil war onset
as ethnic diversity change
dom.
X
X
–
X
X
–
–
–
-1.5
-1
-.5
0
.5
1
* *
* *
* *
* *
*
*
* *
* *
* *
pol.
*
* *
* *
* *
*
*
* *
* *
* *
-2
Marginal Effect of Checks and Balances
1.5
Dependent Variable: Risk of Civil War Onset
0
.1
.2
.3
.4
.5
.6
.7
.8
.9
1
Degree of Fractionalization
* indicate significance at the 10%-level
can be drawn whether constitutions should include guaranteed provisions to form multi-ethnic
or grand coalitions after elections.
In model 7, the positive influence of polarization on the risk of civil war onset is confirmed.
This effect is even increased in countries in which legislative seats or districts are reserved for any
domestic identity group. The marginal effect of both fractionalization and ethnic dominance
is insignificant. Also, the overall marginal effect of reserved seats is insignificant. Contrary
to the former models, the size of the population decreases the risk of civil war onset. As the
significant interaction term with time demonstrates, however, this effect reverses over time.
As time passes, the size of the population looses its decreasing effect – if seats are reserved.
Additionally, the effect of the economic level is also reversed contrary to the other models:
the wealthier a nation, the greater the likelihood of an internal war. However, as time passes,
economic wealth looses its increasing marginal effect.
Finally, in model 8 the effects for Checks and Balances are shown. Since Checks and
Balances are not coded binary, the effects shown are raw coefficients and not marginal effects.
The marginal effects for Checks and Balances for various combinations of ethnic diversity are
demonstrated in figure 3. In a country with no ethnic division, the marginal effect is negative,
i.e. it is less likely that an internal war breaks out. Surprisingly, as the degree of fractionalization
23
increases, the more likely is civil war onset. Therefore, Checks and Balances should only be
implemented in countries with almost no ethnic division.
Conclusion
Table 4 lists marginal effects, coefficients and expectations of all institutional models. One
third of the marginal effects are significant. In 5 of 7 models, the marginal effect of polarization
is reversed to the marginal effects of fractionalization and ethnic dominance. This clearly
demonstrates that polarized societies have to be handled different than fractionalized and ethnic
dominated countries. Whenever a society is characterized by two equally sized groups, finding
a peaceful democratic institutional setting is difficult. Polarization faces the highest threat to
all institutional settings. Some institutions, e.g. proportional representation voting systems,
even increase this threat.
Table 4: Overall-results and expectations on the risk of recurrent civil war
Interaction
Parl. × Pol.
Parl. × Frac.
Parl. × Dom.
PR × Pol.
PR × Frac.
PR × Dom.
Maj. × Pol.
Maj. × Frac.
Maj. × Dom.
Fed. × Pol.
Fed. × Frac.
Fed. × Dom.
Coal. × Pol.
Coal. × Frac.
Coal. × Dom.
rSeats × Pol.
rSeats × Frac.
rSeats × Dom.
C&B × Pol.
C&B × Frac.
C&B × Dom.
Marginal Effect
−
+
+∗∗
+∗∗∗
−∗∗
−∗∗∗
+
+
−
−
+∗
+∗∗
+
−
−
+∗∗
−
−
+
+
+
Coefficient
−
+
+∗∗
+∗∗
−∗∗
−∗∗∗
−
+
+
−
+∗
+∗∗
−
−
+
+
+
−
+
+
+
Expectation
+
−
−
+
±
−
+
−
−
−
+
−
−
−
−
−
+
−
−
−
−
Significance levels:
∗ : 10%
∗∗ : 5%
∗ ∗ ∗ : 1%
Note that expectations refer to the sign of the coefficient, not to the marginal effect.
However, none of the interaction effects have proven non-stationary. Therefore, no change
24
over time from power-sharing to power-dividing institutions – or vice versa – could be determined. Once the institutions are implemented, the (marginal) effects of the institutional
variables remain the same. Hence, Roeder and Rothchild’s (2005) hypothesis of a sequential
effect has been proven wrong in this study.
However, several steps that could improve the analysis (and the empirical answer on nonstationary effects) cannot be made due to a small-N and due to missing data-availability. First
of all, the time-frame should be expanded. As the observations window ends 2005, several
countries made further steps to democracy by introducing new constitutional rules, e.g. in
Africa: DRCongo, Zimbabwe and Sudan. Additionally, several efforts should be made to collect
and improve data. Data is in most cases not missing at random, therefore these cases cannot
be ignored. Third, various institutional variables should be interacted with other institutional
variables, i.e. determining the effect of triple (or even higher) interaction effects. There is
reason to assume that the marginal effect of e.g. federalism and proportional representation
voting systems differs from the effect of federalism in combination with a majoritarian voting
system. However, due to a small-N, these interaction effects could not be incorporated in the
analysis.
Overall, there is no clear answer to the question, if either power-sharing or power-dividing
institutions decrease the likelihood of recurrent civil war. The effects vary per ethnic setting,
i.e. there are no one-size-fits-all ready-made solutions available. Rather a bunch of variables
has to be included in the constitutional engineering process. This analysis demonstrates that
especially ethnicity is one of the key variables in the process of designing institutions. However,
this study also proves that by carefully designing institutions, the risk of recurrent civil war
can be reduced.
25
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Appendix
Table 5: Changes of major independent variables for all post-conflict democratic societies
Country
Argentina
Argentina
Argentina
Argentina
Argentina
Argentina
Argentina
Bangladesh
Bangladesh
Bangladesh
Bolivia
Burundi
Burundi
Chile
Comoros
Comoros
Congo
Costa Rica
Costa Rica
Costa Rica
Croatia
Croatia
Croatia
Croatia
Croatia
Dominican Republic
Dominican Republic
El Salvador
El Salvador
France
France
Ghana
Ghana
Ghana
Greece
Greece
Greece
Greece
Greece
Greece
Greece
Guatemala
Guatemala
India
Indonesia
Madagascar
Madagascar
Mali
Mali
Mali
Moldova
Nepal
Year
1959
1964
1974
1984
1988
1990
1998
1997
2003
2006
1996
1995
2006
1993
1994
1995
1996
1950
1959
1963
1996
2001
2002
2003
2004
1967
1996
1989
1996
1992
2005
1970
1980
1997
1953
1966
1975
1988
1990
1994
1997
1967
1988
1990
2003
1996
2000
1993
1994
1998
2002
1992
Parl.
PR
Maj.
−
+
Federalism
Coalition
+
+
+
+
−
+
−
−
res. Seats
−
+
+
+
+
+
−
+
+
+
−
+
+
−
−
+
−
+
+
−
−
+
+
−
+
+
+
+
−
+
−
+
+
−
+
−
+
+
−
−
−
+
+
+
−
−
−
34
+
−
+
+
Table 5: (continued)
Country
Nepal
Nepal
Nepal
Nicaragua
Nicaragua
Nigeria
Pakistan
Pakistan
Pakistan
Panama
Paraguay
Paraguay
Peru
Peru
Peru
Philippines
Philippines
Philippines
Romania
Romania
Romania
Senegal
Sri Lanka
Sri Lanka
Sudan
Suriname
Suriname
Suriname
Thailand
Thailand
Trinidad and Tobago
Uganda
Uganda
United Kingdom
Uruguay
Uruguay
Uruguay
Venezuela
Venezuela
Note that a change of
Year Parl. PR Maj. Federalism Coalition
res. Seats
1997
−
2000
+
2009
+
1986
+
1992
−
1982
+
1974
−
1990
+
1992
−
+
1996
−
+
1994
+
1999
−
1981
−
+
1986
+
2002
−
1988
−
1989
+
1992
−
−
1991
+
+
1992
+
−
1993
−
2003
−
1974
+
−
+
1990
−
1987
−
−
1990
+
1992
−
1993
+
1981
+
1999
−
1996
−
1981
+
1982
+
2000
+
1986
+
1987
+
1990
−
1984
+
2001
+
the dummy−variables from 0 to 1 is indicated by ”+”; a change from 1 to 0 ”-”.
35