The determinants of election to the United Nations Security Council

Bueno de Mesquita and Smith (2010) briefly analyze the determinants of UNSC membership with a probit model, but the authors focus on the effects of membership. Lim and Vreeland (2013) use a logit model conditioned on year to examine the election of Asian countries. Scharioth (2010) analyzes the election of Western European countries to various UN committees. Part of the work here is based on our earlier working paper (Dreher and Vreeland 2009). Two other working papers on the election of UNSC members that have been presented at conferences include Iwanami (2012) and Schmitz and Schwarze (2012). Thorough qualitative accounts of the selection of specific UNSC members have been published, e.g., Malone (1998, 2000), and Jayakumar (2011).

Strictly speaking, US military aid does not count as official overseas development assistance, according to the Organisation for Economic Co-operation and Development (OECD).

Some background for this section comes from Security Council Report, an independent not-for-profit organization: http://​www.​securitycouncilr​eport.​org. We also draw on Luck (2006).

The “others” in the modern-day WEOG include descendent countries of Western Europe, mainly from the British Commonwealth: Australia, Canada, and New Zealand. The United States also caucuses with this group, as do Turkey and, more recently, Israel (see, e.g., Security Council Report 2011). Before 1966, there were only six elected UNSC members. See Daws (1999) for the development of the UN regional groups.

The Eastern European term begins in even years. The two WEOG terms begin in odd years. The GRULAC and ASIA stagger their two terms; the UNGA elects one from each of these groups every year. Africa’s three terms are also staggered: two begin in even years and one in odd years. The Arab term (shared between Asia and Africa) begins in even years.

We know from the UNGA’s minutes that group chairmen stand up in sequence before the vote and announce the group’s candidates. The chairman position rotates among the region members, and terms last one month. See various issues of the Journal of the United Nations for details on specific elections (http://​www.​un.​org/​en/​documents/​journal.​asp, accessed 5 April 2012).

Sometimes countries announce their intention to run years in advance. Other times they do so much later, even in the midst of elections. The timing of such announcements appears idiosyncratic, and data are not available.

For the 36 election-years (1970–2005) we analyze, the WEOG is the most competitive group, with nine contested elections, and EE is the least competitive, with just five. As we detail further in Footnote 23, we define an election as “contested” if an additional candidate receives ten votes or more. Using this threshold, there are a total of 36 “contested” elections out of 180 total elections, or 20 %.

Africa is the only region for which we have found explicit rules, codified by the African Union in their “Rules of Procedure of the Ministerial Committee on Candidatures within the International System—Doc. EX.CL/213 (VIII).” See African Union (2006: 8).

North Africa and Central Africa rotate one seat every two years; Western Africa has one seat every two years; and Eastern Africa and Southern Africa rotate one seat every two years. See Security Council Report (2011: 6).

According to the Security Council Report (2009: 6), such queue-jumping occurred three times in the sample period: Ghana queue-jumped Liberia in 1985, and Nigeria queue-jumped Niger in 1977 and Guinea-Bissau in 1993.

Bashir and Lim (2013) challenge this assumption.

On the association of democracy with openness, see Hollyer et al. (2011). On the association with justice see Dowding et al. (2004). On the general proclivity of democracies to peace, see Russett and Oneal (2001). For a contrasting view, see Ferejohn and Rosenbluth (2008).

IMF programs themselves come in cycles (Conway 2007). Omitting participation in IMF programs might thus bias our results in favor of finding a turn-taking norm. A substantial literature argues that IMF and World Bank loans might be given for political-economic reasons rather than need (e.g., Kilby 2009, 2013; Reynaud and Vauday 2009; Stone 2002). As for bilateral foreign aid, we limit our attention to the US role for two reasons: (1) its prominent place—both in quantitative magnitude and in the literature and (2) parsimony. If we include foreign aid from all potential countries, degrees of freedom become low in certain regions. Preliminary analyses of foreign aid patterns from other OECD countries did not reveal any statistically significant correlation with UNSC election. We suggest that more in-depth analyses—for example Japan’s use of foreign aid to win favor—be explored in country- or region-specific studies.

We use GNI/capita, as opposed to the more common GDP/capita, as it is the income measure used by the UN in the computation of member state contributions to the General and Peacekeeping budgets. We also follow the UN’s methodology in using US$ exchange rate estimates of GNI. International, rather than domestic, purchasing power is more relevant in this context.

Because of substantial overlap in membership between G77 and NAM, indicator variables for membership in each cannot be included in the same regression equation. Instead we create three separate indicator variables: one for countries that are members of both groupings, and one for countries that are members only of NAM or only of G77, respectively.

Since it impacts foreign aid, UNSC membership may be a channel by which colonial history affects development. See Iyer (2010) and Bruhn and Gallego (2012).

We calculate the percentage of the region according to the number of the countries in the region—minus the country in question—sharing the same ideology (left, center, or right). The variable is coded zero for non-ideological governments. See Beck et al. (1999) for the coding of ideology.

Using the model, which we present in the next section, we tested several possible measures of a turn-taking norm against a benchmark of perfect turn-taking. In a given year, let t _i denote the number of years since country C _ij was last elected to the UNSC (or since it entered the UN, if no such instance), t denote the mean of t _i , and η denote the number of countries, excluding C _ij , eligible for election. The measures we considered were: (1) t _i ; (2) t _i /η; (3) t _i −η; (4) \(\mathbf{1}_{\{t_{i}>\bar{t}\}}\); and (5) \((t_{i}-\bar{t})\mathbf{1}_{\{t_{i}>\bar{t}\}}\), where 1 _{A} is the function taking the value 1 if condition A is true and 0 otherwise. We found the second of these measures to be best suited for capturing turn-taking effects.

Our analysis accounts for the creation of new nations and the disappearance of existing ones. These events change the sample size, which complicates the calculation of the marginal effects. We discuss this issue in depth below.

UNSC membership data are found on its official website (http://​www.​un.​org/​Docs/​sc).

We are grateful to an anonymous reviewer for these two possible extensions.

We compute α _jt using Costa Rica (2005), which contains full UNGA voting records for all UNSC elections prior to 2004. Voting records for 2004 onwards come from the relevant UNGA minutes. Costa Rica (2005) does not explicitly identify the “chairman’s list” countries. In the overwhelming majority of elections, the patterns of voting in the UNGA clearly identify the “chairman’s list” countries (who garner large numbers of votes) from countries that merely are recipients of votes cast in protest or error (who garner only one or two votes). In a small number of cases, the voting patterns identify the “chairman’s list” countries less clearly, as a country garners an intermediate number of votes between five and 15. In these cases we identify the set of “chairman’s list” countries as those that received ten or more votes. Our main results are, however, robust to lowering the threshold down to three votes (thereby counting more elections as contested). Obviously as we employ higher thresholds than ten, there are fewer and fewer elections counted as contested and eventually the model does not converge.

In the sample period 68 countries joined the UN, and four (Czechoslovakia, East Germany, Yemen Arab Republic, and Yugoslavia) left. Table 2 provides further details.

Elections are not independent across time, however. Each year’s election depends on the outcome of the previous year’s election in a recursive manner, owing to the evolution of E _t .

These distributional assumptions are strong but necessary to retain the conditional logit form. Also, when estimating the final likelihood in (6), we can allow for the possibility of within-group clustering. Because we model the probability of choosing C _ij in year t as conditional on the number of eligible countries in year t, our model, like the original conditional logit, implicitly addresses fixed effects for year. For an approach that relaxes our distributional assumptions at some conceptual and computational cost, see Hendel (1999).

The variables with missing values are: United States and Russia voting in the UNGA; debt service; shared regional ideology; control of corruption; and IMF program participation.

Although Voeten’s (2000) analysis suggests much subtler changes between the two periods.

For more on the Mexican case, see, for example, Serrano and Kenny (2006: 298–314). We are grateful to Diego Dewar for this suggestion.

As in other contexts, we are unable to adjust the standard errors for the effective degrees of freedom used by the model selection procedure itself. As such, it is appropriate to urge caution in the interpretation of findings on the margin of statistical significance at conventional levels. We note the necessity of such model selection, however, given the weak steer provided by theory, and the number of potential explanatory variables.

We do not include a separate Cold War intercept because the conditional logit model has the property that any variable that takes the same value for every country in a group in a particular year (like the Cold War indicator) simply cancels out of the numerator and denominator (see (4) and Footnote 26 above).

The regional country-specific effects we allow for are (by region), Africa: Benin, Guinea, Madagascar, Malawi, South Africa, Zimbabwe; Asia: India, Japan, Nepal, Philippines, Saudi Arabia; EE: Bulgaria; the GRULAC: Costa Rica, Mexico, Panama; the WEOG: Austria, Belgium, Switzerland. We allow for a global country-specific effect for Australia, Austria, Burkina Faso, Egypt, Greece, Madagascar, Romania and Slovakia.

These are available in the replication materials.

The estimates for the UNGA in Tables 3a, 3b seem of a different order of magnitude than the estimates in the regional groups. This can be explained with reference to (1), which weights UNGA preferences by α _jt , and group preferences by (1−α _jt ) in the composite utility function. Even for election years with non-zero values of α _jt , that coefficient typically is close to zero; E(α _jt |α _jt ≠0)=0.039, so the apparently large UNGA effects we estimate are offset by the very low weight UNGA preferences receive in the composite preference.

We calculate elasticity and marginal effect estimates for 2005, the final year of our sample, using (4). We evaluate these using the mi predict command in Stata 12, at the group-specific means \(\bar{\mathbf{x}}_{jt}\). Different estimates apply to “clean slate” and “contested” elections. The former are evaluated at α _jt =0, and the latter at E _j (α _jt |α _jt ≠0). We find negligible differences between these estimates, however, so we do not report each separately. Estimates also vary according to n _jt : we report estimates for n _jt =1, but in group-years with n _jt =2, a different estimate based on (5) does apply in practice. Last, the estimates vary across years owing to the evolution of the eligible set. We have evaluated the estimates for 2005 under different assumed eligibility conditions, and find this source of variation to be of minor proportions.

The former British colonies in the WEOG are Ireland (elected twice) and Malta (elected once).

Note that Potrafke (2009) finds that government ideology affects UNGA voting behavior.

We do not control for OIC in Asia due to collinearity with the Muslim variable. When we do include them together, neither variable is statistically significant.