More politicians, more corruption: evidence from Swedish municipalities

Abstract

In the literature on political economy and public choice, it is typically assumed that government size correlates positively with public corruption. The empirical literature, however, is inconclusive, owing to both measurement problems and endogeneity. This paper creates a corruption index based on original data from a survey covering top politicians and civil servants in all Swedish municipalities. The effect of more politicians on corruption problems is analyzed using discontinuities in the required minimum size of local councils. Despite the fact that Sweden consistently has been ranked among the least corrupt countries in the world, the survey suggest that non-trivial corruption problems are present in Sweden. Municipalities with more local council seats have more reported corruption problems, and the regression discontinuity design suggests that the effect is causal.

Full text

More politicians, more corruption: evidence
from Swedish municipalities
Andreas Bergh
1,2
•Gu
¨nther Fink
3
•Richard O
¨hrvall
4,5
Received: 30 June 2016 / Accepted: 15 May 2017 / Published online: 25 May 2017
ÓThe Author(s) 2017. This article is an open access publication
Abstract In the literature on political economy and public choice, it is typically assumed
that government size correlates positively with public corruption. The empirical literature,
however, is inconclusive, owing to both measurement problems and endogeneity. This
paper creates a corruption index based on original data from a survey covering top
politicians and civil servants in all Swedish municipalities. The effect of more politicians
on corruption problems is analyzed using discontinuities in the required minimum size of
local councils. Despite the fact that Sweden consistently has been ranked among the least
corrupt countries in the world, the survey suggest that non-trivial corruption problems are
present in Sweden. Municipalities with more local council seats have more reported cor-
ruption problems, and the regression discontinuity design suggests that the effect is causal.
Keywords Corruption Government size Institutions Local government Political
economy Sweden
Electronic supplementary material The online version of this article (doi:10.1007/s11127-017-0458-4)
contains supplementary material, which is available to authorized users.
&Andreas Bergh
Andreas.bergh@ifn.se
Gu
¨nther Fink
gfink@hsph.harvard.edu
Richard O
¨hrvall
richard.ohrvall@ifn.se
1
Department of Economics, Lund University, Box 7082, S-220 07 Lund, Sweden
2
Research Institute of Industrial Economics (IFN), Grevgatan 34 - 2 fl, Box 55665,
SE-102 15 Stockholm, Sweden
3
Harvard School of Public Health, 665 Huntington Avenue, Boston, MA 02115, USA
4
Department of Management and Engineering, Linko
¨ping University, SE-581 38 Linko
¨ping,
Sweden
5
Research Institute of Industrial Economics (IFN), P.O. Box 55665, SE-102 15 Stockholm, Sweden
123
Public Choice (2017) 172:483–500
DOI 10.1007/s11127-017-0458-4
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1 Introduction
What is the causal effect of government size on political corruption? A large number of
papers have examined the relationship, but no consensus has emerged (Holcombe and
Boudreaux 2015; Kotera et al. 2012; Fan et al. 2009; Gerring and Thacker 2005; Fisman
and Gatti 2002; Goel and Nelson 1998). The contradictory results found in the literature
likely can be attributed to at least three factors: first, differences in identification strategies
chosen to overcome the inherent endogeneity of government size (suggesting that countries
with less corruption may be more willing or able to increase the size of government);
second, the considerable heterogeneity in the measures of government size used; and, third,
the difficulties associated with measuring corruption.
This paper contributes to the relevant literature by using constitutional discontinuities in
the required number of seats in Sweden’s 290 local councils to estimate the causal effect of
council size on corruption. A measure of corruption is created using a detailed survey
developed in 2007 and administered to local politicians and civil servants in 2008. Using
both the ordinary least squares (OLS) method, an instrumental variable (IV) approach and
a regression discontinuity design (RDD), our results suggest that increasing the number of
seats in local councils leads to more corruption problems.
The corruption survey used has several advantages compared to traditional measures of
corruption. First, with the objective of measuring corruption as ‘‘inappropriate use of
common power and authority for purposes of individual or group gain at common
expense’’ (Warren 2004, p. 332), top politicians and officials in all Swedish municipalities
were asked to report (without revealing their identities) any corrupt behavior observed in
the community in the form of bribe offers personally received and corrupt behaviors
observed among fellow civil servants or politicians. The survey thereby aimed to capture
corruption in a broader sense than in many other studies, including in-kind transfers and
petty corruption. It also included questions regarding both perceptions and experiences of
corruption, as well as opportunities to act corruptly. Second, the same survey instrument
was used for all Swedish municipalities. Hence, we have a consistent measurement while
still having 290 different units to analyze. Third, by targeting both politicians and civil
servants, the survey was completed by people likely to know if their municipalities were
plagued by corruption problems. The validity of that design, of course, hinges on the
assumption that the respondents would reveal corruption problems if such existed. The
respondents were granted anonymity, and the results of the survey suggest that they did not
hesitate to report such problems.
Overall, indications of corrupt behavior were more common than anticipated, given
Sweden’s international reputation as a low-corruption country. A total of 642 out of 1074
respondents in the sample analyzed (60%) indicated that officials in their community had
been offered money or other benefits, and 235 out of 1074 respondents (22%) at least
partially agreed with the statement that diverting public resources for their personal benefit
was easy. The survey data thus fit well with recent evidence of corruption at the local level
in Sweden’s public sector (Dahlstro
¨m and Sundell 2014;Wa
˚ngmar 2013; Statskontoret
2012; Erlingsson et al. 2008a,b).
It bears noting that these indications of corruption problems in Sweden are not visible in
the popular Corruption Perception Index (CPI) compiled by Transparency International
(and similar indicators), wherein Sweden typically is ranked among the least corrupt
countries in the world. The CPI is an index based mainly on experts’ views of the level of
corruption in different countries, and it has met widespread criticism. For example, it has
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been argued that the CPI and other cross-national corruption indices are poor proxies for
the actual level of corruption (Razafindrakoto and Roubaud 2006; Heywood and Rose
2014), and that they tend to focus too much on bribes in relation to other forms of
corruption, making them ill-suited for capturing corruption in established democracies with
highly developed economies (Andersson 2017; Heywood 2015).
Swedish municipalities offer an interesting setting for studying political behavior
because of their high levels of political autonomy and the large amounts of public
resources managed at the municipality level. With an average annual budget that is similar
to the US federal government’s budget in per capita terms, Swedish municipalities are the
main provider of child care, education, and elder care for the entire population, and they
are involved heavily city planning activities, including issuing building permits. By using
within-country variation, country-specific factors are held constant, and the results
obtained are less prone to omitted variable bias than standard cross-country regressions (as
noted by Ba
¨ck 2003).
The results presented in this paper add to the extensive literature on the determinants of
corruption (Treisman 2000,2007; Andvig and Fjeldstad 2001; Montinola and Jackman
2002; Lambsdorff 2006; Pellegrini and Gerlagh 2008; Fan et al. 2009; Goel and Nelson
2011a,b; Potrafke 2012). Treisman (2007, p. 211) summarizes that ‘‘quite strong evidence
suggests that highly developed, long-established liberal democracies, with a free and
widely read press, a high share of women in government, and a history of openness to
trade, are perceived as less corrupt.’’ Similarly, Andvig and Fjeldstad (2001) argue that the
most robust finding in the literature is the negative correlation between corruption and
economic development. They also note that some evidence has been reported that
democracy reduces corruption (but probably only slowly—see Treisman 2000) and that
more open economies exhibit less corruption (see also Wei 2000). Using US state-level
data, Goel and Nelson (2011b) find that some results that hold, regardless of how cor-
ruption is measured: greater educational attainment in a state reduces corruption, while
greater judicial employment (i.e., courts and activities associated with courts) adds to it.
Consistent with the cross-country evidence, they also find that perceived corruption
declines with greater economic prosperity.
Our paper also contributes to a separate literature on corruption at the individual or firm
level. Persson et al. (2003) use cross-country data (such as the Corruption Perception Index
and the International Country Risk Guide) to show that larger shares of candidates elected
from party lists are associated with more corruption. They interpret this as an effect of less
individual accountability. Using Italian data, Cingano and Pinotti (2013) find a revenue
premium in politically connected firms. Similarly, Amore and Bennedsen (2013) use
exogenous changes in Danish local municipality sizes to identify a large positive effect of
political power on the profitability of firms related by family ties to local politicians. Their
work—similarly to the results presented in this paper—suggests that having too many
representatives may be socially costly. Auriol and Gary-Bobo (2012) explicitly model the
number of legislative seats as determined by a tradeoff between the need to economize on
decision-making costs and the democratic requirement that decisions should reflect the
citizens’ true preferences. Their estimates suggest that on the national level, the United
States has too few representatives, while France and Italy have too many. Another closely
related paper is Fan et al. (2009), who combine cross-country analysis with firm-level data
to find that in countries with larger numbers of governmental or administrative tiers, and
with more local public employees, reported bribery is more frequent.
The paper proceeds as follows. Section 2contains some theoretical considerations, and
Sect. 3provides some background description of Sweden, describes the data and the
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empirical approach. The main results and some robustness checks are presented in Sects. 4
and 5contains a concluding discussion. An online appendix contains additional findings
from robustness tests.
2 Theoretical considerations
The most basic models of government corruption in the political economy literature (based
on Shleifer and Vishny 1993) argue that governments have incentives to use their regu-
latory powers to extract bribes, and that too many politicians or political units will result in
more corruption than a unitary government body that internalizes the effect of bribing on
the incentives for firms and citizens to produce wealth. Several relevant mechanisms are
described by Rowley and Schneider (2004). In essence, when citizens delegate decision-
making authority to their elected representatives, that delegation creates a principal–agent
problem: In the absence of political constraints, elected representatives may choose to use
their official positions to pursue their own self-interests.
Based on the mechanisms discussed by Rowley and Schneider (2004), it can be con-
cluded that the isolated effect of council size on such corruption problems is ambiguous
theoretically. For example, voters’ costs of monitoring their elected representatives are
lower in larger councils when the number of constituents per representative is smaller. On
the other hand, in larger councils each representative will be less influential, lowering the
incentives for monitoring. Rowley and Schneider note also that a similar ambiguity holds
for vote buying: Votes are cheaper in larger legislatures, but to assemble a majority
coalition, more of them must be bought.
In the context of Swedish municipalities, Pettersson-Lidbom (2012) argues that
increasing the number of legislators may lead to better monitoring and control of the public
administration, but this is true only if council members are able to coordinate their
monitoring efforts. In all, no clear theoretical prediction exists regarding how larger
councils should affect corruption problems, and the issue must be examined empirically.
3 Background, data and empirical approach
3.1 Swedish municipalities
The current structure on Swedish municipalities can be traced back to a process of change
that started in 1952, and ended with massive municipal amalgamations completed in 1974.
Between those years, Sweden’s population increased from 7.2 million to 8.2 million, but
the number of municipalities fell from 2498 to 278; the number of elected local repre-
sentatives fell likewise by 200,000 (Erlingsson et al. 2010). The stated reason underlying
these changes was to increase the efficiency of local service provision. Naturally, concerns
also were expressed that such efficiency would come at the cost of less democratic par-
ticipation (Nielsen 2003; Wollmann 2004). The average size of Swedish municipal
councils reached a maximum in 1988 (47.85 seats on average) and has been falling steadily
since then to 44.1 seats today (after the 2014 election).
Because some municipalities have been split, Sweden currently is (and was at the time
of the survey) divided administratively into 290 municipalities and 20 regions. Accounting
for 62% of public employment in Sweden in 2014, municipalities are the nation’s largest
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administrative units both politically and economically. Financed mainly by a proportional
income tax of approximately 20%, municipalities are responsible for the provision of
schooling, child care, and elder care, leaving only the provision of social insurance, higher
education, and defense to the central government. Municipalities also handle welfare
provision, zoning issues (including building permits and permission to sell alcohol), cul-
ture, and public transport.
Total public municipality consumption accounts for 20% of national GDP, compared to
7% of GDP for the central government. With an average population of 30,000, the mean
annual municipality budget is USD 217 million. Roughly three-quarters of this amount is
spent on child care, primary and secondary education, and care for the elderly.
Each municipality has its own council (kommunfullma¨ktige), responsible for all
municipal activities. They do not, however, pass legislation. Members of the local council
are elected on party lists and meet approximately once a month. Local councils are elected
(together with regional and central elections) every four years (before 1994, every
three years).
3.2 Corruption in Swedish municipalities
To measure the prevalence of corruption problems in Swedish municipalities, Erlingsson
et al. (2008a,b) conducted a detailed anonymous web-based survey of local top politicians
and high-ranking civil servants in 2008. The survey was developed in 2007 in collaboration
with several Swedish corruption experts, relying on both direct and indirect reports of
corruption as well as anecdotal vignettes. For the survey, the top four politicians (the chair
of the executive board, the vice chair of the executive board, the chair of the municipal
council and the chair of the municipal audit department), along with the top three civil
servants (the municipal manager, the budget manager and the staff manager) in each of the
290 municipalities, were identified and invited to participate in an anonymous online
survey. While the politicians surveyed typically are part of the local councils, the civil
servants are not political appointees. The total survey population comprised 2024 indi-
viduals, and 1184 (58%) of them completed the survey. Among the 35 questions asked in
the survey, six dealt directly with corruption problems. These questions are shown in
Table 1, and they all are used in constructing our index of corruption problems (results
using the six questions separately are presented in the appendix).
The decision to let the respondents be anonymous was made in order to improve the
response rate and the quality of the answers. The drawback is that we do not know who has
responded to the survey. Still, by studying the survey responses, we can gain some
information about the character of the nonresponses. The dataset includes responses from
287 out of 290 municipalities, and for 247 municipalities we have responses from both
politicians and civil servants. The response rate does not have any clear geographical
pattern, and no substantial differences are evident depending on the size of the munici-
palities (geographically or in terms of number of inhabitants). If we compare the response
rate by types of municipalities, using the classification of Swedish municipalities that the
Swedish Association of Local Authorities and Regions has constructed based on structural
parameters such as population, commuting patterns, and economic structure, we find that
the response rates vary between 52 and 62%. Furthermore, if we compare the two main
groups of respondents, we find that propensities to fill out the questionnaire are about the
same—59% among politicians and 57% among civil servants. The difference in response
rate by gender is only about one percentage point. Thus, the data we have do not indicate
any severe problems related to nonresponse bias.
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All questions offered multiple response categories, ranging from ‘‘never’’ to ‘‘very
often’’ (questions 1–4) and from ‘‘strongly agree’’ to ‘‘strongly disagree’’ (questions 5 and
6). Figure 1shows the share of respondents reporting inappropriate behaviors, as well as
the share of respondents not answering for each of the six questions in Table 1. Response
rates were highest for the first question, which inquired directly about personal experi-
ences; most respondents appeared to reject indications of personal involvement without
much hesitation. Response rates were lowest for reports on others having accepted bribes.
With respect to actual evidence of corrupt behavior, a remarkable 59.8% of all respondents
indicated that others in their community had been offered bribes while in office, and more
than half (51%) of respondents indicated that bribes had been accepted. A total of 41.4%
Table 1 Questions on corruption problems
Survey question
1. How often are you, in your position as elected representative or in your duty, offered money or other
benefits in order to make a decision in favor of the person/persons offering the benefit?
2. How often do you think other politicians and civil servants in your municipality are offered money or
other benefits in order to make a decision in favor of the person/persons offering the benefit?
3. How often do you think other politicians and civil servants in your municipality have actually accepted
the benefit offered to them?
4. How often have you been subject to violence, threat of violence, or blackmail, where the person exposing
you has demanded that you, in your municipal duty/service, act in a way that you would otherwise have
not?
5. In my municipality, the public procurement process is impartial.
6. If I wanted to, it would be easy for me to bring benefits to me or my close ones, at the expense of the
municipality
Questions have been translated from Swedish by the authors
Fig. 1 Corrupt behavior questions: response rate and fraction reporting corrupt behavior
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suggest that the letting of procurement contracts is not impartial, and 23.3% reported
having been threatened during their official duties.
In our baseline analysis, we aggregated the answers into a single measure of corruption
using principal component analysis (PCA). The first principal component accounted for
35% of the total variation, with the highest factor loadings on questions 2 and 3, where
respondents reported on other politicians being offered or having accepted bribes. We
normalized the PCA score into a z-score variable with mean 0 and standard deviation 1. As
a robustness test, reported in the appendix, we also created and tested alternative indices
without substantial changes in the results.
Corruption reports varied both within and across municipalities, with community-level
variation accounting for approximately 25% of the total variation in reported corruption.
The substantial variation in reporting across individuals may be interpreted as differences
in subjective experiences, but may also reflect heterogeneity in reporting. To use the
provided information most efficiently, we relied on individual reports (aggregated into
individual-level summary scores) in our main empirical specification. As robustness
checks reported in the appendix, we show results for specifications where survey answers
were aggregated at the municipal level, and where answers to the six questions were
analyzed item by item.
3.3 Other municipality data
Data on municipal characteristics were combined from a variety of sources. Population size
and age structure data were collected by Statistics Sweden and compiled by Johansson
(2006). We also compiled information on geographical area, number of council seats,
percentage of population with some college education, ‘‘close’’ recent election results and
whether the same party had been in power since 1995 from the same data source (Jo-
hansson 2006). The reference year for education was 2003; ‘‘some college education’’ was
defined as the fraction of the population 25-to-64 years of age with at least 3 years of post-
secondary education. The reference year for election outcomes was 2006. We defined
elections as ‘‘close’’ if the difference between the two major voting blocs (the four
rightwing parties and the three leftwing parties) was less than five percentage points.
We also entered in our models the total municipality budget and total municipality
income. The former refers to the expenditures of the municipality and is based on Statistics
Sweden’s collection of annual municipality accounts. Municipality income is the regional
gross domestic product (GDPR) in 2008, calculated by Statistics Sweden, divided by the
population of the municipality.
To control for politicians’ genders, we included a variable that measured the share of
women among all local politicians in 2007. The information came from Statistics Sweden’s
survey of elected representatives in municipalities and county councils.
We used information on the number of local newspaper offices in the municipalities to
capture media coverage of local politics. That information was compiled by Bergh et al.
(2013). Furthermore, we also used Statistics Sweden’s information on shares of foreign-
born and shares of urban population within the municipalities. Urban population was
defined as the share of population that resided in the city center. Finally, the variable
‘‘major city’’ was defined as one of the three largest Swedish municipalities, i.e., Stock-
holm, Gothenburg and Malmo
¨.
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3.4 Empirical model
The main empirical model we estimate is
cij ¼aþb1SeatsjþXjcþRi/þdrþeij;
where C
ij
is the (normalized) corruption score of government official iin municipality j,
Seats is the number of council seats, X
j
and R
i
are vectors of municipality and respondent
characteristics, and d
r
are regional fixed effects. At the municipality level, we control for
total spending, population size (linear and quadratic), the share of population of working
age and retirement age (20–64 and 65 and older), municipality income, area (thousands of
square km), the share of adults with at least a college education as proxy for the average
human capital in each municipality, urban population share, and foreign population share.
We also control for factors that have been found to affect corruption in previous studies:
the share of women in local elected politicians (Swamy et al. 2001), the number of local
newspaper offices (Gentzkow et al. 2004) and local political competition (Shleifer and
Vishny 1993).
Though we do not aim to uncover a causal effect of public spending on corruption, we
control for public spending because Pettersson-Lidbom (2012) finds that municipalities
with larger local councils tend to have smaller public expenditures, which may influence
corruption. At the respondent level, we include the following variables collected as part of
the corruption survey: sex, type of appointment (politician or bureaucrat) and educational
attainment. Because those who are new in local politics are less likely to have experienced
corruption problems, we also control for the duration of current appointment.
3.5 Identification strategy
To overcome endogeneity concerns, we make use of the fact that Swedish law Kommu-
nallagen has constitutional status and stipulates a minimum number of seats for each local
council depending on the number of eligible voters residing in the municipality on March 1
the year before the election year. The purpose of the rule is to ensure a high level of
proportionality and representation of minority populations. Electoral rolls are produced
based on a national population register, which is updated daily. Voter cards are sent to all
entitled to vote, and no registration is necessary. The minimum number of seats per
municipality is fixed for a term. The required council sizes have been constant since 1979,
and are summarized in Table 2.
Municipalities are allowed to increase the number of seats above the minimum
requirement, and several have chosen to do so in the past. Given that the law is tied in a
non-linear way to population, the law is in practice most relevant when municipalities
Table 2 Constitutional mini-
mum council seat requirements Number of eligible voters Minimum number of seats
From To
0 12,000 31
12,001 24,000 41
24,001 36,000 51
36,001 100,000 61
Stockholm 101
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grow in population, and must increase their council size when passing a threshold. Figure 2
shows the minimum as well as the actual number of seats at the municipality level. It is
clear that the law is being followed, and it is noted that the requirements are exceeded by a
large number of municipalities (59% overall), while 41% of municipalities currently have
exactly the number of seats legally required.
Table 3shows descriptive statistics for all variables used in the analysis.
4 Results
Basic OLS results are reported in Table 4. In column 1, we show the relation between
corruption and council size conditional on total expenditure, municipality population and
income, newspaper presence, and electoral competition. In column 2, we add controls for
municipality characteristics; in column 3, we add respondent characteristics, and in column
4, we add regional dummies. Overall, the positive effect of seats is stable, and statistically
significant across all specifications. The estimated coefficient is relatively small in mag-
nitude, implying that a 10-seat increase (which typically is required when a threshold is
passed) increases corruption by about 0.15 standard deviations. Corruption increases
strongly with population size, while income per capita has the expected negative sign, but
is not significant.
Among the control variables, it is worth noting that having no change in the ruling party
in the recent past is associated with a small reduction in corruption. This effect could be
generated by respondents systematically reporting more corruption in more contested
areas. We find no effect of media presence, and among the municipality controls not
shown, no effect is found for female politician share—in contrast to Dollar et al. (2001).
Finally, it is worth noting that total government expenditure is negatively associated with
corruption, and increasingly so when more controls are added. While this cannot be
interpreted as a causal effect, the negative sign is still surprising if one expects public
expenditures to cause corruption.
30 40 50 60 70 80
Number of parliament seats in 2006
010000 20000 30000 40000 50000
Number of voters in municipality
Fig. 2 Actual and minimum number of seats in local councils 2007
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4.1 Instrumental variable estimation
Given the cross-sectional nature of the data used in Table 4, we cannot rule out that the
positive conditional associations observed are at least partially the result of unobserved
confounders, or that that the number of seats may respond to perceived corruption (si-
multaneity bias). To address these concerns, we report two-stage-least squares (instru-
mental variable—IV) estimates in Table 5. As discussed above, the instrument used is the
required number of seats. The main logic of the IV approach is to use variation in the
exposure (council size in this case) driven by an exogenous variable (the instrument). As
long as the instrument predicts the independent variable of interest but is orthogonal to the
residual of the main equation, IV estimation will yield unbiased estimates of the causal
effect of interest. As a first step, we examine the predictive power of the instrument, i.e.,
assess whether the constitutionally required number of seats predicts the actual number of
seats conditional on a full set of control variables. This is indeed the case: in all specifi-
cations, the F-statistics for the instrument range between 80 and 118. In column 1, we
control for the main variables of interest only, and gradually add a larger set of controls in
columns 2–4. The overall results remain very similar to OLS, with a small increase in the
Table 3 Descriptive statistics
Variable Mean Standard deviation Min Max
Municipality characteristics
Area (’000 square km) 1.39 2.39 0.009 19.37
Close recent election (\5% point gap) 0.14 0.35 0 1
Female politicians share 41.28 4.30 24.4 52.4
Foreign population share 2010 (%) 10.84 5.32 4.05 39.75
Major city 0.01 0.11 0 1
Municipality income 2008 (US$ ’000s) 40.33 17.58 17.25 174.58
Municipality council seats 2006 45.70 12.08 31 101
Number of local newspaper offices in municipality 1.20 0.92 0 4
Percentage of adults with some college education 12.23 6.03 5 48
Population size (10,000) 3.27 6.45 0.26 77.12
Same party in power since 1995 0.46 0.50 0 1
Share of population age 65 and older 19.36 3.62 9.40 28.90
Share of population age 80 and older 5.90 1.40 1.89 9.89
Total budget 2004 (US$ 100 millions) 2.17 4.61 0.21 57.13
Urban population share 2010 (%) 74.65 14.56 31 100
Working age share (20–64) 56.24 2.51 48.20 65.80
Respondent characteristics
Female 0.30 0.46 0 1
Bureaucrat 0.42 0.49 0 1
Secondary education 0.19 0.39 0 1
Tertiary education 0.70 0.46 0 1
In office for 2–10 years 0.44 0.50 0 1
In office [10 years 0.29 0.45 0 1
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estimated impact of additional seats. Our most tightly controlled model (Table 6, column
4) suggests that a 10-seat increase leads to 0.26 standard deviation increase in corruption.
The slightly larger effect (0.26 standard deviations versus 0.17 in OLS) suggests that the
OLS estimate is biased downwards. A possible mechanism is that less corrupt munici-
palities opt for larger councils. Another plausible explanation is that the exogenous shifts in
council size generated by the law (essentially forcing government size to increase from 30
to 40, or from 40 to 50 members) generates more corruption than a gradual increase in
council size observed outside of the minimum requirement. As such, the estimated IV
coefficients should be interpreted as a local average treatment effect for legally required
increases in council size, which may not apply to more gradual increases in council size.
We explore this aspect in greater detail in the following section, where we show estimates
based on a regression discontinuity design.
4.2 Regression discontinuity design
As an alternative identification strategy, the discontinuities in the required number of seats
can be used to estimate the effect of council size on corruption using a regression dis-
continuity design (RDD). For RDDs to yield statistically meaningful estimates, a
Table 4 OLS results
Dependent variable Corruption principal component z-score
(1) (2) (3) (4)
Number of council seats 0.0127**
(0.00530)
0.0191***
(0.00667)
0.0193***
(0.00670)
0.0168***
(0.00635)
Total government
expenditure 2007
-0.306* (0.157) -0.422**
(0.181)
-0.440**
(0.189)
-0.631**
(0.275)
Population size (millions) 17.17* (9.631) 21.83** (10.92) 23.04** (11.40) 36.49** (16.85)
Population size squared 7.737** (3.298) 10.61***
(3.709)
10.92*** (3.846) 12.63***
(4.794)
Municipality income -0.000197
(0.00217)
-0.00101
(0.00239)
-0.000978
(0.00242)
0.000603
(0.00257)
Same party in power since
1995
-0.139**
(0.0637)
-0.146**
(0.0710)
-0.169**
(0.0730)
-0.176**
(0.0722)
Number of newspaper
offices
0.0167 (0.0362) 0.00873
(0.0385)
0.00950 (0.0382) -0.0110
(0.0452)
Municipality controls No Yes Yes Yes
Respondent controls No No Yes Yes
Region fixed effects No No No Yes
Observations 1074 1074 1074 1074
R-squared 0.029 0.036 0.045 0.061
*** pvalue \0.01; ** pvalue \0.05; * pvalue \0.1. Standard errors are clustered at the municipality
level. Municipality controls included in addition to population and income are population share 20–64,
population share 65 and older, municipality land area, share of population with higher education, indicator
for big cities, foreign population share, urban population share, and female politician share. Respondent
controls included are sex, an indicator for bureaucrat (as opposed to politician), educational attainment
dummies, and tenure length group (2–10 years, [10 years in office)
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sufficiently dense distribution of outcomes around the cutoff is needed. As illustrated in
Fig. 3, the majority of Swedish municipalities have less than 25,000 voters, and the most
suitable cutoff for an RDD design is the increase in the required number of seats at 12,000
voters. Table 6compares average municipality and politician characteristics around that
cutoff. Using a population band of 3000 voters on either side of the cutoff, there are 176
records just below the cutoff and 80 observations just above the cutoff, and these obser-
vations do not differ with respect to any of the variables available.
1
One of the most critical assumptions underlying the RDD design is that the assignment
variable cannot be modified by subjects. In this case, it is worth emphasizing that
municipalities are not involved in the process of deciding where eligible voters vote and
that no one has to register in order to be able to vote. Statistics on the number of eligible
voters are produced by the Swedish Election Authority, based on the national population
register, which is updated daily. To confirm that there is no ‘‘gaming’’ around the 12,000-
voter cutoff, we plot population densities in Fig. 3. If politicians or other stakeholders were
Table 5 IV estimation main results
Dependent variable Corruption principal component z-score
(1) (2) (3) (4)
Number of council seats 0.0139*
(0.00714)
0.0276**
(0.0112)
0.0281**
(0.0111)
0.0264**
(0.0110)
Total government
expenditure 2007
-0.307**
(0.155)
-0.452**
(0.179)
-0.471**
(0.187)
-0.662**
(0.273)
Population size (millions) 16.88* (9.463) 20.37** (10.27) 21.52** (10.70) 34.62** (16.50)
Population size squared 8.158** (3.636) 13.61***
(4.928)
14.03***
(5.014)
16.02***
(5.683)
Municipality income -0.000218
(0.00217)
-0.00131
(0.00247)
-0.00128
(0.00249)
0.000433
(0.00258)
Same party in power since
1995
-0.138**
(0.0635)
-0.146**
(0.0711)
-0.168**
(0.0729)
-0.178**
(0.0713)
Number of news outlets 0.0144 (0.0389) -0.00674
(0.0439)
-0.00654
(0.0435)
-0.0242
(0.0468)
Municipality controls No Yes Yes Yes
Respondent controls No No Yes Yes
Region fixed effects No No No Yes
Cragg–Donald F-stat 943.8 427.6 424.4 445
Observations 1074 1074 1074 1074
R-squared 0.029 0.034 0.043 0.059
First stage F-stat 80.08 100.2 98.99 117.9
*** pvalue \0.01; ** pvalue \0.05; * pvalue \0.1. Standard errors are clustered at the municipality
level. Municipality controls included in addition to population and income are population share 20–64,
population share 65 and older, municipality land area, share of population with higher education, indicator
for big cities, foreign population share, female politician share, and urban population share. Respondent
controls included are sex, an indicator for bureaucrat (as opposed to politician), educational attainment
dummies, and tenure length group (2–10 years, [10 years in office)
1
Unfortunately, the cutoff at 12,000 voters is the only one with a sufficiently large number of municipalities
close to the cutoff. As a result, we cannot say if the effect of additional seats on corruption is linear or not.
494 Public Choice (2017) 172:483–500
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Table 6 Balance around the 12,000-voter cutoff
Below cutoff Above cutoff Equal means
test
N=176 N =80
Voters 9000–11,999
voters
12,000–14,999
voters
Pvalue
Mean SD Mean SD
Public expenditure per capita (SKR) 45,389.7 4220.5 45,667.4 2817.4 0.764
Income per capita 2006 (US$’000 s) 37.2 14.5 34.6 6.2 0.339
Share of population age 65 and older 55.6 1.4 56.1 1.8 0.356
Share of population age 80 and older 19.8 3.1 19.0 3.4 0.474
Area (’000 square km) 1.1 1.8 0.7 0.5 0.155
Percentage of adults with some college
education
10.1 3.0 11.2 4.7 0.336
Foreign population share 2010 (%) 10.2 4.6 9.3 2.4 0.330
Urban population share 2010 (%) 70.4 14.6 74.8 13.0 0.276
Female politicians share in council 40.3 4.6 39.9 5.2 0.749
Female 0.3 0.4 0.2 0.4 0.489
Bureaucrat 0.4 0.5 0.4 0.5 0.742
Secondary education 0.2 0.4 0.2 0.4 0.404
Tertiary education 0.7 0.5 0.7 0.5 0.564
In office for 2–10 years 0.4 0.5 0.4 0.5 0.758
In office [10 years 0.3 0.5 0.3 0.4 0.211
Number of local newspaper offices in
municipality
1.1 0.9 1.4 0.8 0.369
Same party in power since 1995 0.4 0.5 0.4 0.5 0.594
010 20 30 40
Frequency
9000 10000 11000 12000 13000 1400 0 15000
Number of voters in municipality
Fig. 3 The distribution of municipality voters around the 12,000-voter cutoff. Note The figure shows the
number of municipalities in brackets of 500 voters for communities between 9000 and 15,000 voters
Public Choice (2017) 172:483–500 495
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able to modify population numbers to reach specific thresholds of interest, we should see
an increase in the number of municipalities with population sizes just above these specific
thresholds. As Fig. 3shows, this is not the case: the distribution of municipality sizes just
above and below the threshold seems to be almost the same, which suggests that population
estimates have not been modified.
Figure 4shows the estimated increase in the number of seats around the cutoff. The
number of seats increases almost linearly to the left of the cutoff and then jumps by about
two additional seats. The fact that a substantial part of the jump occurs just below the
cutoff suggests that many municipalities anticipate that population will increase, and
increase the number of seats voluntarily.
Figure 5shows the main regression discontinuity results for the corruption z-score.
Average corruption scores are distributed fairly evenly around zero to the left of the cutoff
and have a strong cluster in the 0.3 range just above the cutoff.
30 35 40 45 50
seats
05000 10000 15000 20000
voters
Fig. 4 Number of voters and council size
-.4 -.2 0.2 .4
corrupt_zscore_pca
05000 10000 15000 20000
voters
Fig. 5 Average corruption scores around the 12,000-voter cutoff
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As Fig. 5shows, corruption outcomes are relatively noisy above the cutoff. That sug-
gests that the RDD results are potentially sensitive to different trend specifications. Table 7
shows the results of testing different ranges and trend specifications. While the estimates
fluctuate a bit—particularly when the range is expanded—the average increase in the
corruption score around the cutoff appears to be about 0.4 standard deviations. Given that
the average number of seats increases by about two around the cutoff, this result implies a
marginal increase in corruption of about 0.2 standard deviations for each additional seat.
These estimates are large compared to OLS estimates, suggesting that the exogenous shift
in additional seats at this specific cutoff may trigger a substantial increase in (perceived)
corruption problems in the affected municipalities.
4.3 Additional robustness tests
In an online appendix, we show results from several additional robustness tests that all
confirm the main findings. In summary, we tried creating the corruption index using a
binary coding, where each question is coded as 1 if the respondent’s answer indicated any
type of problem, regardless of frequency or intensity. We also constructed a categorical
index where answers were translated into ordinal scales, so that ‘‘never’’ is coded as 0,
‘‘very rarely’’ as 1, ‘‘rarely’’ as 2, ‘‘sometimes’’ as 3 and ‘‘often’’ as 4, and then summing
the scores to create the index. Both alternative codings generated results similar to the
baseline findings.
Analyzing the six questions separately reveals significant positive findings for the
answers to the questions asking if others in the municipality have been offered money and
whether others have accepted bribes. In contrast, respondents’ views on the impartiality of
public procurement clearly are not driving the results.
Our results also can be reproduced with municipality level regressions using the median
and the maximum level of reported problems in each municipality as dependent variables.
Intuitively, using the minimum level of reported problems, no significant results are found.
Finally, we verified that the results for seats still hold if expenditures are excluded from
the model. The correlation between the number of seats in local councils and public
expenditure per capita is only -0.17, indicating that they are two different dimensions of
government size, and that no bias is introduced by including them simultaneously in the
regressions.
Table 7 Regression discontinuity design estimates (cutoff 12,000)
Dependent
variable
Corruption principal component z-score
(1) (2) (3) (4) (5) (6) (7)
41-seat
requirement
0.409***
(0.138)
0.444***
(0.149)
0.366*
(0.187)
0.276
(0.314)
0.403***
(0.142)
0.474**
(0.198)
0.440**
(0.220)
Trend
specification
linear linear linear linear quadratic cubic quartic
Sample min
voters
2000 4000 6000 9000 2000 2000 2000
Sample max
voters
22,000 20,000 18,000 15,000 22,000 22,000 22,000
Estimated coefficients represent the estimates shift in average corruption z-scores around the cutoff (intent-
to-treat)
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5 Concluding discussion
Corruption is a well-documented social problem, associated with a multitude of undesir-
able social outcomes including slow economic growth (Mauro 1995; Reinikka and
Svensson 2004;Me
´on and Sekkat 2005; Holmberg and Rothstein 2011; Johnson et al.
2011).
2
This paper investigated the empirical relation between local council size and
corruption, using data from a newly collected corruption survey in Sweden. While the
relation between council size and corruption is ambiguous theoretically, our empirical
results are clear: OLS, instrumental variable regressions and a regression discontinuity
design based on discontinuities in the required number of council seats, all suggest that
larger councils lead to more corruption problems.
The ongoing trend towards smaller local councils in Sweden, is thus a factor that has
served to dampen corruption problems there, according to our findings. Our data do not allow
us to identify the exact mechanism that explains the result, but a possible interpretation is
that larger councils mean weaker incentives for monitoring elected representatives.
The findingspresented are well aligned with thosereported by Fan et al. (2009), who find that
bribery is more frequent in countries with larger numbers of administrative tiers. They also fit
well with results for US states reported by Goel and Nelson (2011b), who find that corruption
increases when local public goods and services are delivered by a larger number of govern-
mental units (holding public expenditure constant). The fact that more politicians cause cor-
ruption problems, whereas municipalities with larger public expenditure are not more corrupt,
also sits well with the results in Holcombe and Boudreaux (2015), that it is the regulatory state,
rather than the productive or redistributive state, that is associated with corruption.
The size of the estimated effect is modest, with 10 additional council seats increasing
our corruption index by about one-quarter of a standard deviation (based on the IV-
estimates in Table 5). It bears emphasizing that Sweden consistently has been ranked
among the least corrupt countries in the world, which means that the size of the marginal
effect may well be interpreted as a lower bound internationally. Even in Sweden, more
politicians seem to imply more corruption. While it may make sense from a democratic
perspective not to have local councils that are too small, our results suggest that having
smaller councils is an option worth considering in the fight against corruption.
Acknowledgements The authors gratefully acknowledge the helpful comments from two referees, the
journal’s editor, and financial support from the Swedish Research Council and Torsten So
¨derberg’s
Foundation.
Open Access This article is distributed under the terms of the Creative Commons Attribution 4.0 Inter-
national License (http://creativecommons.org/licenses/by/4.0/), which permits unrestricted use, distribution,
and reproduction in any medium, provided you give appropriate credit to the original author(s) and the
source, provide a link to the Creative Commons license, and indicate if changes were made.
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