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A Dynamic Analysis of Economic Freedom and Income Inequality in the 50 U.S. States: Empirical Evidence of a Parabolic Relationship

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This paper examines the dynamic relationship between economic freedom and income inequality in the 50 U.S. states over the 1979-2004 period. Using fixed effects regression analysis, we find evidence that increases in economic freedom are associated with lower income inequality, but the dynamic relationship between the two variables depends on the initial level of economic freedom. This suggests that there may be an inverted U-shaped relationship between economic freedom and income inequality. The inflection point at which additional increases to economic freedom in a state result in less income inequality is estimated. The results are robust to various time periods and several alternative measures of income inequality.
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A Dynamic Analysis of Economic Freedom and Income
Inequality in the 50 U.S. States: Empirical Evidence of a
Parabolic Relationship
Daniel L. Bennett# and Richard K. Vedder*
#Florida State University – USA, *Ohio University USA
Abstract. This paper examines the dynamic relationship between economic freedom and income
inequality in the fifty U.S. states over the 1979-2004 period. Using fixed effects regression
analysis, we find evidence that increases in economic freedom are associated with lower
income inequality, but the dynamic relationship between the two variables depends on the
initial level of economic freedom. This suggests that there may be an inverted U-shaped
relationship between economic freedom and income inequality. The inflection point at
which additional increases to economic freedom in a state result in less income inequality is
estimated. The results are robust to various time periods and several alternative measures of
income inequality.
1. Introduction
Economic inequality is one of the most divisive
contemporary political issues in the United States.
In a 2011 speech, President Obama invoked the age-
old dogma that the rich are getting richer while the
poor are getting poorer in claiming that “over the
past three decades, the middle class has lost ground
while the wealthiest few have become even wealthi-
er.”1 In doing so, President Obama was making the
case for a second government stimulus since taking
office, with the implicit message that government
redistribution is a corrective mechanism necessary
to reduce the inequalities created under a market
system.
President Obama is correct in suggesting that in-
come inequality has increased over the past 30 years
in the United States, according to most aggregate
1 Remarks of President Barack Obama from his weekly radio
address on October 29, 2011, entitled ``We Can't Wait to Strength-
en the Economy and Create Jobs." Transcript reviewed March 26,
2012 at www.whitehouse.gov/blog/2011/10/29/weekly-address-we-can-t-wait-
create-jobs.
measures. The household income Gini coefficient, a
measure of relative inequality in the distribution of
income among households that ranges from zero
(perfect equality) to one (perfect inequality), in-
creased from 0.403 to 0.469 over the 1980-2010 peri-
od, a 16.5% increase in inequality.2 Figure 1 displays
the trend in the income Gini household measure in
the U.S. from 1980-2010. Across the fifty U.S. states,
a similar pattern of increased income inequality
emerged, with the average state-level household
income Gini measure increasing by 13.2% over the
1980-2010 period.3 Among the states, inequality
grew by the least (6.4%) in Mississippi and the most
(24.6%) in Connecticut.
2 The 80/20 income ratio, or the ratio of the upper income limits
of the 80th to 20th percentile of households, is an alternative
measure of inequality. Similar to the household income Gini
measure, the 80/20 income ratio grew from 4.2 to 5.0 between
1980-2010, an increase in inequality of 18.9%.
3 Average here refers to the simple, non-population weighted
mean percentage change in household income Gini. State data
are from the U.S. Census Bureau's decennial Censuses.
J
RAP 43(1): 42-55. © 2013 MCRSA. All rights reserved.
Economic Freedom and Income Inequality 43
Figure 1: U.S. household income Gini coefficient, 1980-2010.
To the extent that reducing inequality is a policy
objective, gaining a better understanding of the rela-
tionship between economic freedom and inequality
is needed. In this paper, we empirically examine the
relationship between state-level economic freedom,
as measured by the Fraser Institute's Economic
Freedom of North America Index, and relative in-
come inequality for the fifty U.S. states over the
1979-2004 period. We use the family income Gini
coefficients of Galbraith and Hale (2006), hereafter
GH, as our primary measure of income inequality.4
In empirically examining the effect that economic
freedom exerts on income inequality, there are gen-
erally two channels to examine. The first is to ana-
lyze the relationship between levels of economic
freedom and inequality through panel analysis.
Using a variety of methods, we fail to find a statisti-
cally significant relationship between the level of
economic freedom and income inequality in the U.S.
While the static relationship between the two varia-
bles is unclear, fixed effects models exploring the
dynamic relationship between economic freedom
and income inequality are more revealing. Our find-
ings suggest that increases in economic freedom are
associated with lower income inequality, but this
4 GH (2006) calculated annual state Theil statistics from the BEA
industry- and sector-level data from 1969-2004 and used them to
fit family income Gini coefficients using the U.S. Census Bureau's
Current Population Survey data.
effect depends on the initial level of economic free-
dom, implying that there is an inverted U-shape re-
lationship between economic freedom and income
inequality. The results are robust to various time
spans and several alternative measures of income
inequality.
The remainder of the paper is organized as fol-
lows. Section two provides a brief literature review
and is followed by a discussion of the Economic
Freedom of North American Index. In section four,
our methodology and empirical results are present-
ed. Section five discusses the theoretical possibility
of an inverted U-shaped economic freedom-income
inequality curve, along with evidence of its existence
from our analysis. The last section offers concluding
remarks.
2. Literature review
Economic theory does not yet offer clear
guidance on the anticipated relationship between
economic freedom and income inequality. Berggren
(1999) attempted to provide a theoretical foundation
in showing that economic freedom influences
income inequality through various channels, but
concluded that the net effect of economic freedom —
both in levels and changes — on income inequality
is theoretically ambiguous. The ambiguity in Berg-
gren's theory is due to the anticipated differential
44 Bennett and Vedder
effect on inequality of the various components com-
prising a measure of economic freedom.5
Both Berggren (1999) and Carter (2006) suggest
that government redistribution, which reduces eco-
nomic freedom, leads to an increase in equality. As
Barro (2000) notes however, the anticipated negative
relationship between redistribution and inequality
rests on the assumption that the “distribution of po-
litical power is more egalitarian than the distribution
of economic power,” implying that redistribution
must be vertical rather than horizontal to induce
improvements in equality. In addition to Barro's
point that relates to rent-seeking and corruption, it is
also conceivable that redistribution leads to in-
creased inequality through other channels. The rev-
enues used to finance redistribution are largely
raised through distortionary taxation that provides a
disincentive to work. If the disincentives are large
enough, then some -- particularly those at or near
the eligibility level for transfer programs -- may be-
come dependent on the government for transfers
and likely experience stagnation in their income
over time. Meanwhile, those remaining in the labor
force continue to acquire human capital and likely
experience income gains, resulting in an increase in
income inequality over time.6 Vedder, Gallaway,
and Sollars (1988) provide an overview of some of
the other arguments that have been advanced as to
why government redistribution might not reduce
inequality, including the crowding out of private
sector charity and the capitalization of public
transfer payments.7 Thus, it is not clear a priori that
governmental redistribution serves as an inequality-
reducing policy mechanism.
Redistribution is not the only policy related to
economic freedom that potentially exerts an impact
on inequality. Several authors have empirically ex-
amined the relationship between individual compo-
nents of economic freedom and inequality. Clark
and Lawson (2008) found that high marginal tax
5 Berggren's analysis as well as ours adopts the Gwartney, Law-
son, and Hall (2012) definition of economic freedom, which in-
cludes policies and institutions that promote personal choice,
voluntary exchange, freedom to compete, and security of private-
ly owned property.
6 See Cox and Alm (1995) for empirical evidence of income mobil-
ity in the U.S.; Barro (2000) also discusses the disincentive to in-
vest and save created by redistribution, implying that these disin-
centives may act to increase inequality through slowing of eco-
nomic growth.
7 See Gruber and Hungerman (2007) and Hungerman (2005) for
empirical evidence of public sector crowd out; see Gwartney and
Stroup (1986) and Tullock (1986) for discussion of market adjust-
ments to transfers.
rates are negatively related to income inequality,
suggesting that progressive tax and redistribution
policies increase equality, but that other aspects of
economic freedom such as property rights, sound
money, trade openness and limited government act
to reduce income inequality. Berggren (1999) found
that trade openness and financial deregulation exert
a significant negative impact on income inequality.
Scully (2002) indicated that the size of government,
as measured by the government consumption and
transfers and subsidies as a share of GDP ratios, is
associated with greater income equality, but that
government intervention in the form of state-owned
enterprises is associated with greater income ine-
quality. Ashby and Sobel (2008) suggested that min-
imum wage reductions and lower tax burdens
would be the best policies to reduce income inequal-
ity in the United States.
A number of authors have conducted holistic
analyses of the relationship between economic free-
dom and inequality using a variety of methodolo-
gies, but the results have thus far been inconclusive.
Berggren (1999) concluded that across countries the
level of economic freedom is positively associated
with inequality, but changes that enhance economic
freedom over time lead to lower inequality; howev-
er, as Carter (2006) points out, Berggren's model can
be rewritten as a distributed-lag model, and as such,
the regression results imply the opposite of what
Berggren indicates, namely that the short-run effect
of the level of economic freedom on inequality is
negative and the long-run effect is positive.
Using a multiple stage approach, Scully (2002)
found that economic freedom and income inequality
are negatively related across countries and that in-
creases in economic freedom enhance income equali-
ty through the growth of the share of market income
earned by the two lowest income quintiles and a
reduction in the share of the highest quintile. Ashby
and Sobel (2008), in an analysis of the fifty U.S.
states, found that policy changes that enhance eco-
nomic freedom lead to higher levels of income and
income growth for all income groups, acting to
reduce relative income inequality. Concerning the
results of Ashby and Sobel, the Carter critique is
applicable, as their regression models can also be
rewritten as distributed lag models, and as such, the
results are subject to a different interpretation.
Carter (2006) employed a fixed effects approach
in exploring whether there is a parabolic relation-
ship between economic freedom and income ine-
quality across countries, finding that beginning at a
low initial level of economic freedom, increases in
Economic Freedom and Income Inequality 45
economic freedom can exert a negative effect on
income inequality, but that beyond a relatively low
level of freedom, additional increases generate in-
creased inequality. Carter's results contradict those
of Berggren (1999) and Scully (2002) and, as Carter
notes, imply that there is a tradeoff between eco-
nomic freedom and income inequality. Thus, there
remains ambiguity in the relationship between eco-
nomic freedom and inequality.
The contribution of the current research is two-
fold. First, our analysis of the relationship between
income inequality and economic freedom in the U.S.
differs in two significant ways from that of Ashby
and Sobel (2008). Their analysis uses the ratio of the
highest-to-lowest income quintiles as the measure of
relative income inequality, while our paper utilizes
income Gini coefficients, alternative measures of
relative income inequality that account for the entire
income distribution. We also use several alternative
income Gini measures as a robustness check. In ad-
dition, our analysis incorporates lagged dynamic
effects, whereas that of Ashby and Sobel does not.
Next, while Carter (2006) examined the parabolic
relationship between economic freedom and income
inequality across countries, our analysis explores
this relationship across states. Given that some insti-
tutions and policies are established at the national
level, the results from international and subnational
analyses might differ.
3. Economic freedom in the U.S.
The independent variable of interest for this
study is economic freedom. We use data from the
Fraser Institute's annual Economic Freedom of North
America (EFNA) report to measure economic free-
dom for several reasons. First, the economic institu-
tions and policies measured by the index are con-
sistent with the definition of economic freedom ad-
vanced by James Gwartney, Bob Lawson, and their
various co-authors over the years. As noted by its
authors, the EFNA index attempts “to gauge the
extent of the restrictions on economic freedom im-
posed by governments in North America,” includ-
ing the United States (Ashby, Bueno, and McMahon
(2011)). Thus, it attempts to measure the extent to
which states enact policies consistent with free
market principles.
Second, it provides annual state-level data for the
period spanning 1981-2009, making it the only com-
prehensive dataset available for a long enough peri-
od of time to evaluate the dynamic effects that
changes in market-oriented policy exert on income
inequality.8 Next, the index is comprised of reliable,
data-driven measures that are consistent across
states, with the data for the underlying variables
easily accessible. Finally, the data are provided at
two levels: all government, which includes all feder-
al, state, and local government economic activity
within a state, and subnational, which includes only
state and local government economic activity; this
provides an opportunity to examine whether eco-
nomic policy by level of government exerts a differ-
ential impact. The analysis and results reported
here make use of the subnational data only, since
state governments have little control over federal
economic policy.
The EFNA index is comprised of three main are-
as, each with several components. The three areas
are: size of government; takings and discriminatory
taxation; and labor market freedom. Each compo-
nent is transformed to a 0-10 scale and is assigned an
equal weight for the area to which it belongs, and
each area is given an equal weighting in the compo-
site index score. Table 1 provides a breakdown of
the three areas and subcomponents of each.
Table 1. Economic Freedom of North
America Index c
omp
onen
ts.
Area 1: Size of Government
1A: General Consumption Expenditures by
Gov
ernmen
t
as a % of
GDP
1B: Transfers and
Subsidies as a % of
GDP
1C: Social
Securi
t
Pa
y
ments as a% of
GDP
Area 2: Takings and Discriminatory Taxation
2A: Total Tax
Re
v
enue as a % of
GDP
2B: Top Marginal Income Tax Rate & Threshold
at Which it
Applies
2C: Indirect Tax
Re
v
enue as a % of
GDP
2D: Sales
Taxe s
Collected as a % of GDP
Area 3: Labor Market Freedom
3A: Mini
m
um
W
a
g
e
Le
g
islation
3B:
Go
v
ernmen
t
Share of Total
Emplo
y
men
t
3C: Union
Densi
t
y
8 Economic freedom data from 1981 are assigned to 1979 in order
to match the availability of data for other variables in the analysis.
46 Bennett and Vedder
Despite the advantages mentioned above, there
are several limitations to the EFNA data. First, eco-
nomic institutions and policy have a tendency to
change slowly over time. We attempt to overcome
this issue by using quintennial rather than annual
data.9 Next, economic institutions that may exert an
impact on inequality, such as the protection of prop-
erty rights, monetary policy, and trade openness are
relatively homogenous across states. For instance,
the Commerce Clause of the U.S. Constitution pre-
vents individual states of the union from imposing
trade restrictions, and the Federal Reserve controls
the money supply for the entire country. Berggren
(1999) found trade liberalization to be one of the
most important determinants of economic freedom
in reducing inequality, and Scully (2002) indicated
that inflation and inequality are positively related.
Thus, to the extent that economic institutions are
constant across states, the estimated effects of eco-
nomic freedom on income inequality may not have
external validity beyond the United States. Finally,
the EFNA index does not measure any of the nu-
anced regulatory differences across states that may
exert an influence on income inequality. The Merca-
tus Center's Freedom in the 50 States report contains
more comprehensive information on such regula-
tions, but unfortunately it only began to measure
economic freedom over the last few years, which is
not a sufficient duration for a dynamic analysis.
4. Methodology and empirical results
First, the level of income inequality is regressed
on the 10-year change in economic freedom, ,
using fixed state, , and time,, effects, and a
number of contemporaneous control variables, ,
such as per capita income, the unemployment and
college attainment rates, the share of the population
over the age 65, and the Hispanic share of the
population10:
 

 
 

 
 (1)
The results of estimating equation 1, reported in
Table 2, suggest that increases in economic freedom
over the preceding 10 years are associated with
91981 is the first observation year, followed by 1984. Quintennial
data after 1984 (years ending in 4 or 9) are used in order to align
with the income inequality data, which is only available until
2004.
10 Summary statistics for the variables used in the estimations are
reported in the appendix.
lower levels of income inequality in the current
period. Columns (2) and (3) provide a robustness
check of this result using decennial Census measures
of household and family income Gini coefficients as
the dependent variable, respectively.11
The results suggest that a state which increased
economic freedom by a single point, or 1.6 standard
deviations, over the course of the preceding decade
has a family income Gini measure 0.005 lower than a
state with the same initial level of economic freedom
but whose rating remained unchanged over the
period, all else equal. In other words, a 1 standard
deviation increase in economic freedom over the
preceding decade is associated with a 0.104 standard
deviation lower family income Gini coefficient in the
current period, ceteris paribus. Although the point
estimates differ, the two alternative measures of
income inequality reflect the same qualitative and
statistically significant results, suggesting that the
findings are robust.12
Next, we employ two dynamic fixed effects
distributed lag panel models to test whether changes
in economic freedom are associated with changes
in income inequality. Equation 2 provides the gen-
eral structure for these models, which regresses the
5- and 10-year changes in income inequality,∆,
on the initial level of inequality, ,, 5-year interval-
lic changes in economic freedom, including a
lagged change, ∆, where γ is a 1 vector of
coefficients:13
∆ 

, 
 

 
 (2)
The results from estimations of the two dynamic
variations of equation 2 augment the evidence
reported in Table 2. Lagged increases in economic
freedom are associated with reductions in income
inequality. Column 1 of Table 3 reports the
11The Census measures are available decennially from 1979-1999.
Household income Gini measures are available annually from the
American Community Survey beginning in 2006. An average of
the 2006-2010 household income measures is assigned to 2009,
providing an additional observation for it.
12 Estimates given are for the GH (2006) family income Gini
measures. For the Census family and household income Gini
measures, the partial effects are -0.007 and -0.003, suggesting that
a 1 standard deviation increase in economic freedom over the
preceding decade is associated with a 0.163 and 0.08 standard
deviation decline of income inequality, respectively, for the ob-
servations included in the sample used to estimate equation 1
with the respective measures as the dependent variable.
13 k is the number of intervallic 5-year changes in economic free-
dom. k=2 for∆, and k=3 for∆.
Economic Freedom and Income Inequality 47
estimates of equation 2 when 5-year changes in
inequality, , are regressed on the initial level of
inequality, the corresponding 5-year change in eco-
nomic freedom, , and the lagged quintennial
change in economic freedom, . The sign on
both dynamic variables is negative, but only the
lagged change in economic freedom is statistically
significant.
Table 2. Fixed effects regression estimates of equation 1.
(1)
Family (GH)
(2)
Household
(3)
Family (Census)
10EF -0.005* -0.003** -0.00
7
***
(0.002) (0.001) (0.002)
Mean Income 0.138 0.164*** 0.022
(0.134) (0.055) (0.137)
Unemployment -0.065 0.159** 0.049
(0.120) (0.066) (0.123)
College 0.562** 0.284*** 0.389**
(0.234) (0.074) (0.173)
Senior 0.335 0.399*** 0.146
(0.201) (0.102) (0.251)
Hispanic 0.323*** 0.054 0.26
7
***
(0.071) (0.038) (0.057)
d2009 -0.026***
(0.006)
d2004 -0.025*
(0.013)
d1999 -0.016* -0.002 -0.005
(0.008) (0.003) (0.008)
d1994 -0.012***
(0.004)
Intercept 0.185*** 0.261*** 0.280***
(0.057) (0.023) (0.052)
R-squared 0.792 0.873 0.869
N 200 150 100
***Statistically significant at 99% level; **95% level; *90% level.
Fully robust standard errors are in parentheses. Mean income per capita data from t
he
Bu-
reau of Economic Analysis and in constant 2011 $10,000’s of dollars.
Unemployment
rat e
data are from the Statistical Abstract of
the
United States. College is the share of the adult
(25+ years) populat ion with 4 or more years of college, Senior is the 65+ sha re of the
population, and Hispanic is
the
share of the population of Hispanic descent. Data for the
latter three variables are
fr
om
the
decennial Censuses of the Population, with missing
years interpolated using the
annual
compound growth rate between
Censuses.
48 Bennett and Vedder
Table 3: Dynamic regression estimates of Equation 2.
(1)
5Year
Gini
(2)
10Year
Gini
(3)
Gini
1984-2004
(4)
Gini
1979-1999
(5)
Gini
1979-2004
Ginit5 -0.190
(0.154)
Ginit10 -0.399*
(0.202)
Ginit20 0.454* 0.30
7
*
(0.189) (0.137)
Ginit25 0.399*
(0.189)
5EF -0.002 0.00
7
0.004 0.030** -0.006
(0.002) (0.005) (0.013) (0.009) (0.013)
510EF -0.008***
-0.005* 0.031** -0.008 0.034**
(0.002) (0.002) (0.011) (0.009) (0.012)
1015EF -0.008***
(0.002)
1020EF -0.004 0.003
(0.007) (0.004)
1025EF 0.004
(0.005)
d2004 0.010***
0.018***
(0.004) (0.003)
d1999 0.010***
0.009***
(0.002) (0.002)
d1994 -0.001
(0.001)
mw -0.008 0.006 -0.00
7
(0.014) (0.012) (0.014)
ne 0.021 0.019 0.019
(0.015) (0.011) (0.015)
west 0.008 0.019 0.00
7
(0.014) (0.012) (0.014)
south -0.010 0.00
7
-0.004
(0.014) (0.011) (0.014)
Intercept 0.081 0.168** -0.141 -0.099 -0.112
(0.059) (0.077) (0.072) (0.053) (0.072)
R-squared 0.433 0.609 0.48
7
0.529 0.47
7
N 200 150 50 50 50
Model FE FE OLS OLS OLS
***Statistically significant at 99% level; **95% level; *90% level. Ful ly robust standard
errors are in parentheses for the first 2 columns, with normal
standar
d
err
ors
for the
latter 3
c
olumns.
Economic Freedom and Income Inequality 49
When the 10-year changes in inequality,∆, are
regressed on the initial level of inequality and the
previous three quintennial changes in economic
freedom, we find similar results. Lagged increases
in economic freedom, , as well as increases
during the first 5 years following the start of the con-
temporaneous period of measure, , are sta-
tistically significant and negatively associated with
income inequality over a 10-year period. The mag-
nitude of the marginal effect of the latter variable is
greater than that of the former. Recent changes in
economic freedom,∆, do not exert a statistically
significant effect, although the sign is positive, on
the 10-year change in inequality. These results are
reported in column 2 of Table 3. Together, these two
dynamic models suggest that increases in economic
freedom lead to lower income inequality, but the
effect takes time to be realized.
We also examined the dynamics between chang-
es in economic freedom and changes in income ine-
quality over a longer time period using ordinary
least squares regression. The 20-year changes in
income inequality, , are regressed on initial in-
come inequality and changes in economic freedom
in the two most recent 5-year intervals,  and
, change in economic freedom during the
first decade following the initial period, ,
and regional dummy variables, as defined by the
Census Bureau, to control for potential region-
specific effects. Columns 3 and 4 of Table 3 report
the results using 1984 and 1979, respectively, as the
initial period. The results indicate that increases in
economic freedom over the 1994-1999 period are
associated with the long-run increase in income ine-
quality, but that changes during the other intervals
are not statistically significant. A single unit in-
crease in economic freedom during the 1994-1999
period is associated with 0.03 and 0.031 point in-
creases in the income Gini over the twenty year pe-
riods ending in 1999 and 2004, respectively. In other
words, a 1 standard deviation rise in economic free-
dom from 1994-1999 is associated with 0.691 and
0.714 standard deviation rises in income inequality
over the 1979-1999 and 1984-2004 periods, respec-
tively.14
Column 5 of Table 3 reports the estimates when
the 25-year change in inequality, , is regressed
on the initial level of inequality, the two most recent
5-year changes in economic freedom, and changes in
14 The analysis uses the standard deviation of economic freedom
for observations from 1994 and 1999 only and the standard devia-
tion of family income Gini for observations from 1979-2004.
economic freedom for the first 15 years after the
initial period, . The results again suggest
that increases in economic freedom over the 1994-
1999 period are associated with the long-run growth
in income inequality. The results from the last three
regressions suggest that changes in the economy in
the latter part of the 1990s are driving the long-run
results. This period marked the rapid expansion of
the technology sector and an above average 4% real
annual growth rate of the U.S. economy. Economic
freedom and income inequality both increased dur-
ing this period, with 39 of 50 states experiencing an
increase in the former and all 50 states an increase in
the latter. Unfortunately, data on economic freedom
prior to the 1980s is not available. As such, we can-
not examine whether changes in economic freedom
prior to this time exerted an effect on the long-run
change in inequality.
5. Towards an alternative explanation
Our results are suggestive that increases in eco-
nomic freedom are associated with reductions in
inequality, but that the changes in the former take
time to exert an effect on the latter. In other words,
there is a lag between when economic freedom is
enhanced and income inequality declines. The long-
run dynamic regressions muddy this relationship
somewhat, driven by changes to the economy in the
1990s coinciding with the technology boom. Reflect-
ing on this evidence, perhaps the relationship be-
tween economic freedom and income inequality is
not a linear one. Economies are, after all, in different
stages of development, and they have different eco-
nomic institutions and polices in place at any given
point in time.
Simon Kuznets (1955) famously theorized that as
economies grow inequality rises until a certain level
of income is reached and inequality begins to fall,
suggesting that the benefits of growth initially
accrue to the upper end of the income distribution
before trickling down to the lower part of the distri-
bution. Assuming that the Kuznets relationship
holds, one might expect that the same inverted
U-shape relationship exists between economic free-
dom and income inequality, since the former has
been empirically shown to be a positive determinant
of economic growth (cf. Knack and Keefer, 1995;
Berggren, 2003; Dawson, 2003; De Haan, Lundström,
and Sturm, 2006; Gwartney, Holcombe, and Lawson,
2006; Hall, Sobel, and Crowley, 2010; Rode and Coll,
2011).
50 Bennett and Vedder
Thus it is plausible that starting from low levels
of economic freedom, enhancements would induce
growth and provide new economic opportunities
that initially benefit the upper part of the income
distribution more so than the lower part since in-
vestments would likely originate from those with
the physical and human capital necessary to launch
an enterprise or engage in trade. This would result
in an increase in income inequality.15 As economic
freedom continues to expand, growth continues,
providing new economic opportunities to those pre-
viously lacking the capital to take advantage of
emerging economic opportunities. Eventually,
greater economic freedom should result in greater
benefits accruing to the lower part of the distribu-
tion relative to the upper part, resulting in an in-
crease in equality. Berggren (1999) and Ashby and
Sobel (2008) both found evidence that the income-
enhancing effects of positive changes in economic
freedom benefit the bottom of the income distribu-
tion more so than the top over time. Proposition 1
describes this possibility.
Proposition 1: An inverted U-shaped relationship
exists between economic freedom and income
inequality. That is, beginning from low levels of
economic freedom, increases initially lead to more
income inequality, but as enhancements to eco-
nomic freedom continue, an inflection point is
reached such that additional increases lead to more
income equality.
To test proposition 1, we use the static fixed effects
model given by equation 3:
 

 
 

 
 (3)
The results, reported in Table 4, provide evidence
of the existence of an inverted U-shaped relationship
between economic freedom and income inequality
as the coefficients on the linear and quadratic eco-
nomic freedom terms are positively and negatively,
respectively, associated with income inequality. To
check the sensitivity of the results to the measure of
inequality and the time period of data availability,
three different measures of income inequality are
used as the dependent variable in equation 3. All
generate similar statistically significant results, sug-
15Although income inequality is likely to increase during the ini-
tial stages of economic liberalization and growth, other measures
of inequality such as consumption and standard of living may
decline. Data on such measures are limited and beyond the scope
of the present study.
gesting that the parabolic relationship is robust. The
inverted U-shaped relationship depicted by these
estimates is opposite the finding of Carter (2006),
adding complexity to our already limited under-
standing of the relationship between economic free-
dom and income inequality.
The inflection point at which additional increases
in economic freedom are associated with less income
inequality are reported in the last row of Table 4 for
each alternative measure of inequality. Figure 2
plots the average predicted income Gini measure
against the average economic freedom score for the
states using the Galbraith and Hale (2006) family
income Gini measures, depicting an inverted U-
shaped parabolic relationship between the two vari-
ables. The inflection point is 7.319, suggesting that
states with an economic freedom score below this
level will experience an increase in inequality when
economic freedom expands, whereas states with
economic freedom above this level will experience
reductions in inequality for additional increases in
economic freedom.16 We computed the average
economic freedom rating by state over the 1979-2004
period and found that 21 of the 50 states have an
average rating above the inflection point. Proposi-
tion 1 suggests that additional increases in economic
freedom in these states would generate more income
equality.17
In order to further test proposition 1, we sepa-
rately estimate equation 1 for states with economic
freedom above and below the inflection point in
1979. The results, reported in Table 5, indicate that
increases in economic freedom over the preceding
decade are significantly associated with lower in-
come inequality for states with economic freedom
above the inflection point in the initial period.
Meanwhile, the sign on the 10-year change in eco-
nomic freedom variable is negative for the states
with an initial level of economic freedom below the
inflection point, but the coefficient is highly insignif-
icant. Compared to the results obtained for the en-
tire sample, reported in column 1 of Table 2, the par-
tial effect of  on income inequality is much
stronger for the subsample of states with initial eco-
nomic freedom above the inflection point. These
results lend empirical support to the validity of
proposition 1.
16The mean economic freedom over states and time included in
the sample is 7.092, with a standard deviation of 0.695.
17For the interested reader, the states with an average economic
freedom rating to the right of the inflection point are AL, AZ, CO,
DE, FL, GA, IN, IA, LA, MS, MO, NE, NV, NH, NC, SC, SD, TN,
TX, VA, and WY.
Economic Freedom and Income Inequality 51
Table 4. Parabolic FE regression estimates of equation 3.
(1)
Fa mi l y
(GH)
(2)
Household
(3)
Fa mi l y
(Census)
E
F
0.06
7
** 0.051** 0.086***
(0.030) (0.021) (0.028)
EF2 -0.005** -0.003** -0.006***
(0.002) (0.001) (0.002)
d2009 0.049***
(0.002)
d2004 0.045***
(0.003)
d1999 0.034***
0.044***
0.051***
(0.002) (0.002) (0.002)
d1994 0.01
7
***
(0.001)
d1989 0.011***
0.026***
0.032***
(0.002) (0.002) (0.002)
d1984 0.00
7
***
(0.001)
In
tercept
0.13
7
0.211** 0.05
7
(0.107) (0.077) (0.101)
R-squared
0.778 0.901 0.900
N
300 200 150
EF Inflection
P
oin
t
7.319 7.484 7.138
***Statistically significant at 99% level; **95% level; *90% level. Fully robust standard errors are in
p
ar
entheses.
Figure 2. Income inequality vs. economic freedom.
52 Bennett and Vedder
Table 5. Estimates of equation 1, controlling for initial EF.
(1)
Above Inflection, 1979
(2)
Below Inflection, 1979
Δ10EF -0.00
7
** -0.001
(0.003) (0.003)
Unemployment -0.151 -0.018
(0.148) (0.166)
Mean Income 0.140 0.119
(0.163) (0.224)
College -0.045 0.973**
(0.276) (0.314)
Senior -0.05
7
0.528**
(0.350) (0.207)
Hispanic 0.159 0.423***
(0.102) (0.082)
d6 0.013 -0.052***
(0.016) (0.010)
d5 0.008 -0.032***
(0.011) (0.007)
d4 0.003 -0.021***
(0.005) (0.004)
Intercept 0.369*** 0.072
(0.074) (0.060)
R-squared 0.840 0.821
Number of States 1
7
33
Observations 68 132
***Statistically significant at 99% level; **95% level; *90% level. Ful ly robust standard errors
are in parentheses. The sample in column (1) is limited to
states
with an eco n omic f reedom
score above 7.319 in 1979, whereas the sample in column
(2)
includes states with economic
freedom below this level in 1979. Note that
ec
onomic
fr
ee
dom in 1981 is assigned
to
1979.
6. Summary
In this study, the dynamic relationship between
economic freedom and income inequality for the
fifty U.S. states over the 1979-2004 period is ana-
lyzed. Previous literature examining the relation-
ship between income inequality and economic free-
dom has been inconclusive. Most authors have ex-
amined the relationship between the two variables
using a linear framework, with these studies (Berg-
gren, 1999; Scully, 2002; and Ashby and Sobel, 2008)
suggesting that increases in economic freedom are
associated with enhancements of income equality.
Our results tend to support this finding and are
robust to alternative measures of income inequality
and various time periods. Carter (2006) offered a
critique of the interpretation of the previous results
in suggesting that there is a policy trade-off between
economic freedom and income inequality, finding
evidence of the existence of a U-shaped curve be-
tween the two variables in an international analysis.
The current analysis of the U.S. states adds
further complexity to the discussion, as we find evi-
dence of an inverted U-shaped curve between
income inequality and economic freedom. This sug-
gests that beginning from a low level of economic
freedom, increases initially generate more inequality
as the upper part of the income distribution benefits
Economic Freedom and Income Inequality 53
relatively more than the lower part; however, as
enhancements of economic freedom continue, this
reverses and the lower part of the distribution expe-
riences larger relative income gains. This finding is
also robust to alternative measures of income ine-
quality. The preponderance of evidence tends to
support this proposition, which is economically in-
tuitive given that enhancements of economic free-
dom lead to greater growth and development,
which in turn may initially act to increase income
inequality before the benefits trickle down the
income distribution and result in more income
equality.
It should be noted that our results pertain to the
United States and may not extend to an international
analysis. The measure of economic freedom used
can be roughly thought of as only accounting for
differences across states in fiscal and labor market
policies, holding constant other aspects of economic
freedom such as property rights and legal structure,
and monetary and trade policies, which may be im-
portant determinants of the distribution of income.
This is a relatively good depiction of reality in the
U.S., as the aforementioned institutional arrange-
ments are maintained at a national level, and as such
are relatively homogenous across states. The lack of
variation across states in macro-level economic insti-
tutions could explain why our findings are opposite
those reported by Carter.
Our results add to the discussion concerning the
relationship between economic freedom and income
inequality but are far from the final word on the
matter. We suspect that this line of research will
grow in the coming years, as the two variables are of
significant concern among policymakers not only in
the U.S., but also around the world. As such, addi-
tional research is needed to better understand the
relationship between the two variables in order to
better guide public policy.
Acknowledgements
The authors are grateful for useful comments
from James Gwartney, Josh Hall, Martin Rode,
and two anonymous referees. An earlier ver-
sion of this paper was presented at the 2012
Association of Private Enterprise Education
conference in Las Vegas, NV.
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Economic Freedom and Income Inequality 55
Appendix
Table A1. Summary statistics.
V
ariable
Mean
Std.
Dev.
Min.
Max.
N
Family Income Gini (GH) 0.398 0.028 0.338 0.502 300
Δ5 GH 0.009 0.009 -0.018 0.052 250
Δ10 GH 0.019 0.014 -0.004 0.094 200
Δ20 GH 0.038 0.019 0.007 0.115 100
Δ25 GH 0.047 0.022 0.011 0.12 50
Family Income Gini (Census)
0.39
0.031
0.33
0.472 150
HH Income Gini (Census) 0.431 0.028 0.371 0.5 200
Economic Freedom (EF) 7.08 0.682 5.101 8.673 350
Δ5EF 0.014 0.324 -1.078 0.935 300
Δ10 EF 0.034 0.424 -1.384 1.3 250
Δ510 EF 0.047 0.33 -1.078 0.935 250
Δ1015 EF 0.065 0.351 -1.078 0.935 200
Δ1020 EF 0.066 0.47 -1.384 1.3 150
Δ1025 EF 0.098 0.498 -1.23 1.352 100
Unemployment Rate 0.06
0.022
0.025
0.15 350
Mean Income 0.336 0.067 0.194 0.562 350
College Attainment Rate 0.215 0.054 0.106 0.382 350
Senior Share Population 0.123 0.021 0.03 0.185 350
Hispanic Share Population 0.066 0.084 0.005 0.456 350
Economic Freedom data are from the Fraser Institute’s annual
Ec
onomic
Freedom of North America report. Note that EFNA data from 1981 are assigned to
1979
in
order to
match up with the availability of data for other variables. Unemployment
r
ate
is
the average
monthly unemployment rate for a given state and data are from
the
Statistical Abstract of the United
States. Mean income is personal income per capita,
and
data are from the Bureau of Economic Analy-
sis. The college
attainment
rate is the
shar
e
of
adults (25+ years of age) with 4+ years of college.
Senior share of population is
the
percentage of a state’s population age 65 and
over.
College
attain-
ment,
senior, and Hispanic data are from the U.S.
Census
Bureau.
XX
X
;XX
X
.
... These earlier studies, however, are hampered by low coverage on GINI scores-roughly 30 countries, covering mostly the higher-income countries. Others, studying the variation among the 50 states within the United States report a parabolic relationship, where economic freedom increases inequality and then reduces it at the highest levels (Bennett and Vedder, 2013). Other, more recent work, also using the WIID for an expanded set of countries, finds that economic freedom, particularly freedom to trade, raises inequality, but only among rich countries (Bergh and Nilsson, 2010a). ...
... Other, more recent work, also using the WIID for an expanded set of countries, finds that economic freedom, particularly freedom to trade, raises inequality, but only among rich countries (Bergh and Nilsson, 2010a). The most recent study, using far greater coverage and using the standardized GINI scores (SWIID), reports a curvilinear effect between economic freedom and inequality, where higher values of economic freedom above a threshold reduce income inequality-results opposite to that reported by Carter but consistent with studies using only the United States (Bennett and Vedder, 2013;Bennett and Nikolaev, 2017). These studies distinguish between the short-run inequality enhancing effects from the long-run inequality-reducing effects of economic freedom. ...
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The debate about the merits of the redistributive state generally focuses on the proper trade-offbetween more income equality and more total income. The questions are familiar. What is the optimal distribution of income? How much inequality should a just society accept? How much economic growth should we sacrificein order to promote the welfare ofthe poor anddisadvantaged? Ofcourse, trans- fer programs go well beyond the poor. They are aimed variously at farmers whose crops have failed or whose crops "should" sell for more, at governments of nations whose people are worthy ofaid, at corporations that employ many workers and are in financial trouble, and atinnumerable other entities. The proposition that transfer policies promote the welfare of tar- geted groups is generally accepted. Taxes, transfers, and regulatory policies are perceived to be adjustment levers available to fine-tune the economic machine that grinds out goods and services. Ifwe do notlike the allocation ofeconomic benefits, corrective action can be undertaken by moving the levers via the political process. We believe this view of the transfer society is naive. Taxpayers andtransferrecipients arehumanbeings, notsheep whocanbe shorn at will, their wool automatically growing back for the next shearing season. People will adjust their actions for individual advantage, in responseto governmental changes inthe rulesofthegame. Similarly, since the political process, like the market, results from individual choices, itmayor maynotyield its statedgoals, Thus, itis notobvious that income transfers emanating from the political process will pro- mote economic equality or even help the targeted groups.
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The impact of property rights on economic growth is examined using indicators provided by country risk evaluators to potential foreign investors. Indicators include evaluations of contract enforceability and risk of expropriation. Using these variables, property rights are found to have a greater impact on investment and growth than has previously been found for proxies such as the Gastil indices of liberties, and frequencies of revolutions, coups and political assassinations. Rates of convergence to U.S.-level incomes increase notably when these property rights variables are included in growth regressions. These results are robust to the inclusion of measures of factor accumulation and of economic policy.
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The international development community has encouraged investment in physical and human capital as a precursor to economic progress. Recent evidence shows, however, that increases in capital do not always lead to increases in output. We develop a growth model where the allocation and productivity of capital depends on a country’s institutions. We find that increases in physical and human capital lead to output growth only in countries with good institutions. In countries with bad institutions, increases in capital lead to negative growth rates because additions to the capital stock tend to be employed in rent-seeking and other socially unproductive activities.
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Empirical studies provide evidence that economic freedom, as measured by the Economic Freedom of the World Index, is related to economic growth. None the less, identifying which aspects of economic freedom are more conducive to growth has proven difficult, due to multicollinearity among the index areas. A possible explanation is that certain countries score high in all areas, whereas others tend do bad in all of them, simply because the former are more freedom-friendly than the latter. However, it is also true that each country presents a combination of freedoms, and restrictions to freedom, at the level of the individual indicators that make up each area. If some regularity exists with respect to these combinations, empirical detection of the most popular policy combinations would alleviate the collinearity problem, when assessing growth effects. Our article explores this possibility by means of cluster analysis, which we conduct at the individual indicator level. We show that multicollinearity can indeed be reduced in this way and identify policy packages that seem to be more conducive to economic growth than others. Results further indicate that certain policy packages may have only a short-term effect on growth, whereas others seem to have an enduring one. KeywordsEconomic freedom–Economic growth–Aggregation–Cluster analysis
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Interest in religious organizations as providers of social services has increased dramatically in recent years. Churches in the U.S. were a crucial provider of social services through the early part of the twentieth century, but their role shrank dramatically with the expansion in government spending under the New Deal. In this paper, we investigate the extent to which the New Deal crowded-out church charitable spending in the 1930s. We do so using a new nationwide data set of charitable spending for six large Christian denominations, matched to data on local New Deal spending. We instrument for New Deal spending using measures of the political strength of a state's congressional delegation, and confirm our findings using a different instrument based on institutional constraints on state relief spending. With both instruments we find that higher government spending leads to lower church charitable activity. Crowd-out was small as a share of total New Deal spending (3%), but large as a share of church spending: our estimates suggest that benevolent church spending fell by 30% in response to the New Deal, and that government relief spending can explain virtually all of the decline in charitable church activity observed between 1933 and 1939.