More Guns Less Crime
Page 12
The results in table 5.4 for Pennsylvania and table 5.5 for Oregon provide a couple of consistent patterns. 7 The most economically and statistically important relationship involves the arrest rate: higher arrest rates consistently imply lower crime rates, and in twelve of the sixteen regres-
Table 5.4 Crime and county data on concealed-handgun permits: Pennsylvania counties with populations greater than 200,000
Crimes per 100,000 population
Percent change in the crime rate
Violent crime
Murder Rape
Aggravated Property Auto
assault Robbery crime theft
Burglary Larceny
Due to a 1 percent change -5.3%** -26.1%* -5.1%** -4.8%**
in the number of right-to-carry pistol permits/ population over 21 between 1988 and each year since the law was implemented
Due to a 1 percent change -0.79%* -0.37%* -0.08% -0.76%*
in the arrest rate for the crime category
1.2%
-0.12%
1.5%
-1.4%
0.7%
-0.84%* -0.41%** -0.065% -1.1%*
0.13%
Note: While not all the coefficient estimates are reported, all the control variables are the same as those used in table 4.1, including year and county dummies. All regressions use weighted least squares, where the weighting is each county's population. The nondiscretionary-law-times-county-population variable that was used in the earlier regressions instead of the variable for change in right-to-carry permits was tried here and produced very similar results. I also tried controlling for either the robbery or burglary rates, but I obtained very similar results.
*The result is statistically significant at the 1 percent level for a two-tailed t-test. **The result is statistically significant at the 10 percent level for a two-tailed t-test.
Table 5.5 Crime and county data on concealed-handgun permits: Oregon data
Percent change in the crime rate
Crimes per 100,000 population
Murder
Rape
Aggravated assault
Auto Robbery theft
Burglary Larceny
Due to a 1 percent change -37%**** -6.7% -4.8%
in the number of right-to-carry pistol permits/ population over 21 between 1988 and each year since the law was implemented
Due to a 1 percent change -0.34%* -1%* -0.4%*
in the arrest rate for the crime category
Due to a 1 percent change -0.2%* -0.09%* -1.5%*
in the conviction rate for the crime category
-4.7%
12%
2.7%
-0.4%*
-0.19%*
-.04%
-0.37%*
-0.7%*
-0.9%*
-0.27%* -0.86%*
Note: While not all the coefficient estimates are reported, all the control variables are the same as those used in table 4.1, including year and county dummies. I also controlled for sentence length, but the different reporting practices used by Oregon over this period make its use somewhat problematic. To deal with this problem, the sentence-length variable was interacted with year-dummy variables. Thus, while the variable is not consistent over time, it is still valuable in distinguishing penalties across counties at a particular point in time. The categories for violent and property crimes are eliminated because the mean sentence-length data supplied by Oregon did not allow us to use these two categories. All regressions use weighted least squares, where the weighting is each county's population. *The result is statistically significant at the 1 percent level for a two-tailed t-test. **The result is statistically significant at the 5 percent level for a two-tailed t-test. ***The result is statistically significant at the 10 percent level for a two-tailed t-test. ****The result is statistically significant at the 11 percent level for a two-tailed t-test.
sions the effect is statistically significant. Five cases for Pennsylvania (violent crime, murder, aggravated assault, robbery, and burglary) show that arrest rates explain more than 15 percent of the change in crime rates. 8 Automobile theft is the only crime for which the arrest rate is insignificant in both tables.
For Pennsylvania, murder and rape are the only crimes for which per-capita concealed-handgun permits explain a greater percentage of the variation in crime rates than does the arrest rate. However, increased concealed-handgun licensing explains more than 10 percent of the variation in murder, rape, aggravated assault, and burglary rates. Violent crimes, with the exception of robbery, show that greater numbers of concealed-handgun permits lower violent crime rates, while property crimes exhibit very little relationship. The portion of the variation for property crimes that is explained by concealed-handgun licensing is only about one-tenth as large as the variation for violent crimes that is explained by such licensing, which is not too surprising, given the much more direct impact that concealed handguns have on violent crime. 9 The regressions for Oregon weakly imply a similar relationship between concealed-handgun use and crime, but the effect is only strongly statistically significant for larceny; it is weakly significant for murder.
The Oregon data also show that higher conviction rates consistently result in significantly lower crime rates. The change in conviction rates explains 4 to 20 percent of the change in the corresponding crime rates; 10 however, for five of the seven crime categories, increases in conviction rates appear to produce a smaller deterrent effect than increases in arrest rates. 11 The greatest differences between the deterrent effects of arrest and conviction rates produce an interesting pattern. For rape, increasing the arrest rate by 1 percent produces more than ten times the deterrent effect of increasing the conviction rate for those who have been arrested by 1 percent. For auto theft, arrest seems more important than conviction: a 1 percent increase in the arrest rate reduces crime by about ten times more than the same increase in convictions. These results are consistent with the assumption that arrests produce large penalties in terms of shame or negative reputation. 12 In fact, the existing evidence shows that the reputational penalties from arrest and conviction can dwarf the legally imposed penalties. 13 This is some of the first evidence that the reputational penalties from arrests alone provide significant deterrence for some crimes.
One possible explanation for these results is that Oregon simultaneously passed both the nondiscretionary concealed-handgun law and a waiting period. The statistics in table 4.11 suggest that the long waiting period imposed by the Oregon law (fifteen days) increased murder by 5
VICTIMS AND THE BENEFITS FROM P ROTECTI ON / 107
percent, rape by 2 percent, and robbery by 6 percent. At least in the case of murder, which is weakly statistically significant in any case, the estimates from tables 4.11 and 5.5 together indicate that if Oregon had not adopted its waiting period, the drop in murder resulting from the concealed-handgun law would have been statistically significant at the 5 percent level.
The results for sentence length are not shown, but the t-statistics are frequently near zero, and the coefficients indicate no clear pattern. One possible explanation for this result is that all the changes in sentencing rules produced a great deal of noise in this variable, not only over time but also across counties. For example, after 1989, whether a crime was prosecuted under the pre- or post-1989 rules depended on when the crime took place. If the average time between when the offense occurred and when the prosecution took place differed across counties, the recorded sentence length could vary even if the actual time served was the same.
Florida's state-level data showing the changes in crime rates and changes in the number of concealed-handgun permits are quite suggestive (see figure 5.2). Cuba's Mariel Boat Lift created a sudden upsurge in Florida's murder rate from 1980 through 1982. By 1983 the murder rate had return to its pre-Mariel level, and it remained relatively constant or exhibited a slight upward trend until the state adopted its nondiscretion-ary concealed-handgun law in 1987. Murder-rate data are not ava
ilable for 1988 because of changes in the reporting process, but the available evidence indicates that the murder rate began to drop when the law was adopted, and the size of the drop corresponded with the number of concealed-handgun permits outstanding. Ironically, the first post-1987 upward movement in murder rates occurred in 1992, when Florida began to require a waiting period and background check before issuing permits.
Finally, a very limited data set for Arizona produces no significant relationship between the change in concealed-handgun permits and the various measures of crime rates. In fact, the coefficient signs themselves indicate no consistent pattern; the fourteen coefficients are equally divided between negative and positive signs, though six of the specifications imply that the variation in the number of concealed-handgun permits explains at least 8 percent of the variation in the corresponding crime rates. 14 This is likely to occur for several reasons. The sample is extremely small (only 64—89 observations, depending on which specification), and we have only a year and a half over which to observe the effect of the law. In addition, if Arizona holds true to the pattern observed in other states, the impact of these laws is smallest right after the law passes.
The results involving either the mean sentence length for those sen-
A. Murder rates
-6-4-2
-I h
Years before and after implementation of the law
B. Number of permits
-6 -4 -2 u 2 4 6
Years before and after implementation of the law
Figure 5.2A. Cumulative percent change in Florida's murder rate
Figure 5.2B. Concealed-handgun permits after implementation of the law in Florida
VICTIMS AND THE BENEFITS FROM P ROT E CTI O N / 109
tenced in a particular year or the actual time served for those ending their sentences also imply no consistent relationship between sentence length and crime rates. While the coefficients are negative in eleven of the fourteen specifications, they provide weak evidence of the deterrent effect of longer prison terms: only two coefficients are negative and statistically significant.
The Brady law also went into effect during this period. 15 Using the Arizona data to investigate the impact of the Brady law indicates that its only discernible effect was in the category of aggravated assault, where the statistics imply that it increased the number of aggravated assaults by 24 percent and the number of rapes by 3 percent. Yet it is important to remember that the data for Arizona covered only a very short period of time when this law was in effect, and other factors influencing crime could not be taken into account. While I do not believe that the Brady law was responsible for this large increase in assaults, I at least take this as evidence that the law did not reduce aggravated assaults and as confirmation of the belief that relying on this small sample for Arizona is problematic.
Overall, Pennsylvania's results provide more evidence that concealed-handgun ownership reduces violent crime, murder, rape, aggravated assault, and burglary. For Oregon, the evidence implies that murder and larceny decrease. While the Oregon data imply that the effect of handgun permits on murder is only marginally statistically significant, the point estimate is extremely large economically, implying that a doubling of permits reduces murder rates by 37 percent. The other coefficients for Pennsylvania and Oregon imply no significant relationship between the change in concealed-handgun ownership and crime rates. The evidence from the small sample for Arizona implies no relationship between crime and concealed-handgun ownership. All the results also support the claim that higher arrest and conviction rates deter crime, although—perhaps partly because of the relatively poor quality of the data—no systematic effect appears to arise from longer prison sentences.
Putting Dollar Values on the Crime-Reduction Benefits and Private Costs of Additional Concealed-Handgun Permits
By combining evidence that additional concealed handguns reduce crime with the monetary estimates of victim losses from crime produced by the National Institute of Justice, it is possible to attach a monetary value to the benefits of additional concealed-handgun permits. While the results for Arizona imply no real savings from reduced crime, the estimates for
Pennsylvania indicate that potential costs to victims are reduced by $5,079 for each additional concealed-handgun permit, and for Oregon, the savings are $3,439 per permit. As noted in the discussion of table 4.2, the results are largely driven by the effect of concealed handguns in lowering murder rates (with savings of $4,986 for Pennsylvania and $3,202 for Oregon). 16
These estimated gains appear to far exceed the private costs of owning a concealed handgun. The purchase price of handguns ranges from $100 or less for the least-expensive .25-caliber pistols to over $700 for the newest, ultracompact, 9-millimeter models. 17 The permit-filing fees can range from $19 every five years in Pennsylvania to a first-time, $65 fee with subsequent five-year renewals at $50 in Oregon, which also requires several hours of supervised safety training. Assuming a 5 percent real interest rate and the ability to amortize payments over ten years, purchasing a $300 handgun and paying the licensing fees every five years in Pennsylvania implies a yearly cost of only $43, excluding the time costs incurred. The estimated expenses are higher for Oregon, because of the higher fees and the costs in time and money of obtaining certified safety instruction. Even if these annual costs double, however, they are still quite small compared to the social benefits. While ammunition purchases and additional annual training would increase annualized costs, the long life span of guns and their resale value work to reduce the above estimates.
The results imply that handgun permits are being issued at much lower than optimal rates, perhaps because of the important externalities not directly captured by the handgun owners themselves. While the crime-reducing benefits of concealed handguns are shared by all those who are spared being attacked, the costs of providing this protection are borne exclusively by permit holders.
Accidental Deaths and Suicides
Even if nondiscretionary handgun permits reduce murder rates, we are still left with the question of what happens to the rates for accidental death. As more people carry handguns, accidents may be more likely. Earlier, we saw that the number of murders prevented exceeded the entire number of accidental deaths. In the case of suicide, the nondiscretionary laws increase the probability that a gun will be available when an individual feels particularly depressed; thus, they could conceivably lead to an increase in the number of suicides. While only a small portion of accidental deaths are attributable to guns (see appendix 4), the question remains whether concealed-handgun laws affect the total number of deaths through their effect on accidental deaths.
VICTIMS AND THE BENEFITS FROM P ROT E C T I O N / 111
To get a more precise answer to this question, I used county-level data from 1982 to 1991 in table 5.6 to test whether allowing concealed handguns increased accidental deaths. Data are available from the Mortality Detail Records (provided by the U.S. Department of Health and Human Services) for all counties from 1982 to 1988 and for counties with populations over 100,000 from 1989 to 1991. The specifications are identical to those shown in all the previous tables, with the exceptions that they no longer include variables related to arrest or conviction rates and that the variables to be explained are either measures of the number of accidental deaths from handguns or measures of accidental deaths from all other nonhandgun sources.
While there is some evidence that the racial composition of the population and the level of welfare payments affect accident rates, the impact of nondiscretionary concealed-handgun laws is consistently both quite small economically and insignificant statistically. The first estimate in column 1 implies that accidental deaths from handguns rose by about 0.5 percent when concealed-handgun laws were passed. With only 200 accidental handgun deaths nationwide during 1988 (22 accidental handgun deaths occurred in states with nondiscretionary laws), the implication is that enacting concealed-handgun laws in states that currently do not have the
m would increase the number of deaths by less than one (.851 deaths). Redoing these tests by adding together accidental handgun deaths and deaths from "unknown" types of guns produces similar results.
With 186 million people living in states without concealed-handgun laws in 1992, 18 the third specification implies that implementing such laws across those remaining states would have resulted in about nine more accidental handgun deaths. 19 Combining this finding with earlier estimates from table 4.1, we find that if the rest of the country had adopted concealed-handgun laws in 1992, the net reduction in total deaths would have been approximately 1,405 to 1,583.
One caveat should be added to these numbers, however: both columns 2 and 4 indicate that accidental deaths from nonhandgun sources increased by more than accidental deaths from handguns after the nondiscretionary concealed-handgun laws were implemented. To the extent that the former category increased because of uncontrolled factors that also increase accidental deaths from handguns, the results presented here are biased toward finding that concealed-handgun laws have increased accidental deaths from handguns.
Finally, I examined similar specifications using data on suicide rates. The possibility exists that if a person becomes depressed while away from home, the presence of a concealed handgun might encourage that person
Table 5.6 Did nondiscretionary concealed-handgun laws increase the number of accidental deaths? (1982-91 county-level data)
Deaths per 100,000 population
Change in explanatory variable
Nondiscretionary law adopted
Percent change in crime (for Tobit number of deaths per 100,000) for an increase in population of one person per square mile
Percent change in crime (for Tobit number of deaths per 100,000) for an increase in $1,000 of real per-capita personal income