Changes in young adult offense and crash patterns over time

Accident Analysis and Prevention 33 (2001) 117 – 128 www.elsevier.com/locate/aap

Changes in young adult offense and crash patterns over time Patricia F. Waller *, Michael R. Elliott, Jean T. Shope, Trivellore E. Raghunathan, Roderick J.A. Little 1779 Crawford Dairy Road, Chapel Hill, NC 27516, USA Received 17 June 1999; received in revised form 1 March 2000; accepted 13 March 2000

Abstract A study of 13 809 young adult drivers in Michigan examined offenses and crashes (‘incidents’) for an average of 7 years after their original license date. During this period, 73% of subjects committed an offense that resulted in a conviction and 58% had a crash that was reported to the police. Forty-two percent had committed an offense classified as ‘serious,’ and 21% had an ‘at-fault’ crash. The odds of an offense being serious decreased approximately 8% per year of licensure, independent of gender or age at licensure. Similarly, the odds of a crash being at-fault decreased overall about 6% per year of licensure, but the decline was more than twice as fast for women as for men. Examining the empirical rates directly, it was found that the rate for minor offenses increased somewhat with time and then stabilized, while the rate for serious offenses declined. Also, offenses were less likely to be serious the later they occurred in the sequence of offenses for an individual. For crashes, the risk of having an at-fault crash declined more rapidly than the risk of a not-at-fault crash, although the rate of decrease began to equalize after approximately 5 years of licensure. The proportion of crashes that were at-fault did not decline over the sequence of crashes for an individual. Although crashes and offenses are positively correlated, they follow different trajectories over the early years of licensure. © 2000 Elsevier Science Ltd. All rights reserved. Keywords: Youth; Risk-taking; Driving; Crashes; Violations; Offenses

1. Introduction It is well established that beginning drivers of any age show higher violation and crash rates in the early stages of licensure than they do after more experience has been acquired (Mayhew and Simpson, 1990; Maycock et al., 1991). In the US, almost all beginning drivers are also young drivers, and there has been continuing controversy about the major factors contributing to the elevated violation and crash rates of this group. Some consider age to be an important factor, if not the major factor, while others consider experience to be of more importance. There is evidence to support both these hypotheses. In the UK, where the youngest age at which license may be acquired is 17 but where many wait until they are older, Maycock et al. (1991) examined accident * Corresponding author. Tel.: +1-919-9423878; fax: + 1-9199628710. E-mail address: [email protected] (P.F. Waller).

liability (defined as the expected number of accident involvements per year) for beginning drivers of different ages. They found evidence that age was, indeed, important, and that delaying licensure from age 17 to 18 was associated with a 6% decrease in crash risk. A delay from age 18 to 19 resulted in an additional 6% decrease. Subsequent yearly delays were also associated with decreases in crash risk, although they were increasingly smaller. Experience was found to be even more important, at whatever age it occurred, with a 30% decrease in crash risk after the first year of licensure, followed by an additional 17% reduction after the second year of driving. As with age, additional years of experience led to further reductions that were increasingly smaller. In Sweden licensure may not be obtained until age 18, but until 1993 a novice driver could begin driving with supervision at age 17.5. In 1993 the age for permit was lowered to 16, with full licensure remaining at age 18. Evaluation of the impact found that those obtaining permit at 16 acquired more supervised practice, had no

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pre-licensure increase in crash risk, and had a 35% decrease in crash risk upon full licensure at 18 (Gregersen, 2000), giving credence to the value of extended supervised experience. In contrast, in an exhaustive review of available studies, Mayhew and Simpson (1990) concluded that ‘age related factors are more strongly associated with collision risk than are those that are experientially related’. These authors have also made a distinction between risk-taking behavior and risky behavior, with the former characterized by recognition of the risk involved. In contrast, risky behavior is behavior that, objectively, entails risk regardless of whether or not the perpetrator is aware of the risk. Young people are recognized for their greater involvement in risk-taking behavior. Their higher rates of traumatic fatal injury, including motor vehicle injury, reflect this characteristic. Based on such evidence, some experts have advocated simply raising the age of initial driver licensure so that driving will not be initiated until greater maturity is attained. In industrialized nations other than the US, driver licensure does not ordinarily occur until age 17 or 18. In the US opposition to such a proposal comes from both parents and vested economic interests. Parents are often eager to relinquish chauffeuring their youngsters, and grocery stores and fast food restaurants, that hire young people in large numbers, are concerned about their labor pool. In the US most people acquire licensure at age 16 or 17. Consequently age and experience are inextricably confounded, making it difficult to separate the relative effects of each. For example, the correlation between age and duration of licensure by subject-year in our analysis is r =0.92. However, it may be anticipated that experience would have a differential impact on certain kinds of violations and crashes, primarily those that are more likely to be attributable to inexperience, as opposed to deliberate risk-taking behavior. It is assumed that, in general, drivers do not deliberately get into a crash, not even drivers who engage in deliberate high-risk driving behavior. Although virtually all crashes are unintentional (excepting intentional suicide by motor vehicle), the same cannot be said of violations. Some violations, which we define as ‘serious,’ are sufficiently blatant that they can be assumed not only to be under the volitional control of the driver, but also deliberate, e.g., driving at 15 or more miles above the posted speed limit, and driving after drinking. Young drivers usually know that when they engage in such behaviors, they are breaking the law. Although such a distinction is less clear in regard to crashes, for purposes of this study it was assumed that an at-fault crash (a crash which is associated with a violation conviction) may generally be considered to be more attributable to deliberate risk-taking behavior than a not-at-fault crash.

Thus a distinction was made between behaviors that are more deliberate and under the volitional control of the young driver and behaviors involving less deliberate risky behaviors, with the assumption that the latter would be more affected by increasing driving experience. It was hypothesized that less serious offenses that could be assumed to be less volitional should decrease more rapidly as a function of experience than more serious and more obvious offenses. The basis for this hypothesis was that experience should be more beneficial for improving ordinary driving, but that violations that are more clearly identified as deliberate or ‘serious’ would be more a function of personal characteristics and less related to driving experience and thus would persist independent of experience.

2. Methods

2.1. Sample The study sample consisted of subjects in a longrange study of substance abuse prevention. Self-administered questionnaires were collected from 17 191 Michigan students in grades 5–12 between 1984 and 1994 (Shope et al., 1992, 1996). The population from which the sample of students was drawn included rural, suburban, and urban communities in southeast Michigan. Participation rates in the original surveys ranged from 79 to over 90%. Of the 17 191 students who completed questionnaires, 13 809 (80%) eventually obtained a driver license from the Michigan Department of State (DoS) between 1986 and 1997. Information for the analyses was obtained from driver history records, including traffic offenses and reported crashes for these licensed drivers during their entire tenure of licensing in Michigan. Early extractions from the driver history file were merged with later extractions, so that incidents that had been purged from the official record were retained in these analyses. The original survey data are not included in this analysis.

2.2. Measures The DoS data include information on gender, age, license tenure, number and type of violation convictions (referred to subsequently as offenses), and number of and details about reported crashes. Slightly more than one-half (52%) of subjects for whom driver history is available are male, with a mean age of 23.49 1.1 years at the time of the most recent driver history records used (4 June 1997). (The most recent 3 months on the driver history were omitted because of the time delays in getting violation convictions recorded.) Mean duration of licensure was 7.19 1.4 years, a uniquely long

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period for purposes of research. Ninety percent of all subjects held licenses between 4.8 and 9.0 years; the range for all subjects was from 8 days to 13.2 years. The ‘start of driving’ is defined as the original license (OL) date. The 770 offenses and the 129 crashes that occurred before the OL date were deleted; however all regression models include an indicator variable for whether or not an offense occurred before the OL date. The majority of pre-license offenses (63%) were for driving without a valid license. Crashes before the OL date without an offense are not included in order to maintain consistency with the start-of-driving definition. The primary outcomes of interest are offenses and crashes — collectively ‘incidents.’ Offense data are available only for offenses that resulted in convictions, although the original charge is used in this analysis to avoid biases from the subset of respondents whose resources might allow them to have ‘pled down’ to lesser charges. Likewise, only crashes reported to the police are included in the analysis. We focus on incidents indicative of risk-taking behaviors — ‘serious’ offenses and ‘at-fault’ crashes — to distinguish them

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from less serious offenses and not-at-fault crashes that may be more attributable to lack of driving experience or other factors, and to assess changes in risk-taking behavior as drivers become more experienced. A total of 36 407 offenses was recorded, an average of 2.69 3.3 per subject. Subjects averaged slightly less than one offense for every 2 years of licensure. Nearly three-fourths of subjects (73%) had one or more offenses recorded; one in five (18.6%) had five or more, while 1/20 (4.6%) had ten or more. ‘Serious’ offenses are those that meet any of the following criteria: (1) involve use of alcohol (14% of all serious offenses); (2) are classed as ‘serious’ by the Secretary of State’s office (e.g. reckless driving, vehicular homicide [11%]); (3) result in three or more points assigned to a driver (e.g. speeding in excess of 15 mph over the speed limit [73%]); or (4) involve non-driving drug offenses (2%). These offenses were considered to be more volitional and thus more indicative of risk-taking behavior than are lesser offenses. The remainder — typically driving without a license, or without proof of insurance, or speeding less than 15 mph over the speed limit — are classified as ‘minor.’ Table 1 shows that more than four

Table 1 Type of offense/crash, as a percentage of all offenses/crashes and as an overall and sex-specific percentage of subjects incurring one or more offense/crash type during follow-up period (mean follow-up =7.1 years) Type of Incident

Percent of all incidents

Percent of subjects incurring at least one incident All

Female

Male

Offenses

100.0

73.0

62.0

83.0

Serious Speeding \15 mph Disobeyed stop sign/signal OUI/DWIa Reckless drivingb Other

28.7 11.0 9.9 4.0 3.1 0.7

41.8 21.8 20.2 8.3 6.7 2.7

27.3 13.3 13.1 3.3 1.7 1.3

55.1 29.7 26.6 12.9 11.3 4.0

Minor Speeding 15 mph Failure to yield No proof of insurance Failure to display license Drove while unlicensed/suspended Other

71.3 40.9 2.9 6.1 4.2 6.6 10.6

65.3 54.9 7.2 12.1 8.3 9.2 19.0

54.7 44.6 6.3 7.8 4.3 4.2 10.5

75.1 64.3 8.1 16.0 11.9 13.9 26.8

Crashes

100.0

58.3

52.5

63.6

Number of 6ehicles One Two Three or more

15.6 74.8 9.6

13.4 48.8 8.9

9.8 43.2 7.7

16.7 56.8 10.0

Injury Fatality

30.2 0.2

24.7 0.2

21.8 0.1

27.4 0.2

Alcohol involved

3.5

3.1

1.2

4.8

At-fault

25.4

20.7

15.9

25.2

a b

Includes unlawful BAC, person under 21 with BAC, and violations for operating under controlled substances. Includes careless/felonious driving, manslaughter/homicide, fleeing and eluding, and false report violations

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in ten subjects committed at least one serious offense during the tracking period, and that nearly three in ten of all offenses were serious. Males were more likely than females to have committed any specific type of offense. The gender difference was greatest among the most severe offenses, such as operating under the influence of alcohol or drugs (OUI) or reckless driving. A total of 13 521 crashes was recorded, an average of 1.0 91.1 per subject. On average, one crash was observed per 7 years of license. Six in ten subjects (58%) had one or more crashes on record; 9% three or more, and 1% five or more. One in six (16%) were single-vehicle crashes; 75% involved two vehicles, and 10% three or more. Three in ten crashes (30%) had one or more persons injured; 25% of subjects (22% of females and 28% of males) were in a crash with at least one injury. Twenty-one crashes (0.2%) had one or more persons killed. Alcohol was recorded as a factor in 4% of all crashes; 3% of all subjects (1% of females and 5% of males) were in a crash in which alcohol was involved. ‘At-fault’ crashes — crashes in which the subject had an offense on the same date as the crash — made up 25% of all crashes. Twenty percent of all subjects (16% of females and 25% of males) incurred at least one at-fault crash. Crashes that were apparently ‘at-fault’ were considered to be the strongest crash indicator of risk-taking behavior. All subjects at the time of surveys were scheduled to graduate between 1991 and 1994, and thus were typically of an age to be licensed between 1988 and 1992. Because each ‘graduating class’ was not a random subsample of the sample — for example, the classes of 1993 and 1994 were drawn from fewer school districts — a control variable divided into 5 year-of-license cohorts was introduced: 1988 or earlier (n = 1178); 1989 (n =3856); 1990 (n = 4382); 1991 (n =2195); and 1992 or later (n = 2192). (Note that six subjects could not be classified due to lack of original licensure date data.) This variable will adjust for between-cohort differences.

3. Data analysis Analyses were conducted primarily through regression analyses using generalized linear models or survival analysis models, fitted using Statistical Analysis System (SAS) Version 6 for Windows (SAS Institute, Inc., 1998). Bivariate outcomes — risk of incurring an incident (offense or crash) and type of incident (serious versus non-serious or at-fault versus not-at-fault) — were modeled over the duration of licensure using logistic regression (Hosmer and Lemeshow, 1989). Because multiple incidents could be incurred for a given sub-

ject, outcomes are no longer necessarily independent, and estimated standard errors for regression coefficients may be biased. In particular, confidence intervals will often be too narrow and estimates of the probability of Type I error too small. Generalized estimating equation (GEE) methodology was used to account for the dependence among observations incurred by the same subject (Diggle et al., 1994) although in general the within-subject correlation was low. Standard error estimates were constructed using the empirical (sandwich-type) estimator. Log-linear row- and column-effects models (Agresti, 1990) were used to estimate the probability that incidents were serious, conditional on the total number of incidents incurred and the index of the incident (first incident, second incident, etc.). To test whether there is differential change in hazard rates for serious and minor offenses over time, a proportion hazards model was fit of the form: hm(t)= exp(b0 + b1t)hs(t)

(1)

where hs(t) is the hazard rate for ‘serious’ offenses and hm(t) is the hazard rate for ‘minor’ offenses. This model implies hm(t)/hs(t) is a constant with respect to time (proportional hazard) if and only if b1 =0. This is equivalent to fitting a logistic regression on whether the offense was minor or not, conditional on an offense occurring at time t (in years) (Cox and Oakes, 1984). A similar model is fit to compare whether the relative rates of at-fault and not-at-fault crashes change through time. Statistical significance was assessed using likelihood ratio tests for nested generalized linear models (McCullagh and Nelder, 1989) and Wald tests for GEEs (Diggle et al., 1994). All regression analyses have been checked for interactions with gender and with year of license cohort (1988 or before, 1989, 1990, 1991, and 1992 or later). Where gender or cohort interaction is significant at PB 0.05, separate results are reported for men and women and/or by cohort. In addition to these regression-type analyses, non-parametric summary measures of the estimated hazard rate of an incident of type k = {serious, minor} are given by hk(t)= n kp /Tp, where n kp is the number of incidents of type k in a given time period p from the start of driving and Tp is the total licensure time, given by  tpi, where tpi is the amount of time subject i was driving during period p. Confidence intervals for hk(t) and the proportion of incidents that are serious n m p /np are determined under the assumption that, conditional on the total duration of licensing in period p= Tp, the number of incidents of type k that occur (n kp ) follows a Poisson distribution, s and that conditional on np = n m p + n p, the proportion m of incidents that are serious (n p /np ) follows a binomial distribution.

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Table 2 Simultaneous effect of duration of licensing, gender, age at time of license, and early offenses (controlling for cohort) on risk of offense/crash within a single year and expected number of offenses/crashes within a given year [crude odds ratio] (95% confidence intervals)a Predictor

Percent change in odds of having one or more incidents within a 1-year period Offenses

Duration of license (year) Male Age at license (year) Pre-license offense

Crashes

All

Serious

All

At-fault

−8.7 [0.75] (−9.2–−8.1) 112.0 [2.22] (103.7–120.6) 9.0 [1.00] (6.6–11.5) 72.9 [2.13] (57.7–89.5)

−11.5 [0.65] (−12.3–−10.7)

−16.8 [0.50] (−17.5–−16.1)

−19.7 [0.44] (−21.0–−18.4)

163.3 [2.74] (148.7–178.8)

32.9 [1.35] (27.4–38.6)

62.5 [1.68] (50.5–75.4)

5.0 [1.00] (1.9–8.2) 83.1 [2.30] (64.3–104.0)

−5.3 [0.97] (−7.7–−2.8) 23.1 [1.28] (10.7–37.0)

11.4 [1.03] (6.9–16.0) 36.4 [1.64] (16.1–60.4)

a Within-subject dependence corrected using GEE, all PB0.01. [Crude odds ratios for continuous predictors calculated as the ratio between the expected odds of incident for subjects in the 75th percentile of the predictor and the 25th percentile of the predictor.]

4. Results

4.1. Relationship of risk of incident to duration of licensure, age and gender To see the effect of duration of licensure on the risk of committing an offense, we fit a logistic regression model with duration of licensure, gender, age at time of license, and dummy variables for existence of any prelicense offenses as predictors. We also controlled for calendar year of license (cohort). Results are shown in Table 2. All predictor variables are significant (PB 0.01); the effect of cohort appears to be approximately linear and is included as a linear effect. Simultaneously adjusting for all covariates, the odds of men committing an offense in any given year are more than twice as great as for women. The odds of subjects having an offense during a license year increased by about 9% for each year older the subject was at time of license. Increasing duration of licensure decreases the odds of a subject incurring one or more offenses in a year by 9% per year of licensing on average, although this effect was not strictly linear: the odds of an offense in the second year versus the first year increased by 28%, then stabilized and began falling after the third year of experience. The covariate effects of pre-license offenses, duration of license, and gender were more intense for serious offenses, with the effect of gender increasing 46%. However, age at time of license was more weakly, though still positively, associated. For crashes, being male or having a pre-license offense increases the odds of having one or more crashes within a 1-year period after adjusting for age, time-oflicense, and cohort, although the impact was not as strong as it was for the odds of committing an offense. In contrast to the odds of committing an offense, the odds of having a crash actually decline about 5% for each additional year of age at the time of licensing. Increasing duration of license decreases the odds of a

subject having one or more crashes in a year on average by about 17% per year of licensing, adjusting for gender, age at time of license, pre-license offenses, and cohort. This effect is stronger earlier, with a decline in odds of 22% between the first and second year and 19% between the second and third year. As with offenses, pre-license offenses, duration of license, and especially gender, are more strongly associated with at-fault crashes than with crashes in general. Age at time of licensure is positi6ely associated with at-fault crashes, that is, older initial licensees are at higher risk of having an at-fault crash. Significant interactions between gender and duration of license are found for the odds of offense, with the odds of an offense declining faster for men than for women (P= 0.0022). The odds of an offense being incurred in a given year drop by an average of 7% per year for women versus 9% per year for men. However, the effect of duration of license on the odds of a crash is the same for both men and women. Significant interactions between cohort and duration of license are found for offenses and crashes as well. Careful examination of the cohort interactions indicate that they are largely due to the non-linearity in the decreasing risk of incident accompanied by the average duration of licensure differing by cohort.

4.2. Relationship between serious and minor incidents through time Fig. 1 shows the proportion of male and female drivers with one or more minor offenses but no serious offenses (top of each bar), and one or more serious offenses and possibly minor offenses as well (bottom of each bar), by year of license, together with the total sample exposed for at least part of a license year, through 9 years of licensing. Also included in Fig. 1 are the equivalent proportions for not-at-fault and at-fault crashes.

122

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4.2.1. Summary hazard rates To examine the hypothesis that a higher proportion of offenses is due to risk-taking behavior than to inexperience as duration of licensing increases, we compare the serious offense hazard rate hs(t) with the minor offense hazard rate hm(t). If the hypothesis is correct, hs(t) will remain relatively constant while hm(t) declines, implying a non-proportional hazard ratio. Similarly we can examine the hazard ratio for at-fault versus not-atfault crashes. Comparing the non-parametric hazard rates for serious and minor offenses on 6-month intervals in the upper half of Fig. 2, we see that the rate for minor offenses climbed rapidly during the first 2 years after licensing, then stabilized and declined slowly thereafter. The serious offense rate appears to be more stable during early periods and then decline somewhat more rapidly. This effect is clearer when we examine the proportion of offenses that are serious in the lower half of Fig. 2, where we see that this proportion appears to decline more or less linearly with time. Also shown in Fig. 2 is that not-at-fault crashes declined during the first 3 years of driving experience

and then stabilized, whereas at-fault crashes dropped rapidly and then declined more slowly with time. Thus the proportion of at-fault crashes also declined slowly with time.

4.2.2. Test of proportional hazards Table 3 gives the effect of time on the ratio of the hazard rate of serious offenses to the hazard rate for minor offenses, adjusted for presence of pre-license offenses, cohort, gender, and age at time of license. For offenses, cohort effects appeared only between pre- and post-1992 cohorts, and thus were adjusted with an indicator variable for whether license was obtained prior to 1992 or not. Crashes are adjusted with a linear cohort effect. It appears that the odds of an offense being serious decreased by approximately 8% per year of driving time, while the odds of a crash being at-fault declined by about 6% per year. Men were more likely than women to have serious offenses rather than minor offenses and at-fault rather than not-at-fault crashes. After accounting for duration of licensure, older new licensees were less likely to have a serious offense but

Fig. 1. Total number of (a) female and (b) male subjects and proportions committing one or more minor and no serious/one or more serious offenses, by year of license tenure; total number of (c) female and (d) male subjects and proportions having one or more not-at-fault and no at-fault/one or more at-fault crashes, by year of license tenure.

P.F. Waller et al. / Accident Analysis and Pre6ention 33 (2001) 117–128

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Fig. 2. (a) Serious and minor offense rates by license tenure; (b) at-fault and not-at-fault crash rates by license tenure; (c) proportion of offenses that are serious by license tenure; (d) proportion of crashes that are at-fault by license tenure.

more likely to have an at-fault crash. This finding can also be deduced from Table 2, where it is observed that the effect of age at time of licensure is weaker (but still positive) on absolute risk of serious offense than on absolute risk of any offense, while age at time of license is positi6ely associated with risk of at-fault crash but negati6ely associated with risk of any crash. Having a pre-license offense appeared to have little impact on the type of post-license offense, but did appear at least somewhat positively associated with at-fault crashes. Fig. 3 shows the fit of the model-estimated probability that the offense will be serious, corrected for withinsubject correlation and unadjusted for other covariates, together with the observed proportion of offenses that are serious, in 6-month intervals. Also shown in Fig. 3 is the equivalent fit for at-fault crashes. Both models appear to fit reasonably well. The discontinuities among those observed the longest may be due to modest sample sizes in the group with the longest licensure. There was no evidence for interaction between the effect of duration of license and gender (P= 0.54) or

cohort (P= 0.55) on the odds of an offense being serious. However, the effect of time on reducing the odds of a crash being at-fault was twice as strong for women, declining by 9.8% per year versus 4.5% for men (P= 0.0060); no cohort effect was found (P= 0.52). Fig. 3 shows clearly that the odds of a crash being at-fault decline more rapidly for females than for males. Here ‘learning’ appears to be related to gender.

4.3. Relationship between serious and minor incidents by number of incidents Table 4 shows the percentage of offenses that were serious, conditional on the total number of offenses incurred and the ‘index number’ of the offense (first offense, second offense, etc.). If the first (post-license) offense is serious, there is increased risk of subsequent offenses. However, a ‘learning curve’ is evident in the declining probability that an offense will be serious among the later offenses. Fitting a log-linear model that assumes the probabilities in the Table 4 cells are con-

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Table 3 Simultaneous effect of duration of licensing, pre-license offenses, gender, and age at time of license, and early offenses (controlling for cohort) on odds of an offense being serious or crash being at-fault. [crude odds ratios] (95% confidence intervals)a Predictor

Percent change in odds Percent change in odds of an offense being of a crash being serious at-fault

Duration of −8.2** [0.76] license (year) (−9.2–−7.2) Male 32.2** [1.31] (25.2–39.7) Age at license −9.5** [0.96] (−12.3–−6.6) Pre-license −5.8 [0.85] offense (−14.7–4.0)

−6.3** [0.81] (−8.0–−4.6) 31.0** [1.64] (20.3–42.7) 22.1** [1.00] (16.1–28.3) 21.2* [1.89] (0.7–45.9)

a Within-subject dependence corrected using GEE. [Crude odds ratios for continuous predictors calculated as the ratio between the expected odds of incident for subjects in the 75th percentile of the predictor and the 25th percentile of the predictor.] * PB0.05. ** PB0.01.

stant and comparing it with a saturated model indicates that these apparent trends are likely to be real ( − 2L= 73.78 for 8 df, PB 0.0001). Fitting a log-linear rowand column-effects model indicates that the odds that an offense is serious decline by 8.4% (5.7 – 11.0%) relative to the previous offense (P-value for model fit= 0.56). Although multiple offenders show a steady decrease in the proportion of their offenses that are serious, even after three prior offenses this proportion remains higher than that for drivers with a single

offense. There was no evidence of differences between genders (P=0.61) or by cohort (P= 0.95). Here ‘learning’ appears not to be related to gender. Comparing the percentage of crashes that were atfault conditional on the total number of crashes incurred and the index of the crash in Table 5 shows no relationship between either the number of crashes or the index number of the crash ( − 2L = 2.45 for 9 df, P= 0.98). Since previous analysis indicates that there is a downward trend in the probability that later crashes are at-fault, this analysis of the early index of crashes indicates that duration of license (see Table 3) is a better predictor of whether a crash will be at-fault than crash experience itself.

5. Discussion

5.1. Age at licensure It has been recommended that age of initial licensure be increased so that new drivers are more mature. These analyses found a 5% reduction in total crash odds for each additional year of age at time of licensing, consistent with the 6% risk reduction reported by Maycock et al. (1991). However, the older initial licensees actually experienced an increase in the likelihood of an at-fault crash, as well as the proportion of their total crashes that were at fault. In contrast to crash experience, risk of offense increases with increasing age at time of licensure, but the

Fig. 3. Empirical and model-based probabilities that (a) offense is serious, or (b) crash is at-fault, by license tenure.

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Table 4 Percentage of offenses that were serious conditional on the total number of offenses incurred and the ‘index number’ of the offensea Percent of offenses that are serious

Total number of offenses incurred

Index of offense

1

2

3

4/+

First Second Third Fourth (N)

26.5 (26.9)

28.8 (29.2) 28.1 (27.4)

31.3 (31.7) 30.7 (29.8) 29.3 (28.0)

2948

2078

1497

34.0 (34.2) 32.3 (32.3) 31.1 (30.4) 27.5 (28.6) 3551

a Predicted percentage of serious offenses in parentheses assuming log-odds of offense being serious is linear in offense index and total number of offenses.

Table 5 Percentage of crashes that were at-fault conditional on the total number of crashes incurred and the ‘index number’ of the crasha Percent of crashes that are at-fault

Total number of crashes incurred

Index of crash

1

2

3

4/+

First Second Third Fourth (N)

25.4 (25.3)

25.1 (25.3) 25.3 (25.3)

24.2 (25.3) 24.9 (25.3) 25.3 (25.3)

4567

2189

846

27.1 27.3 25.6 24.7 454

(25.3) (25.3) (25.3) (25.3)

a Predicted percentage of at-fault crashes in parentheses assuming log-odds of crash being at-fault is independent of the crash index and total number of crashes.

probability that the offense will be serious decreases. The older initial licensee experiences a higher absolute risk of a serious offense, but the risk of non-serious offenses increases even more, so that the proportion of total offenses that are serious is lower. The odds of having an offense of any kind increase by about 9% for each year older the subject is at time of initial licensure. It is not clear why the risk for total crashes decreases while the risk for offense increases, especially when it is known that offenses and crashes are positively correlated (in these data, r =0.37). It is possible that beginning 18-year-old drivers drive more miles than beginning 16-year-olds, so that exposure to risk would be higher. Such a difference in exposure could account for the increases observed for offenses but not for the decrease in total crashes, unless the miles-driven-adjusted crash risk was dropping very rapidly.

5.2. Duration of licensure The length of time license has been held may be considered a crude measure of driving experience, although it is recognized that holding a license does not guarantee that driving is occurring and that lack of licensure does not mean that no driving is occurring. Nevertheless, it is assumed that time licensed is closely correlated with the miles a subject has driven, and that unlicensed persons account for fewer miles driven.

Length of licensure was found to be related to decreased crash risk, and especially a decrease in at-fault crashes. The odds of having one or more crashes in a year declined about 17% per year of licensing, adjusting for gender, age at time of license, pre-license offenses, and cohort. However, the initial decrease in crash risk was greater (22% in the first year and 19% in the second year) than that for succeeding years of licensure, somewhat lower than the 30% reported by Maycock et al. (1991) for the first year of licensure. However, Maycock et al. included reported estimates of mileage in their model, which were not available in this study. In contrast, Ferdun et al. (1967) found no trend in crash risk among California drivers age 16– 19, a group roughly comparable to those with 0–4 years of driving experience in our analysis.’’ For offenses the picture was somewhat different. While odds of a crash declined at an average of 9% per year over a 9-year period, there was an initial rise in risk of offense, with the second and third year of licensure showing an approximately 30% increase in odds over the initial year, after which risk steadily declined. Thus the risk of offense increases after the first year of licensure even as the risk of crash decreases. The first 4 years of this trend for offenses is similar to that found by Ferdun et al. (1967) for California drivers age 16–19.

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5.3. Gender differences Although the overall trends are similar for men and women, there were gender-specific differences. First, as has been frequently reported, compared to women, men have about twice the risk of committing an offense in any given year, with the difference greater for more serious offenses. Men also incur their first offense sooner than women. However, men showed a more rapid drop in their risk of offense, with the odds of an offense dropping about 9% per year for men compared to 7% for women. Men also have a higher risk of crash and incur their first crash sooner. Furthermore, their crashes are more likely to be at-fault, but the rate of decline in crash risk is comparable for both sexes. If the more rapid decrease in risk of offense observed for men were attributable to their accumulating higher mileage and thus more experience, it would be anticipated that a similar gender difference would be observed in the reduction of crash risk. This is not the case, and total crash risk decreases at comparable rates for both men and women. However, the odds that a crash will be at-fault decrease about twice as fast for women as for men.

5.4. Minor 6ersus serious incidents It was hypothesized that offenses that could be assumed to be less volitional should decrease more rapidly as a function of experience than more serious and more obvious offenses. The basis for this hypothesis was that experience should be more beneficial for improving ordinary driving but have less effect on deliberate risk taking. However, just the opposite is observed. It is the more blatant deliberate offenses that show the most rapid reductions, while the less serious offenses are more persistent over time. Although total offenses and less serious offenses increase after the first year of licensure, the proportion of offenses that may be considered serious begins to decrease immediately and continues to drop as duration of licensure increases (Fig. 2). The proportion of crashes that are at-fault also decreases over time, although not so rapidly as the proportion of offenses that are serious. Although offenses and crashes are positively correlated, it appears that different factors are operating in each. The early rise in both serious and less serious offenses observed during the second and third years of licensure is in contrast to the decline in both at-fault and not-at-fault crashes observed during these same years.

5.5. Earlier 6ersus later incidents As indicated above, serious offenses decline at a more rapid rate than do minor offenses. For drivers

with multiple offenses, the proportion of subsequent offenses that were serious declined. Even so, for those incurring four or more offenses, the proportion of their later offenses that were serious was still higher than that seen for drivers with only one or two offenses. Thus, the occurrence of a serious offense is more predictive of future offenses than the occurrence of a less serious offense. In contrast, previous crash experience is not predictive of whether a future crash will be at-fault. Whether a crash is the first or the fourth, the likelihood of its being at-fault remains the same. Rather, it is duration of license that is related to the likelihood of a crash being at-fault. This likelihood decreases as duration of licensure increases, but the decrease is not so great as that observed for the decrease in proportion of offenses that are serious.

5.6. Age, experience, and seriousness of incident It was hypothesized that more serious incidents would be under greater volitional control and hence less likely to be affected by experience. In contrast, it was hypothesized that less serious offenses and not-atfault crashes would be less likely to be under volitional control and that it was these incidents that would be more likely to be attributable to experience and thus decrease with increased driving. Neither of these hypotheses was confirmed. Quite the opposite, it is the more serious incidents that appear to decrease relative to the less serious with increased length of licensure. Thus, those incidents that appear to be more related to volitional control, i.e. serious offenses and at-fault crashes, are more strongly related to experience, as measured by length of licensure. It is important to note that only offenses resulting in conviction are included in these analyses. It is assumed that the offense behaviors involved occur with much greater frequency than the data would reflect, that is, many if not most offenses result in neither enforcement actions nor crashes. These analyses identify only those young drivers who did indeed suffer consequences for such risk-taking behavior. It might have been anticipated that future serious offenses and at-fault crashes would be more likely to involve other drivers who have not yet experienced consequences of any high-risk driving behaviors in which they may engage. This is not the case — those young drivers who are convicted of serious offenses do indeed show a decrease in the proportion of their future offenses that are serious, but their rate is still higher than that for drivers without previous offenses. In the case of crashes, there appears to be no ‘learning’ effect, in that future crashes are just as likely to be at-fault as are earlier crashes. Because serious infractions decrease more rapidly than less serious ones, it may be that young novice

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drivers truly do not recognize the risk involved in some of their driving behaviors, e.g. very high speeds, driving after drinking. If the risk is recognized, then the willingness to engage in this risk-taking behavior changes as a function of length of licensure. There is also a curious difference between the trajectory of total offenses versus proportion of total offenses that are serious. Total offenses actually increase after the first year of licensure, stabilize between the second and third year, and do not begin to decrease until about the fourth year of licensure. However, the proportion of offenses that are serious begins to decrease immediately, dropping on average about 8% per year of licensure. There was no difference between men and women in the rate of decrease. Driving experience, as indicated by length of licensure, affects serious offenses more than less serious offenses. In contrast, both at-fault and not-at-fault crashes decrease with increasing length of licensure, but the proportion of crashes that are at-fault decreases more than twice as fast for women as for men. Finally, increased age at time of licensure is associated with an increase in risk of an at-fault crash but about a 5% decrease in total crash risk for each succeeding year of age. However, total crash risk drops about 17% for each succeeding year of licensure, with a higher rate of decrease in the early years, suggesting that both age and experience are important, but that, among young drivers, experience contributes more to decreasing crash risk. Different results are found for offenses. Increasing the age at which licensure occurs is associated with a 9% increase in offense risk for each year older the subject is at time of licensing. Furthermore, the absolute risk of serious offense increases, but because total offense risk increases even more, the proportion of total offenses that are serious is lower for these drivers. When only serious offenses and at-fault crashes are considered, both show increases with increased age at time of initial licensure. Thus those behaviors that appear to be more characteristic of risk-taking are actually higher for these later licensees rather than lower, just the opposite of what would be expected if risk-taking declined primarily as a function of age. The findings related to age of initial licensure need to be interpreted with caution. The range of ages we are examining here is limited, with 90% of our subjects receiving their license between 15.8 and 18.6 years of age. Nevertheless, it may be anticipated that the difference in maturity between age 16 and 17 would be greater than the difference between age 18 and 19, so that age differences in these early years might be expected to be more strongly associated with detectable differences in risk-taking behaviors. However, in this society those who delay licensure may be different on other important dimensions than age, e.g. socioeconomic status, cultural norms, or individual abilities. If the effect of experience is attenuated among those who begin driving at an older

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age than is typical in the US, then over a broader range of ages, age may be more important than equivalent experience (Mayhew and Simpson, 1992, as reported in Peck, 1993). A major shortcoming of this study is the lack of driving exposure information. Ferdun et al. (1967) present results indicating that, based on self reports, young California drivers approximately double their annual mileage driven between ages 16 and 18, with further increases leveling off thereafter. If similar exposure patterns hold for our subjects — a likely supposition — controlling for exposure would show that most or all of the observed early increase in offense rates is attributable to increases in driving. Similarly, the early rapid drop in crash rates would be even more extreme if the presumed increase in driving exposure were to have been taken into account. The later more linear trends would presumably be unchanged if driving exposure then becomes constant. Also unchanged would be all measures of relati6e hazards and odds ratios of an offense being serious rather than minor, or a crash at-fault rather than not-at-fault, since these measures are independent of miles driven. Both of these latter measures show decreases beginning from the outset of licensure.

6. Conclusion Offense and crash patterns of young drivers vary as a function of gender, age at licensure, and length of licensure, treated here as a proxy for experience. Men have more offenses and crashes than women and incur their first offenses and crashes sooner than women. These differences are greater the more serious the incident, but the relative decline in risk of serious offense is about the same for both men and women. This is not true for crashes, where the proportion of crashes that are at-fault decreases about twice as fast for women as for men. Those who are older at time of initial licensing are at higher risk of offense during their first year of licensure, but their offenses are less likely to be serious. They are less likely to be in a crash, but, if involved, are more likely to be at fault. These paradoxical findings raise intriguing questions about the learning process in driving. There is only modest evidence of young driver ‘learning’ from specific incidents. Although persons who commit multiple offenses are less likely to have their later offenses be serious, the proportion of offenses that are serious remains higher than that of persons with only a single offense. This trend is not discernible among at-fault and not-at-fault crashes, suggesting that overall driving experience is more important for avoiding atfault crashes than the total number of crashes experienced.

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The initial hypothesis — that risk of offenses and crashes that appear to be more under volitional control and hence more likely to be deliberate risk taking, would be more stable through time, while apparently less volitional offenses and crashes would be more related to experience and would decline-is not borne out by the data. In contrast, incidents that may be considered more related to volitional control appear to decline even more rapidly than those that may be considered to be less volitional and hence more affected by driving experience. This relationship is supported by the fact that drivers who obtain initial licensure at a slightly later age have higher proportions of at-fault crashes (although total crashes decrease). However, it was also found that the proportion of crashes that are at-fault decreases more than twice as fast for women as for men. Women still drive fewer miles than men (Our Nation’s Travel, 2000). If driving experience has a greater impact on volitional behaviors, it would be anticipated that men, who drive more, would show a more rapid decrease. An alternative possibility is that men and women differ in respect to the experience of an at-fault crash, with women ‘learning’ more from the experience. Overall, it appears that driving experience, as measured by length of licensure, has a greater impact on more hazardous offenses and at-fault crashes than on less serious incidents. The data do hint, however, that the drop in risk-taking behavior relative to inexperience may be stabilizing after eight or nine years of driving experience. For the majority of drivers, who obtain license at age 16, this time period coincides with the approximate length of time required for the crash risk to level out. Previous studies suggest that after around age 24 or 25, crash risk generally remains low until after around age 55 (Cerrelli, 1995; Williams and Carsten, 1989). Further observation and analysis will be required to determine if this is indeed the case for these subjects.

Acknowledgements This study was supported by NIAAA Grant AA 09026.

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