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Graduated driver licensing and motor vehicle
crashes involving teenage drivers: an exploratory
age-stratified meta-analysis
Motao Zhu,
1
Peter Cummings,
2
Haitao Chu,
3
Jeffrey H Coben,
4
Guohua Li
5
1
Department of Epidemiology
and Injury Control Research
Center, West Virginia University,
Morgantown, West Virginia,
USA
2
School of Public Health and
Harborview Injury Prevention &
Research Center, University of
Washington, Seattle,
Washington, USA
3
Division of Biostatistics,
School of Public Health,
University of Minnesota,
Minneapolis, Minnesota, USA
4
Department of Emergency
Medicine and Injury Control
Research Center, West Virginia
University, Morgantown,
West Virginia, USA
5
Departments of
Anesthesiology and
Epidemiology, Columbia
University, New York,
New York, USA
Correspondence to
Dr Motao Zhu, Department of
Epidemiology and Injury Control
Research Center, West Virginia
University, PO Box 9151,
Morgantown, WV 26506-9151,
USA; mozhu@hsc.wvu.edu
Accepted 12 October 2012
ABSTRACT
Objective Graduated Driver Licensing (GDL) has been
implemented in Australia, Canada, New Zealand, USA
and Israel. We conducted an exploratory summary of
available data to estimate whether GDL effects varied
with age.
Methods We searched MEDLINE and other sources
from 1991–2011. GDL evaluation studies with crashes
resulting in injuries or deaths were eligible. They had to
provide age-specific incidence rate ratios with CI or
information for calculating these quantities. We included
studies from individual states or provinces, but excluded
national studies. We examined rates based on person-
years, not license-years.
Results Of 1397 papers, 144 were screened by
abstract and 47 were reviewed. Twelve studies from 11
US states and one Canadian province were selected for
meta-analysis for age 16, eight were selected for age 17,
and four for age 18. Adjusted rate ratios were pooled
using random effects models. The pooled adjusted rate
ratios for the association of GDL presence with crash
rates was 0.78 (95% CI 0.72 to 0.84) for age 16 years,
0.94 (95% CI 0.93 to 0.96) for 17 and 1.00 (95% CI 0.95
to 1.04) for 18. The difference between these three rate
ratios was statistically significant: p<0.001.
Conclusions GDL policies were associated with a 22%
reduction in crash rates among 16-year-old drivers, but
only a 6% reduction for 17-year-old drivers. GDL showed
no association with crashes among 18-year-old drivers.
Because we had few studies to summarise, particularly
for older adolescents, our findings should be considered
exploratory.
INTRODUCTION
Motor vehicle collisions are a major source of mor-
bidity and mortality around the world, causing
about 20–50 million injuries and 1.2 million deaths
every year.
1
Motor vehicle crashes are the leading
cause of death among people aged 15–29 years
worldwide.
12
Young novice drivers have the
highest crash rate; per kilometre driven, the crash
rate for 16-year-old drivers is approximately four
times greater than that for drivers ages 30–59 years
in the USA.
3
This excess crash risk is mainly due
to inexperience and risky driving behaviours.
4–6
To
address this issue, some European countries includ-
ing Sweden, Norway, France and Belgium have
implemented an extended learner permit phase,
requiring supervised driving under all conditions
for adolescent drivers before they reach 18 years.
7–9
Australia, Canada, New Zealand, the USA and Israel
have implemented Graduated Driver Licensing (GDL)
laws in some or all states/provinces.
10–15
GDLs in
Australia and Canada apply to drivers of all ages. In
the USA, New Jersey’s GDL system applies to people
entering the licensing process who are under 21, and
some other states apply some restrictions to people
18 and older. In the USA, GDL regulates licensing
and driving behaviours among adolescents younger
than 18 years in three phases: the extended learner
phase, requiring supervised driving under any condi-
tions for 3–12 months; the intermediate phase, allow-
ing unsupervised driving under low-risk conditions
such as daylight or when carrying less than one
young passenger; and the full licensure phase,permit-
ting unsupervised driving all the time.
GDL policies could reduce crash rates if they
reduced risky driving behaviours among those
covered by GDL restrictions. Or they might reduce
crash rates by reducing the amount of driving by
adolescents. Either of these mechanisms would
reduce crash rates for drivers age 16 and 17 years,
while rates for those 18 years would be unchanged.
One effect of GDL programmes is to delay licen-
sure by increasing the minimum permit and/or
intermediate license age, or by placing additional
requirements on teenagers such as minimum prac-
tice driving hours. Another possibility is that GDL
laws may reduce crash rates among those age 16
and 17 years, but crash rates might increase among
drivers age 18 years because they drive fewer miles
and therefore learn fewer driving skills while
driving under GDL laws at ages 16 and 17. In
states that do not apply GDL to people 18 and
older, teenagers might delay licensure until 18 and
thereby have higher crash rates at age 18 because
they are new drivers. Another possibility is that
crash rates are reduced among drivers age 16 and
17 years, and also among drivers age 18 years
because teenagers licensed under GDL would have
a greater amount of practice and gradual introduc-
tion to riskier driving situations and thereby lower
crash rates at age 18.
GDL implementation has been reported to
reduce involvement in a vehicle crash as a driver
by approximately 15–40% for adolescents aged
16 years.
16–20
One reason that studies have found
different effects of GDL is because some GDLs
have more restrictive provisions than others. In
addition, the definition used for a GDL law has
varied among studies.
16 17 19 20
Few studies have
examined 17-year-old and 18-year-old drivers. The
potential effect of GDL on teenagers age 18 years
is primarily a US issue, as other countries have
older licensing ages and most apply GDL to drivers
of all ages. A study of US fatal crashes reported
similar crash rates among teenagers age 18 years
Injury Prevention 2012;00:1–9. doi:10.1136/injuryprev-2012-040474 1
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before and after GDL implementation,
17
but another study of
fatal US crashes found that GDL was associated with a 10%
increase in fatal crash involvements among those age 18.
16
We
conducted a systematic review and meta-analysis to estimate
age-specific associations between GDL implementation and
crash rates.
METHODS
We searched MEDLINE, Transportation Research Information
Service, Web of Science, Google Scholar, and internet sites
maintained by the Insurance Institute for Highway Safety, the
National Highway Traffic Safety Administration, the Centres
for Disease Control and Prevention, and the American
Automobile Association Foundation for Traffic Safety for
studies of GDL polices from January 1991 through December
2011. Our search terms included: (1) GDL; (2) (graduate* or
gradual* or delay* or driver or provisional) in combination
with (permit* or licen* or restrict* or delay*); (3) (teen* or
you* or adolescen*) in combination with (driv*). We used the
‘related citations’feature to capture additional references for
selected articles. We examined the references of articles and
reviews.
18 21 22
Full-text versions of articles or reports were reviewed by the
first author (MZ). Data extraction was conducted by the first
author (MZ) and verified by the second author (PC). For this
study we defined GDL as a new law with a learner phase of at
least 3-months plus an intermediate phase that restricts driving
at night and/or restricts the number of passengers allowed.
23
To be included in the meta-analysis, a study had to use counts
of crashes, injuries or deaths as the outcome. It had to provide
age-specific incidence rate ratio estimates with CI or informa-
tion that allowed us to calculate these quantities. We included
studies from individual states or provinces. We excluded
national studies because within a given nation they have
overlapping time periods and therefore their results are not
independent. In addition, national studies overlap the time
intervals of most studies from smaller jurisdictions such as
states. We examined rates based on person-years, and did
not consider rates based on license-years because license data
for adolescents is not often or consistently reported by states/
provinces. Some jurisdictions include adolescents in the inter-
mediate phase, while others count only those fully licensed.
Not all studies used the same outcomes. When extracting
rate or rate ratio estimates, we first used an estimate based on
the count of crashes in which a teenager was involved as a
driver and at least one person was injured. Our second choice
was a count of injured teenage drivers. Third was a count of
crashes with a teenage driver and at least one death. Fourth
was a count of fatally injured teenage drivers. We extracted the
year of GDL implementation and information about each law.
We identified whether the original manuscript adjusted the rate
ratio for temporal trends using rate data from drivers age 21
and older, who should not be affected by GDL laws but should
be affected by other factors that influence crash rates, such as
changes in speed laws, seat belt use or vehicle design.
Our goal was to pool age-specific rate ratios from each study
to summarise the association between GDL presence and crash
rates. One study (North Carolina)
15
provided an age-specific
rate ratio that was not adjusted for other changes in rates over
time; an adjusted estimate was not available. The needed rate
ratios with CIs were not in the remaining studies. We therefore
extracted from each study age-specific counts of outcome
events (crashes with an injury or counts of injured drivers)
and population estimates before and after GDL passage. These
data came from 1–5.5 years of time before GDL passage and
1–6.5 years after passage. We used age categories of 16, 17 and
18 years as well as a category for older drivers if that informa-
tion was available. We estimated adjusted rate ratios (aRR) for
the association of GDL laws with crash rates using Poisson
regression, with age-specific person-time as offsets. We adjusted
for age group and included interaction terms between
GDL presence and adolescent age groups. Except for NC, we
adjusted for time as a linear term in all models to account for
changes in crash rates over time due to factors other than GDL
laws; but the method used depended upon the available data.
For five studies (California,
24
Florida,
25
Georgia,
26
Nova
Scotia,
11
New York
27
) the adjustment for time was based on
changes in the crash rates of older drivers, because we had only
data from one time period before the GDL law and one after.
For four studies (Maryland,
28
Michigan,
29
Pennsylvania,
30
Texas
31
) the adjustment for time was based on data for teenage
drivers, because data was available for three or more time inter-
vals. For two studies (Ohio,
32
Wisconsin
33
) the adjustment
used data from both teenage and older drivers, because data
was available for three or more time intervals and data was
available for older drivers.
To clarify we will describe what we did for one study from
Nova Scotia.
11
For our meta-analysis we needed to extract an
adjusted rate ratio with a CI for the association between the
GDL law and the crash rate. But this information was not in
the paper. Table 3 of that paper provided data for counts of
Nova Scotia drivers in crashes with an injury for the years 1993
and 1995 (table 1). Nova Scotia’s new GDL began in October
of 1994. We created variables for age group and the interaction
between age 17 and the GDL law. Then we used Poisson regres-
sion to estimate the aRR. This was adjusted for the linear
trend in crash rates for persons age 25 years and older from
1993 to 1995.
We used inverse-variance methods to produce pooled esti-
mates of the aRRs from each study using Stata V.12.
Random-effect estimates were calculated using the method of
DerSimonian and Laird.
34
Fixed effect estimates were also
obtained. The Cochran Q-statistic was used to test the hypoth-
esis that rate ratios were homogeneous across studies.
35
We cal-
culated I
2
, an estimate of the percent of total variation
between studies due to heterogeneity rather than chance.
36
To
test for publication bias, funnel plots were inspected and
Egger’s test for asymmetry was used.
37
To identify characteris-
tics associated with heterogeneity, we used subgroup analyses
for the following variables (1) outcome type (crash with injury,
crash with death), (2) entry age for learner permit phase
(<16 years, 16), (3) entry age for intermediate phase (16 years,
Table 1 Data from one study of graduated driver licensing in Nova
Scotia
Graduated
driver licensing
status Year
Age
(years)
Drivers in
crash with
injury Person-years
Crash rate
per 10000
person-years
Absent 1993 16 167 12700 131.50
Absent 1993 17 201 13400 150.00
Absent 1993 25 or
older
4233 661700 63.97
Present 1995 16 110 12620 87.16
Present 1995 17 193 12501 154.39
Absent 1995 25 or
older
4317 672986 64.15
GDL, Graduated Driver Licensing.
2Injury Prevention 2012;00:1–9. doi:10.1136/injuryprev-2012-040474
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>16), (4) night-time driving restriction (starting at 21 : 00 or
22 : 00; starting at 23 : 00, midnight or 1 : 00; none or no
change) and (5) number of young passengers allowed (0, 1, 2, 3
or more). Because tests of heterogeneity are statistically weak
when there are few studies, we used p<0.1 as our criteria to
judge that there was heterogeneity.
When a GDL is passed, its effects may not be immediate for
several reasons. One is that some people might already have a
license and may be grandfathered in under the law. For
example, a 16-year-old may be allowed to continue to drive
under the old law provisions after the new law is passed. For
this reason, some studies use data until the time of GDL imple-
mentation. Then after the GDL goes into effect, they omit
1 year of data, using information for 16 year olds only after the
law has been effect for over a year. We attempted to deal with
this issue in our analyses. For example, for age 16 years there
were 12 studies. Of these, six omitted the first year after the
law went into effect or presented data in such a way that we
could omit that first year. For a further two studies, the first
year after the law went into effect could not be omitted, but
there were 5 years of data available after the law went into
effect, and thus 16-year-olds were fully covered by the law in
four of the 5 years of post-law data. This means that any dilu-
tion of the GDL effect induced by including the first year after
implementation should be relatively small in 5 years of data.
But in four jurisdictions, Florida, Nova Scotia, Ohio, and
Pennsylvania, we could not exclude a full year of data after the
law went into effect and we did not have 5 years of data after
the law went into effect. For Florida, a 6 month period after
the law went into effect was excluded. For Nova Scotia,
2 months were excluded. For Ohio, 6 months were excluded.
No post-law time was excluded for Pennsylvania. Thus, in
theory, some dilution of the GDL effect might be found in
these four studies. To find out if there was evidence for this
dilution of effect, we used a test of heterogeneity to compare
the pooled aRR in the eight studies where one post-law year
was excluded, or 5 years of post-law data was used, with the
pooled aRR in the four studies where this could not be done.
For drivers age 17 years, we followed the same procedure,
except we tried to exclude 2 years of data after GDL implemen-
tation. There were eight studies for this age group. For three,
we could either exclude 2 years of data after GDL passage or
use 5 years of data after passage. We could not do this for the
other five studies. Again, we compared the aRR estimates from
the first three studies with the other five.
For drivers age 18 years there were only four studies.
California excluded the third year and used the fourth year to
the eighth year after GDL implementation. Florida excluded
only 6 months. Georgia excluded no time, but used 5.5 years of
data after implementation. Wisconsin excluded 2.25 years of
post-law data. We did not perform the test of heterogeneity in
this small group of four studies.
RESULTS
The literature search identified 1397 papers, but 1253 were
excluded as not relevant based upon their titles (figure 1).
Another 97 papers were removed after reading the abstract and
the remaining 47 papers were read. We identified 12 studies that
could provide crash counts and population estimates for age
16,
11 15 24–33
eight for age 17,
11 24–27 30 32 33
and four for age
18.
24–26 33
We conducted new analyses to estimate adjusted rate
ratios with CIs for GDL presence for 11 of the 12 studies for
adolescents age 16 years. For North Carolina
15
the data were not
available for a new analysis. We conducted new analyses for all
eight studies for age 17, and all four studies for age 18.
Adolescents age 16 years
Data were used from 11 US states (California,
24
Florida,
25
Georgia,
26
Maryland,
28
Michigan,
29
New York,
27
North
Carolina,
15
Ohio,
32
Pennsylvania,
30
Texas,
31
Wisconsin
33
) and
one Canadian province (Nova Scotia
11
) (tables 2 and 3). Nine
studies
11 15 24–29 33
were published in peer-reviewed journals and
three
30–32
were reports. In these jurisdictions the earliest GDL
law was implemented in July 1996 in Florida and the latest in
September 2003 in New York. The entry age for a learner permit
stayed the same in 10 jurisdictions, was reduced from 15 to
14 years and 9 months in Michigan, and was reduced from
16 years to 15 years and 6 months in Ohio. The length of the
learner period was extended to 6 months or less in ten jurisdic-
tions, and 12 months in two jurisdictions. The entry age for the
new intermediate phase after GDL implementation was 16 years
in eight jurisdictions, older than 16 and less than 16.5 in two,
and 16.5 in two. North Carolina did not allow beginners to drive
unsupervised after 21 : 00, and night-time restriction was 23 : 00
or later in other jurisdictions. Maryland and New York imple-
mented a night-time restriction before their GDL laws. Three
jurisdictions mandated no more than one young passenger,
New York allowed two and Georgia allowed three; the rest did
not have young passenger restrictions. Of 12 GDL laws first
implemented between 1996 and 2003 in our meta-analysis,
3 (25%) restricted both night driving and the number of passen-
gers, compared with 9/37 (24%) of all US GDL laws during the
same period.
23
The outcome was injury crashes for nine
studies,
11 15 25 27–30 32 33
and fatal crashes for three.
24 26 31
Figure 1 Flow chart of study
selection for data extraction.
Injury Prevention 2012;00:1–9. doi:10.1136/injuryprev-2012-040474 3
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The aRR associated with GDL implementation ranged from
0.64 in Georgia to 0.89 in Florida. The crash rate was less in the
presence of GDL compared with what would have been
expected without GDL: pooled random-effects aRR 0.78 (95%
CI 0.72 to 0.84) (figure 2). The Q-statistic indicated that the
individual aRRs did not estimate the same effect ( p<0.001)
and I
2
was 91%, indicating that most of the difference between
estimates could not be ascribed to chance variation. The pooled
fixed-effects aRR was 0.84 (95% CI 0.82 to 0.85) We found no
evidence of publication bias: p=0.17.
In only one subgroup analysis did we find that pooled
random-effect aRR estimates showed statistically significant vari-
ation (table 4). The pooled aRR was 0.83 using the nine jurisdic-
tions where adolescents could obtain a learner permit before
reaching age 16, and 0.68 in the three jurisdictions where adoles-
cents had to wait until after reaching 16; p<0.001 for a test that
these two estimates differed. Subgroup aRR estimates did not
differ significantly according to outcome type, entry age for the
intermediate phase, category of night-time driving restrictions, or
the allowed number of young passengers.
The pooled aRR was 0.79 among the four jurisdictions
(Florida, Nova Scotia, Ohio and Pennsylvania) where less than
one full year of data right after GDL implementation could be
excluded from analysis and 0.78 using the eight remaining juris-
dictions where at least 1 year of post-GDL data could be
excluded from analysis or there were at least 5 years of data
after GDL passage; p=0.91 for a test that these two estimates
differed.
Adolescents age 17 years
Useable estimates came from seven US states (California,
24
Florida,
25
Georgia,
26
New York,
27
Ohio,
32
Pennsylvania,
30
Wisconsin
33
) and one Canadian province (Nova Scotia
11
)
(tables 2 and 3). Six
11 24–27 33
were peer-reviewed articles
and two
30 32
were reports. All also provided estimates for
16-year-olds, so characteristics of their GDL laws have already
been summarised.
The aRR associated with GDL implementation ranged from
0.81 in Georgia to 1.03 in Nova Scotia. The pooled random-
effects aRR was 0.94 (95% CI 0.93 to 0.96) (figure 3). The
p value for the Q-statistic was 0.44 and I
2
was less than 1%,
indicating homogeneity among jurisdictional aRR estimates.
The pooled fixed-effects aRR was 0.94 (95% CI 0.93 to 0.96).
We found no evidence of publication bias: p=0.85. Subgroup
analyses were not conducted for this age group, because there
was no evidence of heterogeneity between jurisdictional esti-
mates and the pooled association between GDL laws and the
aRR was close to 1.0.
The pooled aRR was 0.92 using the three jurisdictions
(California, Georgia and Wisconsin) where we could exclude
two full years of data after GDL passage or there were at least
5 years of data after GDL implementation and 0.94 among the
remaining five jurisdictions where less than two full years
of data after GDL passage could be excluded from analysis;
p=0.56 for a test that these two estimates differed.
Adolescents age 18 years
Data were available from four US states (California,
24
Florida,
25
Georgia,
26
Wisconsin
33
) (tables 2 and 3). The traffic crash rate
ratio associated with GDL implementation ranged from 0.97 in
Georgia and Wisconsin to 1.17 in California. The pooled
random-effects aRR was 1.00 (95% CI 0.95 to 1.04) (figure 4).
The p value for the Q-statistic was 0.18 and I
2
was 39%, indi-
cating low heterogeneity among aRRs. The pooled fixed-effects
Table 2 Characteristics of the 12 studies which provided useable age-specific data for meta-analysis
Jurisdiction
First author,
year
Graduated
driver licensing
status
Age for
learner permit
(years,
month)
Length of
learner period
(months)
Practice
hours
Night-time
driving
restrictions
No. of young
passengers
allowed
Age for
intermediate phase
(years, month)
Age for full
licensure phase
(years, month)
California Males, 2007 Absent 15 1 None* None* None* NA 16
Present 15 6 50 Midnight 0 16 17
Florida Ulmer, 2000 Absent 15 None* None* None* None* NA 16
Present 15 6 None* 23 : 00 None* 16 18
Georgia Rios, 2006 Absent 15 None* None* None* None* NA 16
Present 15 12 None* 1 : 00 3 16 18
Maryland Kirley, 2008 Absent 15, 9 month 0.5 None* Midnight None* 16 17, 9 month
Present 15, 9 month 4 40 Midnight None* 16, 1 month 17, 7 month
Michigan Shope, 2004 Absent 15 1 None* None* None* NA 16
Present 14, 9 month 6 50 Midnight None* 16 17
New York Zhu, 2010 Absent 16 None* None* 21 : 00 None* 16 17
Present 16 Up to 6 20 21 : 00 2 16, up to 6 month 17
North Carolina Foss, 2001 Absent 15 None* None* None* None* NA 16
Present 15 12 None* 21 : 00 None* 16 16, 6 month
Nova Scotia Mayhew, 2001 Absent 16 2 None* None* None* NA 16, 2 month
Present 16 6 None* 0 Midnight None* 16, 6 month 18, 6 month
Ohio Ohio DPS, 2001 Absent 16 None* None* None* None* NA 16
Present 15, 6 month 6 50 1 : 00 None* 16 17
Pennsylvania Coben, 2003 Absent 16 None* None* 23 : 00 None* 16 17
Present 16 6 50 23 : 00 None* 16, 6 month 17
Texas Willis, 2006 Absent 15 None* None* None* None* NA 16
Present 15 6 None* Midnight 1 16 16, 6 month
Wisconsin Fohr, 2005 Absent 15, 6 month None* None* None* None* NA 16
Present 15, 6 month 6 30 Midnight 1 16 16, 9 month
*No restrictions.
GDL, Graduated Driver Licensing; NA: not applicable.
4Injury Prevention 2012;00:1–9. doi:10.1136/injuryprev-2012-040474
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Table 3 Crude and adjusted age-specific rate-ratios for each study
Jurisdiction Outcome used*
Age
(years)
Count of
outcomes†
Time period
before GDL‡
GDL onset
(month/year)
Time period
after GDL‡
Crude rate
before GDL§
Crude rate
after GDL§
Crude rate
ratio
Adjusted rate
ratio (95% CI)
Adjustment
method¶
New analysis
done**
California Driver deaths 16 208 1/95–6/98 7/98 7/99–12/05 4.6 3.9 0.86 0.86 (0.65 to 1.14) Older drivers Yes
17 348 1/95–6/99 7/00–12/05 7.0 7.1 1.02 0.98 (0.79 to 1.21)
18 610 1/95–6/00 7/01–12/05 11.1 14.0 1.26 1.17 (1.00 to 1.38)
Florida Crash with
injury/death
16 10958 1/95–12/95 7/96 1/97–12/97 3228 2902 0.90 0.89 (0.86 to 0.92) Older drivers Yes
17 14323 4299 4056 0.94 0.93 (0.90 to 0.97)
18 16302 4928 4976 1.01 1.00 (0.97 to 1.03)
Georgia Crash with
death
16 547 1/92–6/97 7/97 7/97–12/02 57.0 36.1 0.63 0.64 (0.53 to 0.76) Older drivers Yes
17 590 54.8 44.4 0.81 0.81 (0.69 to 0.96)
18 724 62.6 60.4 0.97 0.97 (0.84 to 1.13)
Maryland Injured drivers 16 920 1/96–12/98 7/99 1/01–12/03 294 142 0.48 0.78 (0.52 to 1.19) Teenage drivers Yes
Michigan Crash with
injury/death
16 23824 1/96–12/96 4/97 1/98–12/01 4517 3038 0.67 0.80 (0.76 to 0.85) Teenage drivers Yes
New York Injured drivers 16 162 1/01–12/01 9/03 1/05–12/05 81.4 55.3 0.68 0.69 (0.49 to 0.96) Older drivers Yes
17 313 134.0 131.2 0.98 0.99 (0.77 to 1.28)
North
Carolina
Crash with
injury
16 NA 1/96–12/96 12/97 1/99–12/99 370 270 0.72 NA None No
Nova Scotia Crash with
injury/death
16 277 1/93–12/93 10/94 1/95–12/95 1315 872 0.66 0.66 (0.52 to 0.84) Older drivers Yes
17 394 1500 1544 1.03 1.03 (0.84 to 1.26)
Ohio Crash with
injury
16 57149 1/90–12/97 7/98 1/99–12/99 4110 3622 0.88 0.88 (0.86 to 0.91) Older and teenage
drivers
Yes
17 67832 4757 4441 0.93 0.94 (0.91 to 0.96)
Pennsylvania Crash with
injury
16 17384 1/96–12/99 1/00 1/00–12/00 2238 1465 0.65 0.68 (0.64 to 0.71) Teenage drivers Yes
17 25090 3065 2881 0.94 0.97 (0.93 to 1.01)
Texas Crash with
death
16 360 1/00–12/01 1/02 1/03–12/04 31.2 23.2 0.74 0.77 (0.40 to 1.47) Teenage drivers Yes
Wisconsin Crash with
injury
16 8336 1/99–12/99 9/00 1/02–12/03 4014 3073 0.77 0.85 (0.81 to 0.89) Older and teenage
drivers
Yes
17 6148 1/99–12/99 1/03–12/03 4061 3431 0.84 0.94 (0.89 to 0.99)
18 6180 1/99–12/99 1/03–12/03 3994 3473 0.87 0.97 (0.92 to 1.02)
*We used age-specific counts of drivers in this order of preference: (1) crash with at least one injury, (2) injured drivers, (3) crash with at least one death, (4) drivers who died.
†Overall count of outcome events.
‡Rate information came from these time periods before and after GDL onset.
§Rate per 100 000 person-years based on population estimates.
¶Except for North Carolina, rate ratios were adjusted for change in crash rates over time using data from one of three groups: (1) older drivers; (2) teenage drivers; (3) older and teenage drivers.
**Yes means that we extracted data to re-estimate an adjusted rate ratio or CI that was not in the original report.
GDL, Graduated Driver Licensing.
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aRR was 1.00 (95% CI 0.97 to 1.02). There was no evidence
of publication bias: p=0.41. Subgroup analyses were not con-
ducted for this age group, as there were only four studies, and
the pooled association between GDL laws and crash rates was
close to one.
Variation of age-specific pooled estimates
The random-effects pooled aRR associated with GDL laws was
0.78 for those age 16 years, 0.94 for those age 17 and 1.00 for
those age 18; a test that these estimates differed was statistic-
ally significant, p<0.001. We also tested whether the aRR for
16-year-olds was different from that for 17-year-olds among
the eight studies with estimates for both these age groups. The
pooled random-effects aRR was 0.78 (95% CI 0.71 to 0.86) for
age 16 and 0.94 (95% CI 0.93 to 0.96) for age 17; p value for a
test of difference was <0.001. Among the four studies with
results for all age groups, the aRR was 0.82 (95% CI 0.75 to
0.90) for age 16 years, 0.93 (95% CI 0.91 to 0.96) for age 17
and 1.00 (95% CI 0.95 to 1.04) for age 18; these estimates were
statistically different, p<0.001.
DISCUSSION
Our meta-analysis estimated that adolescents aged 16 years
experienced a 22% (95% CI 16% to 28%) reduction in crash
rates, while among 17-year-olds the rate reduction was smaller,
6% ( 95% CI 4% to 7%), and among teenagers aged 18 years
Figure 2 Adjusted rate ratios for traffic crashes comparing graduated driver licensing (GDL) presence with absence for adolescents age 16 years.
Table 4 Subgroup estimates of the random-effect pooled adjusted rate ratios for crash rates under GDL laws compared with expected rates without
GDL laws, for adolescents age 16 years
Subgroup Category
Random-effect pooled
adjusted rate ratio 95% CI
p Value for a test that the adjusted
rate ratios are the same*
Outcome type Crash with injury 0.79 0.73 to 0.85 0.44
Crash with death 0.72 0.58 to 0.90
Entry age for learner permit phase <16 years 0.83 0.79 to 0.87 <0.001
16 years 0.68 0.64 to 0.71
Entry age for intermediate phase 16 years 0.81 0.75 to 0.86 0.41
>16 years 0.74 0.61 to 0.90
Night-time driving restriction Starting at 21 : 00 or 22 : 00 0.72 0.62 to 0.84 0.54
Starting after 22 : 00 0.79 0.73 to 0.86
None or no change 0.72 0.56 to 0.94
Number of young passengers allowed 0 0.86 0.65 to 1.14 0.19
1 0.85 0.81 to 0.89
2 0.69 0.49 to 0.96
3 or more 0.76 0.69 to 0.84
*p Value for a test of homogeneity using the inverse variance method. A small p value is evidence that the adjusted rate ratios vary more than expected if the GDL laws had similar effects
in similar populations.
GDL, Graduated Driver Licensing.
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the rate changed little after GDL implementation. The appar-
ent difference in GDL effects among those age 16 and those
age 17 years is probably not due to the different study jurisdic-
tions, as the pooled random-effects aRR for 16-year-olds was
0.78 using the eight studies with results for both ages 16 and
17, the same as the pooled aRR of 0.78 using all 12 studies.
Perhaps the most important limitation of this meta-analysis
was the small number of studies available with age-specific
Figure 3 Adjusted rate ratios for traffic crashes comparing graduated driver licensing (GDL) presence with absence for adolescents age 17 years
Figure 4 Adjusted rate ratios for traffic crashes comparing graduated driver licensing (GDL) presence with absence for adolescents age 18 years.
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data for ages 17 and 18 years; there were reports available
from 12 jurisdictions for age 16 years, but only from eight jur-
isdictions for age 17 and four for age 18. Because we had few
studies to summarise, all results are exploratory and those for
adolescents age 18 years are particularly unreliable due to the
small number of studies. Second, it would be ideal to study
the same outcomes for every jurisdiction as it is possible that
associations with GDLs might vary with the type of crash
outcome. Among the 12 studies for age 16 years, the rate ratio
for GDL laws varied somewhat according to whether the
outcome involved injury (aRR 0.79) or death (aRR 0.72), but
the difference in these rate ratios was modest and not statis-
tically different ( p=0.44, table 4). Another limitation is that
we could not exclude 1 to 3 years of data after GDL imple-
mentation for all studies, allowing us to examine whether
GDL effects might be stronger when all drivers of a given age
are covered by the laws. To address this problem, we compared
aRRs based on excluding 1 or 2 years of data after GDL
passage or based on at least 5 years of data after GDL passage,
with estimates from jurisdictions where the years immediately
after GDL passage had to be included. These estimates were
similar. A national evaluation of 1996–2007 US fatal crashes
compared the immediate effect of GDL with 1-year delay for
age 17, and 2-year delay for age 18, and reported similar esti-
mates.
17
Another limitation is that our estimates may be
subject to residual confounding. GDL effectiveness should
ideally be estimated with many repeated measures of crash
rates before and after GDL implementation. We conducted
new analyses for 11 of the 12 studies for age 16 by using
methods to control for temporal changes in crash rates unre-
lated to GDL policies. However, some traffic safety factors
may affect adolescents and adults differently, and our ability
to control for temporal trends was limited by the available
data years in the original research. Nevertheless, our estimates
were comparable to nationwide US evaluations using more
than 10 years of crash rates.
16 17 19 20
Another limitation is
that none of the studies selected for meta-analysis contained
any information about the amount of driving done by
adolescents.
Our estimate of a 22% (95% CI 16% to 28%) reduction of
traffic crash rates among 16 year olds is consistent with previ-
ous studies for this age group for the entire USA. A review of
collision claims for vehicles less than 4-years-old during 1996
through 2006 reported a 22% reduction (95% CI 18% to 27%)
for states with good GDL ratings from the Insurance Institute
for Highway Safety and a 17% reduction (95% CI 12% to
31%) for states with fair GDL ratings.
20
In an analysis of
crashes with a death during 1994–2004, GDL polices that
included five or more of seven GDL components were asso-
ciated with an 18–21% reduction in crash rates.
19
Another
study of all fatal crashes during 1996–2007 reported a 41%
reduction in crash rates for states with a good GDL rating and
an 18% reduction for states with a fair GDL rating.
17
Another
analysis of 1986–2007 fatal crashes in the USA reported that
among those age 16 years, GDL was associated with a 16%
reduction (95% CI 6% to 25%) in crash rates for jurisdictions
with either a night-time driving restriction starting before 1: 00
or a passenger restriction allowing no more than one young
passenger, and a 26% reduction (95% CI 16% to 35%) for juris-
dictions with both restrictions.
16
In a Cochrane review, five
GDL programmes were summarised with a median decrease of
15.5% (range 8% to 27%) for the adjusted rate of all crashes
during the first year and 21% (range 2% to 46%) for the
adjusted rate of crashes with an injury among 16 year olds.
18
Our estimate of association (aRR 0.94, 95% CI 0.93 to 0.96)
between GDL and traffic crash rates among 17 year olds is con-
sistent with previous studies for this age group for the entire
USA. A study of all fatal crashes during 1996–2007 reported a
rate ratio of 0.81 for states with a good GDL rating and 0.97
for states with a fair GDL rating.
17
Another study of fatal
crashes during 1986–2007 reported an aRR of 0.98 (95% CI
0.92 to 1.04) for GDL with night-time driving or passenger
restriction, and an aRR of 0.91 (95% CI 0.83 to 1.01) for GDL
with both restrictions.
16
Our study did not find that GDL implementation was related
to traffic crash rates among 18 year olds (aRR 1.00, 95% CI 0.95 to
1.04). A study of all fatal crashes for 18-year-olds in the USA
during 1996–2007 reported a rate ratio of 0.96 for states with a
good GDL rating and 1.03 for states with a fair GDL rating.
17
However, another study of all fatal crashes in the US during 1986–
2007 reported that GDL with either a night-time driving restric-
tion or passenger restriction was associated with a 10% increase
(aRR 1.10, 95% CI 1.03 to 1.18) in fatal crash involvements
among those age 18.
16
Further research is needed to evaluate
whether GDL negatively affects 18 year olds.
CONCLUSION
GDL implementation was associated with a 22% reduction in
traffic crash rates among 16 year olds, but only a 6% reduction
in rates among 17 year olds. GDL implementation was unre-
lated to crash rates among 18 year olds, but this exploratory
finding was based upon a sample of only four jurisdictions and
should be treated with caution.
Acknowledgements We express appreciation to Michele Fields and Laurel Sims
at the Insurance Institute for Highway Safety for their assistance in describing
graduated driver licensing laws; and to anonymous reviewers for their constructive
comments.
Contributors MZ originated and designed the study, conducted data analysis, and
led the writing. PC verified data extraction, provided expertise in meta-analysis,
offered suggestions for data analysis, and critically reviewed the manuscript. HC,
JHC and GL critically reviewed and substantially revised the manuscript.
Funding MZ and JHC received support from grants (R21CE0018205, R49CE001170)
from the US Centers for Disease and Prevention, National Center for Injury Prevention
and Control (http://www.cdc.gov/injury/). The funders had no role in study design,
data collection and analysis, decision to publish or preparation of the manuscript.
What is already known on the subject
▸Graduated Driver Licensing has been reported to reduce
involvement in a vehicle crash as a driver by approximately
15–40% for adolescents aged 16 years.
▸Few studies have examined 17 and 18 year old drivers.
What this study adds
▸Adolescents age 17 years received less benefit from GDL
laws than adolescents age 16 years.
▸GDL implementation showed little association with crashes
among 18 year olds, but this exploratory finding was based
upon a sample of only four jurisdictions and should be
treated with caution.
8Injury Prevention 2012;00:1–9. doi:10.1136/injuryprev-2012-040474
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Competing interests None.
Provenance and peer review Not commissioned; externally peer reviewed.
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doi: 10.1136/injuryprev-2012-040474
published online December 4, 2012Inj Prev
Motao Zhu, Peter Cummings, Haitao Chu, et al.
an exploratory age-stratified meta-analysis
vehicle crashes involving teenage drivers:
Graduated driver licensing and motor
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