O R I G I N A L R E S E A R C H
Population-based incidence of proximal radial and ulnar
fractures among adults in a Canadian metropolitan area
David M. Shepsa,b, Kyle A. R. Kempb and Kevin A. Hildebrand
ABSTRACT
Background
The lack of North American population-based incidence studies
of fractures of the proximal radius and ulna creates challenges in
the assessment of outcomes and corresponding complication
rates. Such data may help to establish consensus regarding
optimal treatments. The present study’s goal was to determine
the population-based incidence of proximal radial and ulnar
fractures in a large metropolitan area.
Methods
Over a 3-year period (April, 2002--March, 2005), cases of
proximal radial and ulnar fractures were documented and
classified according to the AO/OTA system. Overall, ageadjusted, age-specific, gender-specific, and fracture-specific
rates were calculated according to patient demographic and
2001 Canadian census data. Rates were reported as per 10,000
persons per year.
Results
1030 proximal radial and ulnar fractures were identified.
Fractures occurred at an overall rate of 5.09 (95%CI: 4.78 to
5.40), while the age-adjusted incidence was 5.14 (95% CI:
5.05--5.23). The most common fracture types observed were
B2.1 (simple articular fracture of radius, n ¼ 374), B1.1 (unifocal
articular fracture of ulna, n ¼ 280), and A2.2 (simple extraarticular fracture of neck of radius, n ¼ 145). Fracture incidence
was similar among all age groups (approximately five), with the
exception of patients ages 80 years and older (8.70; 95% CI:
6.24--11.16). Males and females had similar fracture incidences
at all ages.
Conclusions
As our results indicate similar incidences across age and gender
groups, our data is likely generalizable to the general populaa
Division of Orthopaedics, University of Calgary, Calgary, Alberta,
Canada
b
Division of Orthopaedic Surgery, University of Alberta, Edmonton,
Alberta, Canada
Funding for this study was obtained from the University of Calgary, the
Calgary Health Region, the Canadian Institutes for Health Research (CIHR),
the Calgary Orthopaedic Research Education Fund, and the Calgary Surgical
Research Development Fund. The authors would like to thank Sandy
Doolaar, Quality Improvement and Health Information, Calgary Health
Region, for her assistance with data collection for this study.
Financial Disclosure: The authors report no financial and conflicts of
interest.
Correspondence to David M. Sheps, MD, MSc, FRCSC, 10839 124th
Street NW, Edmonton, Alberta, Canada T5M 0H4
Tel: þ 780 453 2953; fax: þ 780 453 2964;
e-mail: [email protected]
1940-7041 r 2012 Wolters Kluwer Health | Lippincott Williams & Wilkins
364
Current Orthopaedic Practice
tion, which may provide further insight into the assessment of
outcomes and complication rates of such injuries.
Keywords
Canada, fracture, population-based incidence, proximal radius,
proximal ulna
INTRODUCTION
T
he study of traumatic injuries using an epidemiologic
approach is an evolving process. As with various
fractures of the upper limb, there are limited population-based reports on the incidence of proximal radial and
ulnar fractures. As most published research does not separate
proximal fractures from those occurring mid-shaft or
distally,1--5 this limits the ability to calculate the true
incidence of proximal radial and ulnar fractures. Additionally,
previous studies have employed a variety of classification
systems in the description of the type and distribution of
fractures,6--13 which may introduce observer error. Because of
these conflicting observations of incidence and patterns of
distribution of proximal radial and ulnar fractures within the
orthopaedic literature, an accurate assessment of the true
epidemiology of these injuries is difficult to ascertain.
As fractures of the proximal radius and ulna may result in
a substantial amount of disability and subsequent functional impairment, concerted efforts to document the
incidence of these fractures may improve treatment outcomes. Therefore, the goal of the present study was to
determine the overall, age-adjusted, age-specific, genderspecific, and fracture-specific incidence of proximal radial
and ulnar fractures over a 3-year period in a major Canadian
urban center. These estimates may be used as denominators
in the calculation of outcome and complication rates for
specific injuries. With the application of this information,
the orthopaedic community may gain a better understanding as to why such injuries do poorly, with the ultimate goal
of improving treatment results.
MATERIALS AND METHODS
Before commencing the study, ethics approval was obtained
from our institutional Research Ethics Boards (REB). The
present study included proximal radial fractures, ulnar
fractures, or both during a 3-year period (April, 2002--March,
2005) within a large Canadian metropolitan health region.
Inpatient discharges and ambulatory care visits among
individuals ages 18 years and older were included, as
identified with the aid of the regional administrative database
Volume 23 Number 4 July/August 2012
www.c-orthopaedicpractice.com | 365
Current Orthopaedic Practice
used in the region. This database encompasses health records
of patients treated in the three adult acute care hospitals, the
two urgent care facilities, and the three rehabilitation
hospitals within the region. These sites provide the majority
of orthopaedic care and almost all of the complex orthopaedic
care within the region.
During the 3-year period of the study, the health region
used the International Statistical Classification of Diseases
and Related Health Problems, version 10 (ICD-10) coding
system to record the diagnoses of patients at each site. Each
case was given at least one discharge diagnosis by the
treating physician most responsible for the patient’s care.
The discharge diagnosis was converted to the appropriate
ICD-10 code by the health records department, for inclusion
in the health records database.
Before conducting the database search, ICD-10 codes for
proximal radial and ulnar fractures were identified. Cases
captured from the database included those that contained at
least one eligible ICD-10 code. To ensure that duplicate cases
were not included, cases from each site were cross-referenced with all other sites, using personal health numbers,
dates of service, and type of injury.
Demographic data (date of birth, date of injury, gender)
were collected for each case. These were used to calculate
age-specific and gender-specific incidence rates and for the
standardization of incidence to the Canadian population
from the 2001 national census, which includes both age and
gender-specific population rates.14 Each patient’s address
was verified to ensure that residence was within the city
under study. Patients with a residential postal code from
outside of the city were excluded from review.
Radiographs for each identified case were collected for
review. Pretreatment radiographs also were compiled when
available. For cases in which pretreatment radiographs were
not available, post-treatment radiographs were compiled
along with relevant patient chart documents describing the
pretreatment injury. For the purpose of injury classification,
two authors, both practicing upper extremity orthopaedic
surgeons, reviewed the radiographs and relevant patient
chart documents. Each fracture was classified using the
comprehensive AO/OTA system (Table 1).15 In addition,
other classification systems were applied at various fracture
sites, whenever applicable. These include the Mason (radial
head),6 Mayo (olecranon),16 Regan and Morrey (coronoid),17
and Bado (Monteggia).18
Statistical analysis was performed using STATA version 8.0
(STATA Corporation, College Park, TX) and SPSS version 13.0
(SPSS Inc., Chicago, IL). Confidence intervals were calculated
for all incidence rates using a Poisson distribution. Standardization of the results was performed using the direct form of
standardization, using the stratum-specific rates in the study
population and the stratum structure (percentages) of the
standard population. The age-adjusted rates were weighted
averages of the age-specific rates, with the weights equal to the
proportion of the Canadian population in each age category
from the 2001 national census. Both the crude rates and the
standardized rates are included in the results section. During
the 2001 national census, the adult (18 years and older)
populations of Canada and the studied city were 23,040,965
and 674,695 persons, respectively.14
TABLE 1. AO classification system for proximal radial
and ulnar fractures
A1 Extraarticular fracture, radius intact
1 avulsion of the triceps insertion from the olecranon
2 metaphyseal simple
3 metaphyseal multifragmentary
A2 Extraarticular fracture, of the radius, ulna intact
1 avulsion of the bicipital tuberosity of the radius
2 neck simple
3 neck multifragmentary
A3 Extraarticular fracture of both bones
1 simple of both bones
2 multifragmentary of one bone and simple of the other
3 multifragmentary of both bones
B1 Articular fracture, of the ulna, radius intact
1 unifocal
2 bifocal simple
3 bifocal multifragmentary
B2 Articular fracture, of the radius, ulna intact
1 simple
2 multifragmentary without depression
3 multifragmentary with depression
B3 Articular fracture, of the one bone, with extraarticular fracture of
the other
1 ulna, articular simple
2 radius, articular simple
3 articular multifragmentary
C1 Articular fracture, of both bones, simple
1 olecranon and head of radius
2 coronoid process and head of radius
C2 Articular fracture, of both bones, the one simple and the other
multifragmentary
1 olecranon multifragmentary, radial head simple
2 olecranon simple, radial head multifragmentary
3 coronoid process simple, radial head multifragmentary
C3 Articular fracture, of both bones, multifragmentary
1 three fragments of each bone
2 ulna, more than three fragments
3 radius, more than three fragments
RESULTS
Over the 3-year study period, a total of 1,030 patients (507
males, 523 females; mean age ¼ 43.8 ± 17.5 years) with a
fracture of the proximal radius, ulna, or both were
identified. The most common fracture types observed were
B2.1 (simple articular fracture of radius, n ¼ 374), B1.1
(unifocal articular fracture of ulna, n ¼ 280), and A2.2
(simple extraarticular fracture of neck of radius, n ¼ 145).
The overall incidence of proximal radial and ulnar fractures
was 5.09 per 10,000 persons per year (95% CI: 4.78--5.40), while
the age-adjusted incidence was 5.14 (95% CI: 5.05--5.23). Men
and women had similar incidences (5.07 compared with 5.11,
respectively). Based on the AO classification, the incidence of
fracture was similar for all age groups between 18 and 79 years
of age (incidences ranging 4.62--5.53). In comparison, the
80-years-old or older cohort showed an increase in incidence
(8.70, 95% CI: 6.24--11.16). The complete list of age-specific,
gender-specific, and age-adjusted fracture incidence rates is
provided in Table 2, while fracture counts according to AO
group, age, and gender are displayed in Figure 1.
According to the other fracture sub-site classifications, the
overall rate of Mason type-I fractures (radial head) was 2.01
per 10,000 persons per year (95% CI: 1.82--2.21). Olecranon
fractures occurred at a rate of 1.12 per 10,000 persons per
Volume 23 Number 4 July/August 2012
366 | www.c-orthopaedicpractice.com
TABLE 2. Age, sex, and age-adjusted incidence rates
of proximal radial and ulnar fractures (per 10,000
persons per Year)
Group
18-29
30-39
40-49
50-59
60-69
70-79
80 þ
Females
Males
Age-adjusted
Total
Incidence Rate
95% Confidence Interval
5.12
4.75
4.62
5.53
5.43
4.90
8.70
5.11
5.07
5.14
5.09
4.49--5.76
4.12--5.37
4.00--5.24
4.66--6.40
4.28--6.57
3.62--6.19
6.24--11.16
4.67--5.55
4.62--5.51
5.05--5.23
4.78--5.40
year (95% CI: 0.97--1.26), with Mayo sub-type IIa being the
most common. Coronoid fractures occurred at a rate of 0.75
per 10,000 persons per year (95% CI: 0.63--0.87), while
Monteggia fracture-dislocations had a rate of 0.23 per
10,000 persons per year (95% CI: 0.23--0.27).
DISCUSSION
To our knowledge, this study is the first comprehensive
North American population-based study to present the
incidence of proximal radial and ulnar fractures using
expert-based classifications. Our results highlight the most
common fracture types seen, as well as the similarities in
rates between genders.
In a previous American study, Baron et al.19 presented the
population-based incidence of fractures of the proximal
radius and ulna in patients between the ages of 65 and 89.
However, their results were based on gender and race, and
overall rates were not presented. Since the authors did not
provide the population at risk, it is difficult to calculate the
combined rate. As such, the authors presented rates of
1.5--8.6 per 10,000 persons per year, which compare
favorably to the rates of 5.43--8.70 per 10,000 persons per
year in the present study.
Examining the counts and incidence rates of proximal
radial and ulnar fractures separated by AO group provides
some insight into the frequency of occurrence of these
fractures. Group B fractures are by far the most common
type, occurring at a rate more than four times that of group
A fractures and more than 13 times that of group C. Within
group A, the incidence rates show a similar rate of
occurrence of these fractures in all age groups and in both
sexes, with rates that range from 0.72--0.96 per 10,000
persons per year. The most common type in group A, type
A2.2 or fractures of the radial neck, occurs in a bimodal
distribution. The rates in the younger and older age groups
are closer to 0.8 per 10,000 persons per year, compared with
rates of 0.5 per 10,000 persons per year in the middle age
groups. However, there is a significant drop-off in rates in
the oldest age group (0.18 per 10,000 persons per year). This
FIGURE 1. Observed fracture counts according to AO group, age and gender. Type A (RA) refers to extraarticular fractures. Type B (RB) are articular
fractures of a single bone, with or without an extraarticular fracture of the other bone. Type C (RC) are articular fractures of both bones.
www.c-orthopaedicpractice.com | 367
Current Orthopaedic Practice
may again be attributable to the small size of this group,
where an increase in one A2.2 injury would have had a
dramatic affect on the incidence rate for this group. In
addition, there is overlap of the confidence intervals,
particularly in the older age groups.
Within group B, the incidence rates are similar in all age
groups at approximately four per 10,000 persons per year,
with the exception of individuals age 80 and older in whom
the rate almost doubles (7.25 per 10,000 persons per year).
The two most common types in group B, types B1.1 (unifocal
intraarticular fractures of the proximal ulna), and B2.1 (simple
undisplaced and displaced fractures of the radial head) occur
in opposite distribution patterns. The incidence rate of B1.1
fractures increases with increasing age, with a six times greater
rate in the group age 80 and older compared with the younger
age groups (6.34 compared with 1.02 per 10,000 persons per
year). The incidence rate of B2.1 fractures decreases with
increasing age, with a 2.5 times greater rate in the younger age
group than the older age group (2.48 compared with 0.91 per
10,000 persons per year). The incidence rate patterns for
fractures of the proximal radius and ulna, separated by AO
group and subgroup, previously has not been described in the
literature.
The overall incidence of observed fracture was similar across
all age groups between 18 and 79 years of age, as well as
between genders. It is worth noting, however, that the rate of
proximal radial and ulnar fractures sharply increases among
persons 80 years and older. Although the mechanism of injury
was not captured as part of our methodology, this increase
may be from falls or poor bone quality among this age group.
There were limitations in our study that warrant discussion.
These include, but are not limited to, misclassification of cases
at initial treatment, errors in the population database and the
possibility of selection bias on the part of the authors.
First, by looking at fracture injuries, we were able to
estimate the presence of misclassification. Although the
study was successful at measuring the accuracy of the
current coding system, it did not determine at what point
any miscoding might have occurred. Miscoding may be the
result of an incorrect discharge diagnosis with the correct
associated code, a correct discharge diagnosis with an
incorrect associated code, or both an incorrect discharge
diagnosis and an incorrect associated code. Although the
measured percent agreement provides an estimate of the
degree of miscoding, the true source of the miscoding could
not be determined. The present study was done as part of a
larger project, in which the percent agreement between the
authors and health region’s database coding system was
determined to be 59.03%.
Second, as classification of the radiographs and chart
notes by the authors is the foundation upon which the
estimates of the incidence rates are based, we attempted to
identify sources of misclassification that may exist when
identifying injuries from a database such as the region’s
health records database. However, misclassification may also
occur during the authors’ classification of the identified
injuries. Although both authors can be considered experts in
the field under study, the past clinical experience of the
authors affects the classification of the injuries based on the
radiographs and chart notes. The relatively poor reliability
of the classification systems would suggest that a percentage
of the classified cases would be open for discussion and that
another set of observers may choose to classify the cases
differently.20,21 Ultimately, this would lead to a degree of
misclassification of the identified cases by the study authors
and would affect the estimated incidence rates. Since a
reference standard for the authors was not used, this
misclassification could not be accurately measured.
Third, the denominator data used for the calculation of
injury incidence rates were taken from the 2001 Canadian
census, which provides estimates of the overall, age-specific
(by decade), and sex-specific population of the city in the
present study in 2001. During the period under study (from
2002--2005), there was a change in the population of the
city, with an increase in the overall population and an
increase in certain age ranges (most likely younger age
ranges). Therefore, for the purposes of this study, the use of
the 2001 census data would result in an overestimation of
the incidence rates, because of the increase in the population of the city during the period under study. This would
have resulted in an increase in the value of the denominator
used in the incidence rate calculations, and thus a decrease
in the calculated incidence rate. However, the presentation
of the data with the associated 95% confidence intervals
provides the reader with a range of possible values for the
estimates of the incidence rates and takes into account the
uncertainty associated with the use of the 2001 census data
as the denominator in the incidence rate calculations.
Lastly, potential sources of selection bias in this study
merit consideration. There were limitations in the health
records database in regard to capturing all cases. For
example, if a fracture was managed by a primary care
physician, a private practice orthopaedic surgeon, or under
the auspices of the Worker’s Compensation Board, the
health region’s database may not capture that injury. This
would have resulted in an underestimation of the actual
incidence of some injuries. Because records outside of the
region’s health records database cannot be sampled because
of privacy and logistical reasons, it is impossible to measure
the potential magnitude of this selection bias.
In conclusion, the information presented in this study
highlights the relative magnitude of the clinical burden of
proximal radial and ulnar fractures. The incidence rates
presented provide a population-based denominator that
may be used for future research, allowing for a better
estimation of the percentages of positive and negative
outcomes as well as complications.
REFERENCES
1. Contreras L, Kirschbaum A, Pumarino H. Epidemiology of
fractures in Chile. Rev Med Chil. 1991; 119:92--98.
2. Donaldson LJ, Cook A, Thomson RG. Incidence of fractures in a
geographically defined population. J Epidemiol Community
Health. 1990; 44:241--245.
3. Garraway WM, Stauffer RN, Kurland LT, et al. Limb fractures in a
defined population. I. Frequency and distribution. Mayo Clin
Proc. 1979; 54:701--707.
4. Sahlin Y. Occurrence of fractures in a defined population: a
1-year study. Injury. 1990; 21:158--160.
Volume 23 Number 4 July/August 2012
368 | www.c-orthopaedicpractice.com
5. van Staa TP, Dennison EM, Leufkens HG, et al. Epidemiology of
fractures in England and Wales. Bone. 2001; 29:517--522.
6. Hotchkiss RN. Fractures and dislocations of the elbow. In:
Rockwood CA, Green DP, Bucholz RW, et al., eds. Rockwood and
Green’s Fractures in Adults. Philadelphia, PA: Lippincott-Raven;
1996:529.
7. Mason ML. Some observations on fractures of the head of the
radius with a review of one hundred cases. Br J Surg. 1954;
42:123--132.
8. Cutler CW. Fractures of the head and neck of the radius. Ann
Surg. 1926; 83:267--278.
9. Poulsen JO, Tophoj K. Fracture of the head and neck of the
radius. Follow-up on 61 patients. Acta Orthop Scand. 1974;
45:66--75.
10. Stankovic P. Surgical management of fractures of the proximal
end of the radius. Chirurg. 1978; 49:377--381.
11. Johnston GW. A follow-up of one hundred cases of fracture of
the head of the radius with a review of the literature. Ulster Med
J. 1962; 31:51--56.
12. Gutierrez G. Management of radial head fracture. Am Fam Phys.
1997; 55:2213--2216.
13. Morrey BF. Management of radial head fracture. In: Morrey BF,
ed. The Elbow and its Disorders. Philadelphia, PA: WB Saunders;
2000.
14. Statistics Canada. Age (123) and Sex (3) for Population, for
Canada, Provinces, Territories, Census Metropolitan Areas
and Census Agglomerations, 1996 and 2001 Censuses---100%
15.
16.
17.
18.
19.
20.
21.
Data. Retrieved from: http://www12.statcan.ca/english/census01/
products/standard/themes/RetrieveProductTable.cfm?Temporal ¼
2001&PID ¼ 55521&APATH ¼ 3&METH ¼ 1&PTYPE ¼ 55440&TH
EME ¼ 37&FOCUS ¼ 0&AID ¼ 0&PLACENAME ¼ 0&PROVINCE
¼ 0&SEARCH ¼ 0&GC ¼ 99&GK ¼ NA&VID ¼ 0&VNAMEE ¼
&VNAMEF ¼ &FL ¼ 0&RL ¼ 0&FREE ¼ 0&GID ¼ 431637.
Muller ME, Nazarian S, Koch P, et al. The comprehensive
classification of fractures of long bones. Berlin: Springer-Verlag;
1994.
Cabanela ME, Moorey BF. Fractures of the olecranon. In:
Moorey BF, ed. The Elbow and its Disorders. Philadelpha, PA:
WB Saunders; 2000:365--379.
Regan WD, Morrey BF. Coronoid process and Monteggia
fractures. In: Morrey BF, ed. The Elbow and its Disorders.
Philadelphia, PA: WB Saunders; 2000:396--408.
Bado JL. The Monteggia lesion. Clin Orthop Relat Res. 1967;
50:71--86.
Baron JA, Karagas M, Barrett J, et al. Basic epidemiology of
fractures of the upper and lower limb among Americans over 65
years of age. Epidemiology. 1996; 7:612--618.
Morgan SJ, Groshen SLB, Itamura JM, et al. Reliability evaluation of classifying radial head fractures by the system of Mason.
Bull Hosp Jt Dis. 1997; 56:95--98.
Wainwright AM, Williams JR, Carr AJ. Interobserver
and intraobserver variation in classification systems for fractures of the distal humerus. J Bone Joint Surg. 2000; 82-B:
636--642.