Systematic Review of Methods and Results of Studies of the Genetic Epidemiology of Ischemic Stroke
Background and Purpose— To design appropriate molecular genetic studies, we first need to understand the genetic epidemiology of stroke. We therefore performed a systematic review of the literature to assess the heritability of stroke according to methodological quality of studies and to determine any heterogeneity in findings between studies and possible publication bias.
Methods— We searched for twin studies and studies of family history of stroke using bibliographic databases and by hand-searching reference lists and journals. Odds ratios (ORs) for family history as a risk factor for stroke were calculated within studies and combined by meta-analysis. Heterogeneity between studies, methodological quality of studies, and the influence of the age at which stroke occurred in both probands and relatives were assessed.
Results— We identified 53 independent studies (3 twin, 33 case-control, and 17 cohort). Monozygotic twins were more likely to be concordant than dizygotic twins (OR, 1.65; 95% CI, 1.2 to 2.3; P=0.003). A positive family history was a risk factor for stroke in both case-control (OR, 1.76; 95% CI, 1.7 to 1.9; P<0.00001) and cohort (OR, 1.30; 95% CI, 1.2 to 1.5; P<0.00001) studies. However, there was major heterogeneity between studies (cohort P=0.0001; case-control P<0.00001), with much stronger associations in small studies and methodologically less rigorous studies. Moreover, studies that reported insufficient data to allow meta-analysis tended to have found weaker associations. Family history of stroke was more frequent in studies that were confined to probands or relatives aged <70 years. However, few studies considered the number of affected and unaffected relatives, only 2 studies considered stroke phenotypes in detail, and only 19 studies (38%) adjusted associations for intermediate phenotypes. No twin study, only 5 cohort studies (26%), and 20 case-control studies (61%) differentiated between ischemic and hemorrhagic stroke in the proband. Family history of stroke was more frequent in large- and small-vessel stroke than in cardioembolic stroke. There were very few data on the influence of family history on stroke severity and no data on stroke recovery.
Conclusions— Twin studies suggest a small genetic contribution to stroke, but reliable interpretation of published family history studies is undermined by major heterogeneity, insufficient detail, and potential publication and reporting bias. More detailed large-scale genetic epidemiology is required.
Afamily history of stroke is regarded as an important risk factor for the development of cerebrovascular disease, and strokes cluster in families.1 Animal models suggest that susceptibility to ischemic stroke is influenced by genetic factors,2 and there are a number of rare mendelian stroke syndromes in humans.3,4 However, several dozen candidate gene studies have produced no consistent results, and data on the genetic epidemiology of stroke are conflicting.4–6 One reason for this is that although ischemic stroke is a highly complex trait, few studies have assessed stroke subtypes, and no studies have been properly powered to do so. Moreover, many studies have combined ischemic and hemorrhagic strokes. It is unlikely that these very different pathological conditions are under the same genetic influences. Of equal importance, little account has been taken of the role of intermediate phenotypes (eg, hypertension, hyperlipidemia, diabetes, carotid stenosis), many of which have a substantial genetic component themselves. Furthermore, a positive family history could be the result of shared genes, shared environment, or both.
Several genome screens and other detailed molecular genetic studies of ischemic stroke are planned. However, if these are to be properly targeted, it is essential that we first understand the basic genetic epidemiology. Recent reviews of the genetics of ischemic stroke have not sought to be systematic and have cited only 11 to 16 family history studies and 1 twin study.4–6 We performed a systematic review of all published twin studies and studies of family history as a risk factor for stroke. We sought to estimate effect sizes, to determine the extent of heterogeneity in the strength of associations between studies, and to explain any differences by assessing the quality of studies, searching for possible publication bias, and exploring whether effects varied between stroke subtypes, age of stroke onset in probands and relatives, and the presence of intermediate phenotypes.
Materials and Methods
We sought to identify articles that reported on family history as a potential risk factor for stroke. Studies were identified by 2 independent observers from MEDLINE+ and EMBASE (Silverplatter Winspirs 4.0 online and Entrez PubMed NIH 06/08/2001 for 1966 to May 2003) with the following search terms: family history AND (stroke OR CVA OR TIA OR cerebrovascular) and twin AND (stroke OR CVA OR TIA OR cerebrovascular). No restriction was made on the language of publication. Journals that yielded >10% of all studies identified electronically were systematically hand-searched for further relevant studies published after 1980. The reference lists of all articles that met the inclusion criteria were searched. We contacted authors personally if their publications were unavailable in the United Kingdom.
Articles were included in the review if they fulfilled the following criteria: (1) study was a prospective cohort, case-control, or twin study; and (2) study reported on the frequency of a positive family history of stroke for patients with stroke and a control group for case-control studies, or the risk of stroke (either undifferentiated or ischemic) in patients with and without a positive family history of stroke for prospective cohort studies, or the concordance for stroke in monozygotic versus dizygotic twins.
The following data were extracted from eligible reports by 2 independent observers with a structured questionnaire: (1) type of study (case-control, cohort, or twin); (2) setting (hospital or population based); (3) length of follow-up (for cohort studies); (4) details of patient selection (Were patients recruited consecutively or by type of stroke? Were patients with transient ischemic attack [TIA] or subarachnoid hemorrhage included?); (5) details of collected family history (Did studies also collect information on family history of ischemic heart disease, hypertension, peripheral vascular disease, TIA, or diabetes mellitus? Were only fatal events included? Which relatives were included in a positive family history? Were the age and number of affected relatives and the size of the family taken into account? What method was used to collect a family history?); (6) details of stroke classification in patients (ischemic versus hemorrhagic; subtyping of ischemic strokes according to the TOAST7 [or similar] criteria); (7) consideration of whether other potential risk factors (age at event, personal history of hypertension, diabetes mellitus, ischemic heart disease, peripheral vascular disease, carotid disease, cardioembolism, smoking, cholesterol, alcohol consumption, body mass index, use of hormone replacement therapy and/or oral contraceptive pill, social class/education, employment status, marital status, amount of exercise) collected and related to the presence of family history of stroke; (8) results and conclusion of each study (Was a positive family history found to be a risk factor for stroke? If so, was there a difference for the various stroke subtypes? Was a positive family history associated with any confounders or established risk factors or environmental factors? Did the study correct for these potential confounders? Did the study consider genetic versus environmental factors in a positive family history?); and (9) consideration of whether studies collected information on the influence of family history of stroke on stroke severity and recovery.
Studies that gave absolute numbers for patients with family history of stroke were identified. Only patients with ischemic stroke were counted if the subtypes were differentiated. A parental family history of stroke was used for the meta-analysis in which family history was reported separately for whether strokes had occurred in parents or siblings and in which it was not possible to derive an overall estimate for family history in first-degree relatives (FDR).
Odds for a positive family history as a risk factor for stroke were calculated within individual studies. Where appropriate, odds ratios (ORs) from separate studies were combined by fixed-effects meta-analysis according to the Mantel-Haenszel method. Heterogeneity between studies was calculated with the χ2 method. The possibility of publication bias was assessed by checking funnel plots of the inverse standard error (precision) of studies versus the OR for asymmetry and by simple linear regression analysis of the standard normal deviate (lnOR/SE) versus precision (1/SE). The predicted regression line assuming no bias was compared with the actual simple regression fit (P<0.10 was regarded as significant).8 Both cohort and case-control studies measure the odds of stroke that are conferred by a positive family history of stroke. We therefore tested for potential publication bias of both study types individually and combined.
We predefined a simple score to assess the methodological quality of studies, giving 1 point each if a study (1) recruited patients consecutively; (2) defined the relatives who contributed to the family history; (3) separated family histories of stroke, hypertension, or myocardial infarction (MI); (4) took the age of onset of stroke in relatives into account; (5) took the number of affected relatives into account; (6) distinguished between ischemic and hemorrhagic strokes; and (7) subtyped ischemic strokes by etiology. Studies were then stratified into 3 strata if they met <3, 3 to 5, and ≥6 quality criteria. Heterogeneity of OR estimates between the resulting strata was calculated. To detect heterogeneity in relation to the power of studies, we stratified them into tertiles according to the inverse of their standard errors. A possible correlation between study quality and precision was also tested for by the rank correlation test of Spearman.
To identify other sources of heterogeneity, year of publication (divided into studies published before and after 1990), type of study (prospective cohort or case-control), median age of subjects, maximum age of subjects, maximum age of relatives, whether hemorrhagic strokes were included or not, and family history of fatal stroke only were tested versus the natural logarithm of the OR weighted by the inverse variance.
The electronic literature search yielded 889 publications with the search terms family history AND (stroke OR CVA OR TIA OR cerebrovascular) and 105 publications for the search terms twin AND (stroke OR CVA OR TIA OR cerebrovascular). Duplicate records were removed, and abstracts were reviewed. Ninety-seven publications were judged potentially relevant by 1 or both observers. Review of reference lists of these articles and a hand-search of the journal Stroke for the years 1980–2003, which was the only journal that met the criteria for hand-searching, yielded an additional 127 publications. Thus, a total of 224 articles (9 in languages other than English) were considered in detail. Two of the articles were published in Spanish, 2 in Italian, 2 in German, and 1 each in Russian, Chinese, and Hungarian.
Sixty publications satisfied our inclusion criteria. They consisted of 3 twin studies that investigated the concordance of stroke in monozygotic versus dizygotic twins (1 group published a follow-up report on their original article as an abstract),9–12 19 articles that published data on 17 independent cohort studies,13–28,65–67 and 37 articles that published data on 33 independent case-control studies.29–64,68 Of these, 9 cohort studies15,18,20,22–25,28,66 and 6 case-control studies33,34,36,40,43,54 only reported relative risks or ORs and gave no absolute numbers and therefore could not be included in any meta-analysis. However, details of these studies are given in Table 1. Thus, 3 twin studies,9–12 9 cohort studies,* and 27 case-control studies† reported sufficient data to allow inclusion in the meta-analysis. One additional cohort study was not included because outcome strokes in probands were defined only as MRI evidence of infarction without data on clinical stroke outcomes.69
|Study||Cohort Type||Country||Subjects, n||Age, y||Length of Follow-Up, y||FHx of Event|
|TIA indicates transient ischemic attack; sibs, siblings; FHx, family history; N/A, not applicable; FDR, first-degree relatives; SD, standard deviation; SDR, second-degree relatives; MI, myocardial infarction; RR, relative risk; MRR, multivariate adjusted relative risk; HTN, hypertension; OR, odds ratio; IHD, ischemic heart disease; DM, diabetes mellitus; PVD, peripheral vascular disease; SAH, subarachnoid hemorrhage; ICH, intracerebral hemorrhage; CI, cerebral infarction; Cnsct, patients were recruited consecutively.|
|A. Prospective cohort studies|
|Sesso et al 200122||Professional||USA||22071 M 39876 F||40–84 M ≥45 F||13 M 6.2 F||MI|
|Voko et al 200025||Population||Netherlands||7603||?||?||Stroke|
|Menotti, Giampaoli 199820||Population||Italy||1527||40–59||35||MI, HTN, DM|
|Simons et al (1998)24||Population||Australia||2805||>60||?||IHD|
|Kiely et al (1993)18||Population||USA||(1) 4933 parents||30–62||36||Fatal stroke|
|(2) 2317 offspring of parents in (1)||>30||16||Stroke or TIA|
|Stroke or TIA or IHD|
|(3) 604 sibships of subjects in (2)||Stroke or TIA|
|Shaper et al (1991)23||Community||England||7735||40–59||8||Fatal stroke, IHD|
|Harmsen et al (1990)15 (cf Wilhelmsen )28||Community||Sweden||7495||47–55||11.8||Stroke|
|Okada et al (1976)66||Community||Japan||4737||40–79||7||Stroke, HTN, IHD|
|Study||Setting||Country||Cnsct||No. of Patients/ Controls||Age, y||FHx of Event||Relatives||Differentiation of Strokes|
|B. Case-control studies|
|Becher et al (2000)33||Hospital||Germany||?||197/197||65 (22–80)||Stroke||FDR||Ischemic|
|Halim et al (1998)43||?||USA||?||?||>39||Stroke||FDR||Ischemic|
|Carrieri et al (1994)36||Hospital||Italy||Yes||164/164||40–75||Stroke||Not defined||Ischemic|
|Bharucha et al (1988)34||Population||India||Yes||111/111||?||Stroke||FDR||Ischemic|
|Marshall (1971)54||Hospital||England||?||201/national mortality statistics||?||Fatal stroke||Parents, siblings||Nonembolic ischemic|
|Gertler et al (1968)40||Hospital||USA||Yes||185/?||62 ±13||Stroke, IHD, HTN, DM||Parents, siblings||Undifferentiated|
|Relatives||Differentiation of Outcomes||Result|
|Parents||Undifferentiated||M, MRR 1.03 (0.67–1.60); F, MRR 1.45 (0.80–2.62)|
|FDR||Undifferentiated||Adjusted: (1 FDR) RR 1.3 (1.0–1.64)|
|(>1 FDR) RR 1.5 (1.0–2.4), (one FDR <65) RR 1.6 (1.1–2.2), (>1 FDR <65) RR 2.0 (0.6–6.4)|
|Parents||Fatal||No significant association for either FHx|
|Unspecified||Ischemic or fatal||No influence|
|Parents||Undifferentiated||Crude RR 1.07 (0.89–1.28), adjusted (age, sex) RR 1.06 (0.88–1.27), multivariate RR 0.99 (0.82–1.19)|
|Parents||Undifferentiated||Crude RR 1.9 (0.95–3.74), adjusted (age, sex) RR 1.68 (0.85–3.33), multivariate RR 1.56 (0.76–3.19)|
|Parents||Undifferentiated||Crude RR 3.6 (1.49–8.67), adjusted (age, sex) RR 2.99 (1.23–7.26), multivariate RR 3.33 (1.27–8.72)|
|Siblings||Undifferentiated||Sibship size adjust. RR 1.50 (0.80–2.82), adjusted (age+sex) RR 1.23 (0.66–2.32), MRR 1.19 (0.61–2.32)|
|Siblings||Atherothrombotic||Sibship size adjust. RR 3.39 (1.41–8.16), adjusted (age+sex) RR 2.52 (1.05–4.94), MRR 1.83 (0.68–4.94)|
|Parents||Undifferentiated||No effect on risk of stroke|
|Parents||Differentiated||OR (SAH) 1.7 (0.6–5.3), (ICH) 2.1 (0.7–6.0), (CI) 1.4 (0.9–2.5), (all stroke) 1.5 (1.05–2.1), multivariate adjusted (ischemic strokes) FHx not significant|
|Parents||Undifferentiated||No significant association on multivariate analysis|
|Not specified||Differentiated||Cerebral thrombosis (age+sex adjusted): RR FHxstroke 0.71, FHxHTN 1.80, FHxIHD 1.41 all nonsignificant|
|Cerebral hemorrhage (age+sex adjusted): RR FHxstroke 1.44 P<0.05, FHxHTN 1.77, FHxIHD 0.93|
|Crude OR 1.54 (0.83–2.85), attributable risk 0.09 (−0.04–0.19)|
|RR 2.2 (1.4–3.6) independent of HTN, smoking|
|Crude OR (pt<55) 12.18 (7.55–58.03), OR (pt>55) 1.87 (1.1–4.7), adjusted (age, BP) OR (pt<55) 5.08 (?), OR (pt>55) 1.67 (0.92–3.12)|
|No overall excess of FHx|
|Higher than expected in normal population|
The characteristics and main results of the studies are summarized in Table 2. The 2 Scandinavian twin studies9,12 used national death or hospital discharge registries to identify twins with stroke. The original report of the American study employed a mailed questionnaire10; the follow-up report additionally used a telephone health screen and a mortality review.11 None of the twin studies distinguished between stroke subtypes or assessed confounding by other risk factors. In the meta-analysis, monozygotic twins were more likely to be concordant for stroke than dizygotic twins (OR, 1.65; 95% CI, 1.2 to 2.3; P=0.003; heterogeneity P=0.3; Figure 1).
|MZ indicates monozygotic twins, DZ, dizygotic twins; Df, degrees of freedom.|
|Bak et al (2002)9||Danish Twin Registry (≈11 500 same-sex twin pairs with known zygosity 3852 MZ, 7712 DZ followed up for stroke death or hospitalization for stroke)||Stroke types not differentiated||Concordance for stroke death|
|Age 14–73 y||MZ 0.18 (0.14–0.22)|
|Follow-up up to 51 years||DZ 0.10 (0.08–0.13)|
|Heritability estimate 0.32|
|Concordance for stroke hospitalization or stroke death|
|MZ 0.19 (0.15–0.24)|
|DZ 0.13 (0.10–0.16)|
|Heritability estimate 0.17|
|Brass et al (1996)11||National Academy of Science-NRC twin registry (4345 MZ, 5240 DZ, and 1261 unknown twin pairs)||Stroke types not differentiated||Concordance for stroke or stroke related death|
|Age 58–68 y||MZ 12.8%|
|Follow-up up to 10 y||DZ 8.0%, χ2 = 4.085 P<0.05|
|Telephone health screen, mortality review||Df1|
|Brass et al (1992)10||National Academy of Science-NRC twin registry (15 948 male twin pairs born between 1917 and 1927; 9475 twins responded to mailed questionnaire, 1382 MZ, 1221 DZ, 119 unknown complete twin pairs included)||Stroke types not differentiated||Concordance for stroke|
|Age 58–68 y||MZ 17.7%|
|Cross-section||DZ 3.6%, χ2 = 4.94 P<0.05|
|de Faire et al (1975)12||Swedish Twin Registry (11 000 same-sex twin pairs born between 1886 and 1925 alive in 1961, 3654 MZ, 6842 DZ, 439 unknown twin pairs followed up for cause of death)||Stroke types not differentiated||Concordance for fatal stoke|
|Age 37–52 y||Similar for MZ and DZ twins|
|Follow-up up to 12 y|
Characteristics of the case-control studies included in the meta-analysis are summarized in Table 3. Of these, 15 independent studies differentiated between ischemic and hemorrhagic strokes,‡ and 4 studies subtyped ischemic strokes further into cardioembolic, large-artery disease, small-artery disease, and undetermined,49,59 into cortical artery occlusion and perforating artery occlusion,60 or included only patients with large-vessel stroke.46 Most studies defined which relatives were considered in a family history (mostly FDR: parents and/or siblings), but 3 studies did not.29,55,58 Two studies used a pooled family history of stroke or hypertension or MI.48,53
|Study||Setting||Country||Consec.||Age, y (Range)||FHx|
|TIA indicates transient ischemic attack; sibs, siblings; FHx, family history; N/A, not applicable; FDR, first-degree relatives; SD, standard deviation; SDR, second-degree relatives; MI, myocardial infarction; DM, diabetes mellitus; HTN, hypertension; PVD, peripheral vascular disease.|
|Jerrard-Dunne et al (2003)49 (cf Hassan44)||Hospital||England||Yes||64.4 (SD 8.7)||Stroke, MI|
|Hassan et al (2002)44 (cf Jerrard-Dunne49)||Hospital||England||Yes||65.43 (SD11.4)||Stroke|
|Polychronopoulos et al (2002)59||Hospital||Greece||Yes||67.6 (SD11.8)||Stroke|
|Starr et al (2001)62||Hospital||Scotland||Yes||70.3 (36.6–94.5)||Fatal stroke, HTN, DM, other vascular diseases|
|Peng et al (1999)57,58||Hospital||China||?||62.6±8.9||Stroke, MI, HTN, DM|
|Caicoya et al (1999)35||Hosp+primary care||Spain||Yes||70.8 (40–85)||Stroke, MI, HTN, DM|
|Feigin et al (1998)38||Population||Russia||Yes||67.8±9.2||Stroke|
|Kubota et al (1997)50||Hospital||Japan||Yes||58 (SD 8.9)||Stroke|
|Liao et al (1997)52||Cross-section of population cohort||USA||N/A||60 (SD 6.6)||Stroke|
|Toyoshima et al (1997)68||Cross-section of rural population||Japan||N/A||35–74||Stroke, HTN, DM, cancer, TB|
|Vitullo et al (1996)64||Hospital||Italy||? Yes||30–69||Stroke, MI|
|Graffagnino et al (1994)42||Hospital||Canada||Yes||64.6 (SD 8.7)||Stroke, MI|
|Margaglione et al (1994)53||Hospital||Italy||?||63.8 (31–86)||Stroke or MI|
|Shintani et al (1993)60||?||Japan||?||61.9 (25–73) stroke onset <65||Stroke|
|Fonte et al (1993)39||Hospital||Italy||Yes||77.3±7.3||Stroke|
|Muñiz et al (1993)56||Hospital||Spain||?||62.6±13.4||Stroke, HTN|
|Spriggs et al (1990)61||Hospital||England||Yes||74 (33–97)||Stroke, MI, HTN, PVD, DM|
|Matias-Guiu et al (1990)55 (cf Alvarez et al 32)||Hospital||Spain||Yes||42.9 (15–50)||Stroke, MI, HTN, DM|
|Li et al (1990)51||Population cross-section||Rural China||N/A||?||Stroke, HTN|
|Thompson et al (1989)63||General practice||UK||Yes||45–69||Stroke, MI|
|Hu et al (1989)47||Cross-section, cluster sampling||Taiwan||N/A||37–85||Stroke, MI, HTN, DM|
|Diaz et al (1986)37||Hospital||Canada||Yes||69.3 (SD 6.6)||Stroke, MI, HTN, DM|
|Herman et al (1983)45||Hospital||Netherlands||Yes||40–74||Stroke, MI, HTN, DM|
|Abu-Zeid et al (1977)29||Hospital||Canada||Yes||66.99±14.1||Stroke|
|Alter and Kluznik (1972)31 (cf Alter 30)||Hospital||USA||?||?||Stroke, MI, DM, HTN|
|Heyden et al (1969)46||Hospital||USA||?||58 (42–81)||Fatal stroke, MI|
|Issaeva and Mikheev (1967)48||Hospital||Russia||?||30–70||Stroke or HTN or MI|
|Gifford (1966)41||Hospital||USA||?||Median 60–64||Fatal strokes, MI|
|Characteristics of Affected Relatives Considered||Relatives||Differentiation||Control Group|
|Yes||Yes||Parents/sibs||Ischemic subtyped into large artery, small artery, cardioembolic, and undetermined||Spouse and community sampling controls (age- +sex-matched)|
|Yes||Yes||Parents/sibs||Ischemic||Spouse and community sampling controls (age- +sex-matched),|
|No||No||FDR||Hemorrhagic and ischemic (ischemic subtyped into large artery, small artery, cardioembolic, and undetermined)||Matched controls|
|Yes||No||Parents||Not differentiated for FHx of stroke||Population controls (age-/sex-/parental occupation-matched)|
|No||No||Not defined||Ischemic||Hospital controls (age- +sex-matched)|
|No||No||FDR||Not differentiated for FHx of stroke||Population controls (age- +sex-matched)|
|No||No||Parents/sibs||Ischemic||Population controls (age- +sex-matched)|
|No||No||Parents/grandparents||Ischemic and hemorrhagic||Outpatient controls (age- +sex-matched)|
|No||No||FDR||Not differentiated||(Some overlap with ARIC study21)|
|No||No||FDR, SDR separately||Ischemic (cardioembolic excluded)||Population controls (age- +sex-matched)|
|Yes||No||Parents/sibs||Ischemic||Outpatient controls with cardiovascular risk factors|
|No||No||Parents/sibs||Ischemic, cortical, and lacunar differentiated||“Normal control subjects”|
|<65||No||FDR||Ischemic both stroke and TIA||Hospital controls (age- +sex-matched)|
|No||No||Parents/sibs||Not differentiated||Hospital controls (age-/sex-/residence-matched)|
|No||No||FDR||Not differentiated||GP register controls (age- +sex-matched)|
|No||No||Not defined||Ischemic||Controls (age- +sex-matched)|
|No||No||FDR, grandparents||Ischemic and hemorrhagic|
|No||No||Parents/sibs||Not differentiated||GP practice controls (age- +sex-matched)|
|Yes||No||Sibs||Ischemic, TIA included||Spouse controls|
|No||No||Parents/sibs||Not differentiated||Hospital controls (age- +sex-matched)|
|No||No||Not defined||Ischemic and hemorrhagic||Hospital controls (age-/sex-/residence-matched)|
|Yes||Yes||Parents/sibs||Not differentiated||Spouse controls|
|Yes||No||Parents||Ischemic, large-vessel||Hospital controls (age-/sex-/race-matched)|
|No||No||Parents||Not differentiated||Hospital controls|
|No||No||Parents/sibs||Not differentiated||Hospital controls|
The combined OR for a positive family history of stroke as a risk factor for stroke was 1.76 (95% CI, 1.7 to 1.9; P<0.00001; 5991 patients). However, there was significant heterogeneity between studies (P<0.00001; Figure 2). The funnel plot of the OR versus the variance was asymmetrical (Figure 3), and there was evidence of possible publication bias on simple linear regression of the standard normal deviate against precision (P=0.003). However, studies cited in recent reviews4–6 had an overall similar association of family history of stroke with stroke risk (OR, 1.56; 95% CI, 1.4 to 1.8; P<0.00001; heterogeneity P=0.05; 1665 patients). The OR did not change significantly when only the 15 studies exclusively investigating ischemic stroke were combined (OR, 1.75; 95% CI, 1.6 to 2.0; P<0.00001; heterogeneity P=0.005; 3800 patients).
Studies that fulfilled at least 6 methodological quality criteria found a weaker association with family history of stroke (OR, 1.28; 95% CI, 1.1 to 1.5; P=0.01; heterogeneity P=0.27; 1221 patients) than studies that fulfilled more than half but <6 (OR, 1.88; 95% CI, 1.7 to 2.0; P<0.00001; heterogeneity P=0.0001; 4280 patients) and studies that fulfilled less than half the methodological quality criteria (OR, 2.40; 95% CI, 1.8 to 3.2; P<0.00001; heterogeneity P=0.0001; 490 patients). The difference between the 3 strata was significant (P=0.00006; Figure 4). There was no correlation (r=0.085, P=0.67) between the inverse standard error (precision) of a study and the quality score.
Ten studies adjusted the ORs for a family history of stroke at least partially for the presence of other established risk factors of stroke in patients,§ and 1 study62 adjusted the ORs for the presence of risk factors for stroke in relatives. They found with 3 exceptions35,49,58 that adjustment either attenuated the association with family history of stroke or that family history was no longer a significant risk factor after adjustment (Table 4). Two studies that investigated the influence of environmental factors other than smoking on family history of stroke as a risk factor for stroke found no significant association.35,64 Ten∥ of 12 studies56,59 investigating a possible interaction between family history of stroke and hypertension in patients found a significant association; 3 studies found an association with smoking,35,44,52 3 found an association with diabetes mellitus,37,44,52 4 found an association with the presence of ischemic heart disease or vascular disease,31,37,46,52 and 1 found an association with the presence of hyperlipidemia.42 Two studies found that family history of stroke was significantly related to a cluster of vascular risk factors.42,37
|OR Crude (95%CI)||OR (Age, Sex) (95%CI)||OR Multivariate (95%CI)|
|Morrison et al21||1.11 (0.85–1.43)||1.05 (0.81–1.37)|
|Jousilahti et al16|
|Men||1.53 (0.89–2.65)||1.51 (0.88–2.61)|
|Women||1.71 (1.03–2.84)||1.79 (1.08–2.97)|
|Wannamethee et al26||1.6 (1.2–2.1)||1.4 (1.1–2.0)|
|Peng et al58||5.5 (1.9–16.1)||6.2 (1.3–32.3)|
|Caicoya et al35||1.74 (1.27–2.56)||1.79 (1.25–2.56)|
|Kubota et al50||1.41 (0.83–2.39)||0.82 (0.39–1.74)|
|Vitullo et al64||1.4 (0.9–2.1)||1.3 (0.8–2.1)|
|Jerrard-Dunne et al49||1.20 (0.98–1.48)||1.22 (0.90–1.39)|
|Polychronopoulos et al59||2.41 (1.67–3.48)||2.40 (1.65–3.36)||2.06 (1.39–3.04)|
|Liao et al52||1.73 (1.17–2.56)||1.60 (1.08–2.39)||1.56 (1.02–2.38)|
Only 2 recent studies49,59 compared the association of a positive family history of stroke between ischemic stroke subtypes. Their findings and the findings of 1 additional study75 that did not include a control group and could therefore not be included in the meta-analysis are summarized in Table 5. No case-control study investigated the influence of family history of stroke on stroke severity or recovery.
|Study||Jerrard-Dunne et al (2003)*†49||Polychronopoulos et al (2002)§59||Meschia et al (2001)75|
|Proportion of patients with a family history of stroke (FHx) are given for each ischemic stroke subtype together with adjusted odds ratios (OR) for having a family history of stroke compared with controls.|
|*Adjusted for age, sex, hypertension, smoking, diabetes mellitus, and cholesterol.|
|§Adjusted for age, sex, hypertension, smoking, and diabetes mellitus.|
|†FHxstroke ≤65 years.|
|% with FHx||18.3% (44/240)||49% (64/130)||42% (32/77)|
|Adjusted OR||OR 1.67 (1.08–2.66) P<0.05||OR 2.05 (1.24–3.38) P=0.005||No control data|
|% with FHx||16.2% (36/222)||50% (47/94)||48% (32/67)|
|Adjusted OR||OR 1.49 (0.94–2.37) NS||OR 2.76 (1.55–4.91) P=0.0006||No control data|
|% with FHx||6.3% (7/111)||40% (28/70)||45% (25/55)|
|Adjusted OR||OR 0.60 (0.26–1.39) NS||OR 1.35 (0.73–2.52) P=0.34||No control data|
|% with FHx||11.3% (32/283)||51% (29/57)||51% (53/103)|
|Adjusted OR||OR 1.11 (0.70–1.77) NS||OR 1.71 (0.85–3.42) P=0.13||No control data|
|% with FHx||14.5% (137/944)||48% (168/351)||47% (145/310)|
|Adjusted OR||OR 1.38 (1.01–1.90) P<0.05||OR 2.06 (1.39–3.04) P=0.0003||No control data|
Characteristics of the cohort studies included in our meta-analysis are summarized in Table 6. One study21 only reported ischemic stroke outcomes, and 1 study examined a cohort of TIA patients.14 None of the remainder distinguished between hemorrhagic and ischemic strokes. All but 3 studies13,14,67 defined which relatives were considered in the family history (FDR, mainly parents).
|Study||Setting||Country||Age, y||Follow-Up, y||FHx||Characteristics of Affected Relatives Considered||Relatives||Stroke Differentiated|
|TIA indicates transient ischemic attack; sibs, siblings; FHx, family history; N/A, not applicable; FDR, first-degree relatives; SD, standard deviation; MI, myocardial infarction.|
|Morrison et al (2000)21||Population (probability sample)||USA||45–64||6||Stroke||No||No||Parents||Ischemic|
|Berger et al (1998)13||Occupational cohort||Germany||51.7 (30–65)||7.2||Stroke||No||No||Not defined||Not differentiated|
|Jousilahti et al (1997)16||WHO MONICA cohort||Finland||44 (25–64)||7;12||Stroke||<60||Yes||Parents||Not differentiated for FHxstroke|
|Kobayashi et al (1997)67||Health screen volunteers||Japan||57.5±9.2||1–6||Stroke||No||No||Not defined||Not differentiated for FHxstroke|
|Wannamethee et al (1996)26||General practice cohort||UK||40–59 at baseline||14.8||Fatal stroke, fatal MI||Yes||Yes||Parents||Not differentiated|
|Lindenstrøm et al (1993)19 (cf Boysen et al 65)||Population cohort||Denmark||>35 at baseline||12||Stroke||No||No||Parents||Not differentiated|
|Brass and Shaker (1991)14||TIA patients||USA||69 (14–99)||1–4||Stroke, MI,||No||No||Not defined||Ischemic|
|Welin et al (1987)27||Birth cohort||Sweden||54 at baseline||18.5||Fatal stroke||Yes||No||Parents||Not differentiated|
|Khaw and Barrett-Connor (1986)17||Population||USA||50–79 at baseline||9||Stroke||No||No||FDR||Not differentiated|
Individuals with a positive family history of stroke had a slightly higher risk of subsequent stroke than those without a family history (OR, 1.30; 95% CI, 1.2 to 1.5; P<0.00001; 1906 stroke outcomes). However, there was major heterogeneity between studies (P=0.0001; Figure 5). The funnel plot of the OR versus the variance was asymmetrical (Figure 3), and there was borderline significant evidence of potential publication bias (P=0.10). However, those studies cited in the recent reviews4–6 reported significantly (P=0.00007) higher odds (OR, 1.76; 95% CI, 1.5 to 2.1; P<0.00001; heterogeneity P=0.18) than studies that were not cited (OR, 1.12; 95% CI, 1.0 to 1.3; P=0.16; heterogeneity P=0.05). There was also possible reporting bias in that the majority of cohort studies that we could not include in the formal meta-analysis because they gave no detailed numbers failed to find significant associations of family history of stroke with stroke risk.15,20,22–24,28,66
Two studies examined the influence of family history of stroke on stroke severity and found a greater influence on less severe strokes,21,26 but no study investigated an association between family history of stroke and stroke recovery.
Five studies adjusted at least partially for potential confounding by established stroke risk factors in the subjects. One study did not report the effect of adjustment.17 The association was attenuated after adjustment in 2 studies21,26 but did not show any major change in the other 2 studies16,27 (Table 4). Two studies also adjusted for potential environmental confounding other than smoking. One of these found that the risk conferred by a positive family history of stroke was independent of the socioeconomic status,16 and the other26 found that patients with a family history of stroke were more likely to be manual workers but found no interaction between family history of stroke and heavy alcohol drinking or physical activity.
Three studies found a significant interaction between family history of stroke and presence of hypertension in the studied probands.16,17,26 One study also found a positive association between family history of stroke and increased body mass index and cholesterol levels but failed to find an association with smoking or diabetes mellitus.26
Influence of Age
Four case-control and 3 prospective cohort studies reported a family history of stroke in relatives at young age. Jerrard-Dunne et al49 and Fonte et al39 provided details about stroke onset in FDR before the age of 65 years, Margaglione et al53 reported about stroke or MI in male FDR aged <55 years and female FDR aged <60 years, and Diaz et al37 reported family history of stroke onset in siblings aged <70 years. The combined OR for these 4 studies was nonsignificantly (P=0.35) higher (OR, 1.82; 95% CI, 1.4 to 2.3; P<0.00001; heterogeneity P=0.78; 1311 patients; Figure 6) than the remainder (OR, 1.61; 95% CI, 1.5 to 1.7; P<0.00001; heterogeneity P<0.00001). Both cohort studies26,27 that stratified their analyses by whether parental stroke occurred before or after age 70 years found a larger effect on stroke risk for parental stroke before 70 years. It was not possible, however, to perform a meta-analysis on these studies because they did not report absolute numbers for each stratum separately. A third study that only considered family history of stroke in parents aged <60 years16 found an overall OR for stroke of 1.89 in men and 1.80 in women.
Five case-control studies41,48,50,63,64 recruited only patients aged <70 years, 1 study recruited patients with stroke onset at <65 years,60 1 study recruited patients with stroke onset at <50 years,55 and 1 study reported family histories separately for patients aged <54 and ≥54 years.68 The combined OR for these younger patient groups was nonsignificantly (P=0.10) higher (OR, 1.93; 95% CI, 1.7 to 2.2; P<0.00001; heterogeneity P=0.0001; 1240 patients; Figure 6) than the remainder (OR, 1.69; 95% CI, 1.6 to 1.8; P<0.00001; heterogeneity P=0.00001).
Two prospective cohort studies14,17 reported the influence of family history of stroke on the subgroup of individuals who had their strokes when aged <65 or 70 years, respectively, during follow-up. They found a reduced effect in younger probands (OR, 0.41; 95% CI, 0.2 to 0.8; P=0.29; heterogeneity P=0.85; 38 stroke outcomes); however, the number of strokes was very small. In addition, Wannamethee et al26 reported that the increased risk of stroke with parental death from stroke was apparent in older men (aged 50 to 59 years at baseline) but not in younger men (aged 40 to 49 years at baseline). In contrast, Jousilahti et al16 found a higher relative risk of family history of stroke for both men and women aged <50 years compared with older individuals.
Influence of Number of Affected Relatives
Two case-control studies46,64 and 1 prospective cohort study26 reported the effect if both parents had been affected by stroke26,64 or vascular disease.46 The combined OR for a family history of stroke in the case-control studies increased significantly (P=0.018) when they were analyzed separately for a family history in only 1 parent (OR, 1.08; 95% CI, 0.7 to 1.6; P=0.77; heterogeneity P=0.35) versus a family history of both parents (OR, 2.45; 95% CI, 1.4 to 4.2; P=0.002; heterogeneity P=0.18). The number of subjects with 2 affected parents in the cohort study was too small to allow meaningful analysis.
Test for Heterogeneity
Cohort studies (P=0.069), higher maximum age of patients (P=0.071), and higher maximum age of relatives (P=0.059) were associated with a lower OR on univariate analysis of potential sources of heterogeneity. These parameters still had a significant inverse association if tested together in a multivariate model (type of study, P=0.032; maximum age of relatives, P=0.021; maximum age of patients, P=0.018) and accounted for approximately 34% of the total heterogeneity between studies (adjusted R2=0.344).
There was highly significant (P<0.0001; Figure 3) evidence of possible publication bias if both case-control and cohort studies were tested together. When stratified into tertiles according to the inverse of the standard error (precision) of their individual estimates, the most power-ful (largest) studies found significantly (P<0.00001) smaller ORs (OR, 1.48; 95% CI, 1.4 to 1.6; P<0.00001; heterogeneity P<0.00001) than studies of intermediate power (OR, 1.82; 95% CI, 1.6 to 2.0; P<0.00001; heterogeneity P=0.005) and low power (OR, 2.30; 95% CI, 1.9 to 2.8; P<0.00001; heterogeneity P=0.24; Figure 4).
A number of epidemiological studies of twins, affected sibling pairs, and family history data have suggested that there are significant genetic influences in stroke.¶ Previous nonsystematic reviews of genetic epidemiology of stroke had identified 11 to 16 family history studies and 1 twin study. Ours is the first systematic review, and we identified 50 independent family history studies and 3 independent twin studies.
Twin studies provide the most reliable evidence of genetic influence in complex diseases, and they are best suited to disentangle genetic influences from influences of a shared environment. However, there are difficulties in conducting twin studies in stroke patients. Stroke most commonly affects old people, which makes it challenging to recruit enough twin pairs and increases the chance of twins dying of other unrelated diseases. Thus far, only 3 twin studies have been reported, and they reached different conclusions. Our meta-analysis showed a small genetic influence on the risk of stroke, with monozygotic twins being 1.6 times more likely to be concordant for stroke than dizygotic twins. This should be compared with a 3- to 5-fold better correlation of blood pressure among monozygotic twins versus dizygotic twins70,71 and a 10-fold higher concordance of monozygotic versus dizygotic twins in studies of multiple sclerosis.72 The most frequently quoted twin study of stroke, which used a mailed questionnaire for case ascertainment, found a 17.7% concordance in monozygotic twins versus 3.6% in dizygotic twins.10 This higher concordance might partially be explained by the younger age of the twins in this study compared with the other twin studies, which mainly explored concordance of fatal stroke, suggesting an increase in the relative influence of hereditary factors on stroke risk in younger patients. However, this initial study was based on only 8 stroke-concordant twin pairs, and on 10-year follow-up this difference had diminished to 12.8% and 8.0%, respectively, which is in good agreement with the other 2 studies. Interestingly, in a study on heritability of coronary heart disease among Swedish twins, Marenberg et al73 found a very strong association of age and genetic liability with a strong genetic component among young twins and a weak component among old twins. None of the twin studies of stroke differentiated between stroke subtypes or collected information on other risk factors to assess potential confounders.
We also found that a family history of stroke was only a moderate risk factor for stroke in case-control and cohort studies. Case-control studies found a higher risk compared with prospective cohort studies, but the results of the highest quality case-control studies were very similar to the findings of the prospective cohort studies (OR, 1.28; 95% CI, 1.1 to 1.5 versus OR, 1.30; 95% CI, 1.2 to 1.5). Prospective cohort studies eliminate recall bias, which is one of the most important biases in family history studies, and they are therefore more likely to determine the true underlying effect. Nevertheless, there was major heterogeneity between studies within both the case-control and the prospective cohort groups.
There was evidence of possible publication bias in both case-control and prospective cohort studies, with larger studies reporting more conservative estimates than small studies. There was also evidence of reporting bias. However, it is likely that there was true heterogeneity between studies. First, there is some evidence that genetic factors are less important in strokes that occur later in life. Studies that only considered a family history of stroke to be positive if the affected relatives were younger than a certain age16,39,53 found a stronger association than the pooled estimate. Moreover, studies that stratified their analysis by the age at which strokes occurred in relatives found a stronger effect of family history of stroke in the group of relatives affected at a younger age.16,25–27,37,49 There is more uncertainty in studies investigating young stroke patients. Studies investigating only young and middle-aged patients with stroke41,48,50,55,60,63,64 found a greater influence of family history of stroke compared with studies including older patients, but there was disagreement between the studies that stratified their patients by age and reported on the effect of family history of stroke in different age strata.
Second, not all stroke types are likely to be equally heritable, and case mix is therefore likely to influence the results. For instance, subarachnoid hemorrhage, which was included in some of the studies, might have a different genetic component compared with other strokes.74,50 Many studies did not differentiate between ischemic and hemorrhagic stroke, and only 2 studies49,59 subtyped ischemic stroke sufficiently to compare a family history between subtypes. Large- and small-vessel disease was more strongly associated with a positive family history of stroke than the remainder of ischemic strokes. One additional study75 gave details of a family history of stroke in carefully subtyped stroke patients but did not include a control group. The exact stroke phenotype among affected family members is very difficult to ascertain, and this could mask stronger associations of less frequent stroke subtypes.
Third, few studies investigated the potential influence of the number of affected relatives or controlled for family size. The impact of a family history of stroke increased with the number of affected FDR in the studies25,46,64 that reported on this. Few studies corrected their findings for this effect.
Fourth, the influence of family history on the risk of stroke might vary in different populations and ethnic groups and may also vary over time as environmental factors, dietary habits, and levels of deprivation change. A considerable degree of the heritability of stroke appears to be conferred by the heritability of risk factors and intermediate phenotypes (see below). The introduction of preventive treatments such as antihypertensive drugs, statins, and carotid endarterectomy could have attenuated the penetrance of stroke. Interestingly, the strongest associations were found in studies from the 1960s,41,46 mainland China,51,58 and Russia.48
Fifth, genetic factors could have an influence on stroke severity. Only 2 studies21,26 investigated the influence of family history of stroke on stroke severity. Both found an increased influence on less severe strokes. However, there was no consistency in the results between studies that only considered a family history of fatal stroke. Four studies26,27,41,46 found a positive association with family history of fatal stroke, whereas 4 others18,23,54,62 failed to find any association. Reed et al69 found a positive correlation between a family history of stroke and MRI stroke volume on purely imaging-defined strokes in probands.
Finally, several other methodological issues could potentially affect the strength of association between family history of stroke and risk of stroke. The choice of control groups (eg, whether they included patients with other vascular diseases) is likely to affect the magnitude of association. In addition, the ascertainment of family history is difficult, and different methodologies to collect family history are prone in varying degrees to bias and inaccuracies.76,77
Most established risk factors and intermediate phenotypes for stroke, such as hypertension, diabetes mellitus, ischemic heart disease, hypercholesterolemia, smoking, and carotid stenosis, are likely to run in families.78–85 A family history of stroke was associated with a higher proportion of hypertension# and other conventional risk factors in patients, including smoking, and most studies found that adjustment for vascular risk factors diminished the associations between family history and stroke. In addition, Nicolaou et al86 found a higher occurrence of hypertension and stroke in parents of hypertensive probands compared with controls; Williams et al1 found that positive family histories of ischemic heart disease, stroke, hypertension, and diabetes mellitus were significantly associated with each other; and finally, Lestro-Henriques et al87 found that a family history of cardiac or cerebrovascular disease was more common in hypertensive stroke patients than in normotensive stroke patients. Furthermore, in 2 other studies, a family history of stroke was significantly related to a cluster of vascular risk factors.37,42 We were unable, however, to precisely estimate the contribution of the heritability of risk factors to the heritability of stroke. Most studies reported that heritability of stroke was independent of hypertension (or at least independent of a history of hypertension or a single measurement of blood pressure).16,17,26,27,49,52,58,59 The collection of family histories of risk factors poses further methodological difficulties since hypertension, type II diabetes mellitus, and hypercholesterolemia were significantly underdiagnosed in the past, and diagnostic criteria have changed. The contribution of these risk factors to the heritability of stroke may therefore be greater than previously thought.
We were not able to assess the role of environmental factors, eg, smoking, eating, and drinking habits or social status in a positive family history of stroke and to what degree they modulate genetic influences.
We identified 53 independent studies of the genetic epidemiology of ischemic stroke. Taken together, these studies suggested a small genetic contribution to stroke occurrence, but there was major heterogeneity between studies, with much stronger associations in small studies and methodologically less rigorous studies. Moreover, studies that reported insufficient data to allow meta-analysis tended to have found weaker associations. It is possible that the genetic contribution is larger for certain subtypes of stroke. However, only 2 studies considered stroke phenotype in detail, and many studies did not differentiate between ischemic and hemorrhagic stroke in the proband. There was some evidence that the genetic contribution is greater for stroke occurring at a younger age, with stronger associations in analyses confined to probands or relatives aged <70 years. However, few studies considered the number of affected and unaffected relatives, and only a minority of studies adjusted associations for intermediate phenotypes. There were few data on the influence of family history on stroke severity and no data on stroke recovery. More detailed large-scale genetic epidemiological studies are required.
This study was supported by the United Kingdom Medical Research Council (Drs Floßmann and Rothwell) and the Wellcome Trust (Dr Schulz). We would like to thank Robert Cuffe for advice on statistical analysis. We are very grateful to Professor Hugh S. Markus for providing us with the raw numbers from his study of family history in subtypes of ischemic stroke.49
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