Stroke incidence in the young: evidence from a Norwegian register study

Recent studies have suggested that the incidence of stroke, one of the main contributors to premature death and disability, has declining in the developed world [1‐5]. However, there is concern that this overall trend masks an alarming increase in incidence among the young—often defined as those below the age of 55 [6]. Because stroke incidence has a very strong age gradient, with an exponential growth curve as the individual ages, most strokes originate from the older cohorts. Hence, even a moderate decrease in age-adjusted stroke incidence for the elderly could potentially disguise a significant increase in stroke incidence among the younger segments of the population. As stroke etiology differs between the old and the young, [7] the successful efforts in recent decades to better detect and manage stroke risk factors may not be as effective in the younger segments of the population as in the elderly.

The influential Journal of the American Heart Association recently printed an editorial titled Rising Stroke Incidence in Young Adults: More Epidemiological Evidence, More Questions to Be Answered by Y. Béjot et al. with a call for more research into this phenomenon [8]. This editorial states that ‘[t]here is now a large amount of evidence that ischemic stroke incidence is on the rise in young adults, and the reasons for this trend are probably multiple’[8]. Several possible reasons for a rise in stroke incidence among the younger cohorts in the developed countries are suggested. An increase in classical vascular risk factors, alongside the emergence of new risk factors such as substance abuse, is discussed alongside possible confounders such as an increased awareness and better stroke recognition.

Articles cited in the editorial as evidence for increasing stroke in the young are listed in Table 1, as well as additional relevant studies. Many of the articles cited as supporting an increasing incidence offer alternative explanations for the observed increase, and are substantially less categorical in their conclusions. For example, Kissella et al. highlight that, while their study demonstrated an apparent increase in stroke incidence (including subarachnoid haemorrhage) in the young, they also documented that the fraction of patients for whom an MRI was obtained roughly doubled during the study period (from 37.5 to 70.0% for 20–44 years old, and from 31.0 to 63.4% for 45–54 years old.) Hence, a temporal trend in the health systems’ ability to correctly detect strokes in the younger segments of the population could be a possible confounder.

Another important point in this context is that several of the studies indicating an increase in stroke incidence, only consider hospitalised strokes, which does not imply that stroke incidence is increasing. In particular, if the severity of hospitalised strokes show a trend towards milder strokes, what we could be observing is not an increase in stroke incidence, but rather an increase in the health systems’ ability to detect, diagnose, and treat more of the milder strokes. Assuming an underlying constant stroke incidence, we would expect an increase in the number of diagnosed stroke events given the remarkable development in imaging techniques, and their increased use in clinical practice.

It is also problematic that the category young adults is not uniformly defined, and that methods and classification vary greatly between studies, both with respect to incidence rates and aetiology [20]. Hence, it is hard to judge the evidence presented in the articles in Table 1 collectively. The individual risk of stroke increases exponentially with age [21‐23]. Estimated incidence rates can therefore be quite sensitive to choice of age strata during analyses. For example, if 20–44 years old are grouped together in a regression model, and one aim is to estimate the association between year and stroke incidence, even a moderate shift of the age-stratified composition of the 20–44 age group—i.e., towards ‘more 44 years old and less 20 years old’—could be confounded with a temporal trend towards increased cerebrovascular accidents (CVA) incidence, due to the underlying exponential increase in individual risk for stroke or transient ischemic attack (TIA).

Some contemporary articles point to evidence of either a stationary, or even declining, trend [24, 25]. Notably, to our knowledge, no studies have demonstrated increased stroke-fatality rates in the young. This is true both in the sense that there has been no increase in fatalities attributable to strokes, nor has there been an increase in the case fatality rates—the proportion of registered strokes resulting in death—quite the contrary [15].

While the recent shift in acute stroke treatment—from life-supporting and palliative care to effective and advanced interventions—has been a great success in terms of better outcomes for survivors, very few studies have shown improved survival. Therefore, if stroke incidence is on the rise—within any sub-population—we would expect an accompanying rise in fatalities. In the case of the young, the observed combination of increased incidence and declining mortality could be fully explained by increased sensitivity to milder strokes, whereas assuming an underlying increase in stroke incidence would need to be substantiated with a separate rationale for the lack of corresponding increase in fatality.

Vaartjes et al. found that while stroke incidence appears to be increasing in the Netherlands, stroke fatalities are declining—both in absolute terms (i.e., deaths following a stroke per 100k)—and as a percentage of those with a stroke [15]. The authors note that the second statistic shows the most dramatic decline, and acknowledge that they do not have case mix data: i.e., their finding is consistent with an increase in minor, less severe strokes only. They also remark that TIAs are increasing more than other diagnoses during 1995–2010, and tentatively conclude that minor strokes may be the reason behind the observed overall stroke increase. However, no consideration is given to the possibility that the observed increase may be due to an increased propensity for hospitalising less severe strokes, and not that there are necessarily more of them. For example, in a recent Norwegian study, it was shown that during the six months following a mass media intervention aimed at increasing public awareness of stroke symptoms, the average number of ED arrivals on indication suspected stroke almost doubled [26].

Due to uncertainties regarding current changes in stroke incidence in the young, the aim of this study is to investigate temporal trends in childhood and young stroke incidence and fatality, during recent years, using a large data set of all CVAs obtained from the Norwegian Patient Registry (NPR) and the Cause-of-death Registry (NCDR) in Norway for 6 years, 2010–2015 (received May 2017).

This study provides another piece of the puzzle of recent temporal trends in CVA incidence in children and the young, broken down by TIA, ischemic, haemorrhagic, and undetermined stroke. In this way, we provide a more complete picture than is commonly provided in the literature, and also hope to set a standard for the reporting on CVA sub-types together, so that possible shifts between e.g., TIA and ischemic stroke are not confounded with an increase or decrease. On this note, this is also the first study to consider temporal trends in TIA incidence in the young on a relatively large data set; the only former large study with a separate focus on TIA that we know of considered all ages [27].

The data used in this study is described in greater detail elsewhere [28]. Briefly, we combined data from two sources: the NPR, and the NCDR. NPR collates hospital records, including demographics, ICD-10 codes, and admission times from all hospitals in Norway that treat CVA. NCDR collates corresponding information regarding all deaths in Norway. Our data set therefore contains hospital and fatality records for all persons recorded with at least one stroke diagnosis (ICD-10 codes I61, I63, I64) or TIA (ICD-10 code G45) in either NPR or NCDR between 2010 and 2015; the longest time period for which data from both registers were available and could be linked. This dataset was combined with Statistics Norway records for the total Norwegian population, broken down by year, 1-year age groups, and sex. Recorded stroke episodes for the same individual¹ were merged when on adjacent dates. This was done separately by sub-diagnosis (ischemic, haemorrhagic, undetermined and TIA), and jointly for strokes. Subsequent to merging, episodes were assigned to year by their start date. As such the data set does contains all events of hospitalised stroke, in addition to all events where a stroke lead to a fatality. We have little reason to believe that there are many non-fatal strokes that do not lead to hospitalisation in particular not among children or the young. Since TIAs do not lead to deaths by definition, our counts of TIAs is for hospitalised TIAs. However, the threshold for admitting TIAs are low (a recent study showed up to 40% of admissions to a large stroke unit in Norway were stroke mimics [29]) and mandated by national guidelines. This means that we believe our counts are fairly complete, and that only a negligible share of the CVAs experienced by the 0–54 cohorts will not be represented in our data set.

The following ICD-10 diagnosis codes were used for case ascertainment: I63 (ischemic stroke), I61 (haemorrhagic stroke), I64 (undetermined stroke) and G45 (TIA).² Strokes denote the aggregated category ischemic, haemorrhagic, and undetermined, but when computing the counts, events that occurred within 7 days of each other was collapsed onto one stroke event to avoid double counting. Cerebrovascular accidents (CVAs) denotes strokes and TIAs combined. We believe it is prudent to perform all analyses for all of these categories where the data permits; in particular when researching the question about possible temporal trends, since diagnostics, stroke awareness, and hospitals may vary in their practice and ability to distinguish between these categories.

The main outcome of interest was CVA incidence per 100,000 person year at risk (PY), pooled over all CVAs and for the specific CVA sub-types. Furthermore, we analysed mortality trends by two statistics: stroke mortality—here defined as total number of 30-day case fatalities per 100,000 PY, and, case fatality rates. The former is the number of recorded deaths within 30 days of a recorded stroke (including non-hospitalised cases). The case fatality rate is the fraction of the strokes that led to death within 30 days, and as such is relative to the incidence statistic. All three outcomes convey different information, are sensitive to different confounders (actual CVA incidence, case mix, health systems sensitivity for CVA and their ability to distinguish them, treatment options), and may exhibit different temporal trends.

In this study, we are investigating temporal trends in the age and sex-adjusted CVA incidence rates in the young. To answer this question, several concerns must be addressed: first, the definition of ‘the young’. We have considered both childhood stroke—defined as stroke in individuals aged 0–18—and stroke in the young—defined as individuals aged 15–54. This delineation was chosen because it corresponds well with most of the classifications from the literature on stroke in the young.³

Since we are modelling count data it is reasonable to consider the negative binomial distribution; i.e., assuming that \(C_{y,a,s}\sim \mathrm {NB}(\mu _{y,a,s},\theta )\), where \(C_{y,a,s}\) is the number of CVA events of interest for year y, age a, and sex s. This leads to the model specification \(\mu _{y,a,s}=\text {PY}_{y,a,s} e^{\beta _0+\beta \mathrm {Year}+ \beta _a\mathrm {Age}+ \beta _s\mathrm {Sex}}\). That is, the expected number of events is modelled as a function of year, age, and sex, with a factor PY to account for the population size in the bracket. The residuals distributed as a negative binomial with dispersion parameter \(\theta\). The negative binomial model is parameterised so that the variance is \(\mu + \mu ^2/\theta\) and converges to the Poisson distribution as \(\theta\) become large. The main coefficient of interest here is thus the \(\beta\) on the year variable, which should be able to capture temporal trends.

These were fitted both with the counts of stroke events and 30-day stroke fatalities as the dependent variable. This was done for childhood stroke (0–18), and for stroke in the young (15–54), and furthermore, separate models were fitted for each 10-year sub-group 15–24, 25–34, 35–44, 45–54. We fitted models for each of the stroke categories I61, I63, I64, G45, and for the aggregate categories strokes and CVAs. Furthermore, we specified separate models for men and women [30].

We also analysed overall 30-day case fatalities, because this outcome should be sensitive to an increase in underlying incidence rate. Strokes are life-threatening, and under the assumption that the number of stroke fatalities is more readily available than the number of stroke events, it seems reasonable to expect that an increase in stroke incidence would be accompanied by an increase in stroke fatalities. As sensitivity analyses, we also performed Mann–Kendall non-parametric test for temporal trends, [31, 32] and compared the incidence rates in 2010–2011 vs. 2014–2015 by a standard Chi square test for difference in proportion over the same groups. We also computed the relative share of haemorrhagic, ischemic, undetermined, and TIA of all CVAs, stratified by age group and inspecting for temporal trends.

All analyses were performed with the statistical software R in the RStudio environment, [33, 34] relying on the Plotly and Stargazer packages for graphing and rendering regression outputs [35, 36]. The negative binomial regression model was fitted with the MASS-package’s glm.nb-function [37].

We find no evidence for a temporal trend of increased stroke incidence in the young, nor for childhood stroke. Rather, we find weak evidence for a decline in CVAs in the young (15–54), although this trend most likely pertains only to the Strokes in the 45–54 stratum (and TIAs), and appears stationary across the 15–44 group (Table 4 [model 2], Table 5 and Fig. 2). Furthermore, we find some evidence for a decrease in both childhood (0–18) and young stroke mortality, consistent with the literature, [17] while childhood CVA incidence appears stationary. This is mirrored in the stationary case-fatality rates for young adults, and an indication of a drop in childhood case-fatality rates.

In Norway, ten years into the millennium, all facts considered, stroke incidence and mortality appear to be stable for children and the younger cohorts, while the decline in stroke incidence for the middle aged that has been observed for quite some time has still not stabilised at a new incidence-level [38].

Our findings thus do not support recent publications suggesting an increase in stroke incidence in the young, and our findings contrast the findings reported on in the studies summarised in Table 1.

As for the clinical relevance of these estimates, keeping in mind that when \(\beta\) is close to zero, it can approximately be interpreted as a percentage change per year, we see that the main regression model predicts a decline in CVA incidence of about 4% per year, but only in the 45–54 cohort. In Norway, there are about \(7.1\times 100\mathrm {k}\) individuals in this age bracket; from which we currently expect \(67.3\times 7.1=478\) CVA events in a given year. About 10% of these patients will die without admission, [28] meaning that the yearly absolute reduction is given by \((0.9\times 478\times 0.96^y)\times 0.04\) between years y and \(y+1\). The model thus predicts that the number of CVA admissions of patients in the bracket 15–54 will drop from the current 430 to 275 by 2025. It is reasonable to assume—given the ageing of the population, combined with the fact that most CVAs originate with the 55+ cohorts –that stroke units will not see a clinically relevant reduction due to this decline.

Our results are valid for Norway—indeed we have data on every CVA known to Norwegian health authorities. We also believe that our results likely generalise to the Scandinavian countries, and possibly other northern European countries (e.g. the Netherlands). We do not venture to assume anything beyond this. It follows that we believe that cited results (Table 1) on increasing trends may be spurious, and that continued inquiry is warranted. See also Mallick et al.’s study for a further discussion of spurious temporal trends [39]. Crucially, such studies should include all results, and follow a pre-set (preferably pre-registered—here we are failing ourselves) protocol for which age-brackets to analyse and sub-analyses to undertake.

Admittedly, our data does not contain more than six years, and although we believe that future data in our series will not negate our results, we are hoping to confirm these findings as more data become available from the NPR and NCDR.

The incidence and mortality rates we report are in keeping with previously published studies. As we report on all CVAs (excluding sub-arachnoid haemorrhage), and not only hospitalised CVAs, some variation throughout the literature is to be expected. We do not know of any recent studies that are incompatible with our results. E.g. Marini et al. [40] reported incidence rates for first-ever strokes (including sub-arachnoid haemorrhage) – pooled over 29 studies—at 4.2/100k PY for 15–24; 10.7 for 25–34; and 30.7 for 35–44 (males). These are lower than our estimates at, respectively 4.9,  13.2,  and 40.6, but as we include both recurrent strokes and non-hospitalised strokes registered through the NCDR, these numbers square well. The figures for females compare similarly. Putaala et al. report a 95% CI of (8.4–13.0) for ages 15–49 for first-ever ischemic strokes; we calculate⁹ 20.7 for first-ever and recurrent including non-hospitalised cases. Ischemic- and haemorrhagic stroke mortality in children—0.0 and 0.1/100k PY in our study—also compare well with figures of 0.0 and 0.1 from Krishnamurthi et al. [17]. Béjot et al. reports incidences for first-ever strokes by category for 0–55 as 18.1 (ischemic), 2.2 (haemorrhagic) and 0.0 (undetermined); [6] we find 20.7,  5.3, and 0.9—which fits well with the inclusion of recurrent strokes and the NCDR data on non-hospitalised, fatal strokes.

There are several reasons why others may have found rising trends, and which we believe may have been discussed insufficiently. Chief among these reasons is increased stroke awareness; changes in clinical practice, particularly regarding how patients are handled when minor strokes are suspected; and improved diagnostics during the preceding decade. These mechanisms all work towards biasing estimates on trends in an upward direction. Our study is well positioned to untangle some of these effects, but we cannot rule out that our observation of a temporally stationary incidence profile throughout 2010–2015 is really a moderate decrease in stroke incidence. The observed decline in mortality, which includes non-hospitalised CVA events, corroborates this result [41].

Our study shows that children in Norway have never been safer from stroke fatalities than today. Little seem to have changed with regards to incidence and mortality rates since the late 70s. [10, 39] Furthermore, our results suggest that one is never at a lower risk of stroke, ischemic or haemorrhagic, than when roughly four years of age. The observed stationary or declining incidence in stroke incidence comes despite it being shown that improved imaging techniques may have resulted in earlier estimates being biased downwards [42]. Furthermore, although sex was not a significant predictor in childhood stroke, the Sex-coefficients are all negative (indicating lower incidence for females). There is good reason to believe that male sex is a risk-factor for childhood stroke [10].

Regarding childhood stroke, we see that haemorrhagic strokes are relatively more common in children than adults, as expected. Our results replicate the approximate 50%–50% distribution between ischemic- and haemorrhagic reported for childhood stroke [43]. The incidence rates we find for all CVA in children (3.83/100kPY), and the 30-day case fatality rate (0.13/100kPY) suggest that mortality rates have continued the rather steep decline between 1950 and 2000 in the UK reported on by Mallick et al.; they found a drop from about 0.7/100kPY to about 0.5/100kPY [10]. Broderick et al. and Schoenberg et al. reported CVA 30-day case fatality of about 0.5[43, 44].

We have not paid much attention to discussing the \(\varDelta\)-estimates (Tables 5 and 6). Although we find slightly different results by this method, than with the regression modelling, we mainly included this analysis to counter possible critiques for not comparing the sub-periods’ point estimates, since this method has been employed in other studies.

A further strength is the use of data that has not been stratified prior to analysis. In our regression models, age has been entered as a continuous variable. This precludes any bias introduced by aggregating by year, in the presence of shifts of the within bracket distribution. As discussed briefly above, aging of the younger combined with the exponential increase in stroke risk that age confers (from the age of four and onwards; recall Fig. 1a), means that any aggregating across age-brackets is susceptible to introduce bias in a temporal trend due to auto-correlation of time and within-bracket age-shift.

Also, it is a strength of this study that we performed a pre-defined battery of analyses (regression modelling, \(\chi ^2\), Mann–Kendall), and report on all outcomes. Depending on our own biases, we could have reported an increase in TIA in males aged 35–44 (\(p=0.024\)), or a decrease in TIA in males aged 15–24 (\(p=0.050\)). However, when these ‘results’ are shown in Table 5 alongside the other values, it becomes apparent that these are part of a general body of evidence for a stationary, or possibly declining, temporal trend.

The main limitations to our study, despite covering all of Norway, is the sample size; the sub-analyses on 10-year age-groups each contain roughly four million PYs and 50–500 stroke events. Hence, the study could be underpowered to detect an actual temporal trend specific to some of the younger age-brackets investigated. Any such temporal trend, however, cannot be pronounced. Furthermore, our data is limited to the relatively short time frame of 2010–2015.

With respect to 30-day stroke-mortality, however, we maintain that there is weak evidence of a slight decline, and fairly strong evidence against an increase. Furthermore, the second regression model, explicitly including the \(\mathrm {Age}_\ge 45\)-dummy, provided a point estimate of 0.001 for the overall temporal trend (95% CI − 0.025 to 0.027); which is certainly compatible with a stationary temporal trend.

Another limitation is that the data, in particular data from the NCDR, may be inaccurate. By this we mean that death certificates may be written by physicians without personal knowledge about the anamnesis of the deceased.

Furthermore, we do not have stroke sub-type (e.g. TOAST classification), nor do we distinguish between recurrent and first-ever strokes. On the other hand, we do include TIAs, which we believe is important to get the full picture on CVA trends, and to our knowledge this is the first study to present as detailed numbers on TIA incidence in children and the young as we do. Nevertheless, we believe our results comply with Feigin et al.’s recommendations where the data allows, [30] and it is a further strength of our study that we juxtapose incidences for ischemic, haemorrhagic, and TIA for ready comparison.

Norwegian CVA incidence rates appear stationary, with a weakly significant decline for the 45–54 age group. Stroke mortality appears to remain stable or declining, for both sexes, and for all age categories. Case-fatality rates are declining in children, and stationary for the young. These results contrast recent results that have raised a concern about a possible increasing trend in the incidence for the young, while confirming the stationary or declining mortality. From a Norwegian perspective, therefore, it seems there is little reason to fear that the young are experiencing increased exposure to stroke risk.

An important caveat is that although we have access to all CVA-events—hospitalised or leading to death—based on more than 21M PYs, our study is not powered to detect slight temporal trends. Therefore, the continued research into this question is warranted, and, wherever possible, data sets pooled over several countries should be employed. Such research should preferably be pre-registered, and should report on all sub-analyses, and should report on the recommended mid-decade age bands.