If mobile phone use causes brain tumors, the marked increase in prevalence of use over a 20-year period will eventually influence the time trends of the incidence rates of these tumors. The populations of the Nordic countries were among the first to adopt mobile phones extensively, and since 2005 have had more than one subscription per inhabitant.1–3 In these countries, high-quality nationwide population-based cancer registries have been maintained for the past half century,4 with mandatory and effective recording of all incident tumors including benign brain tumors. Hence, these registries provide excellent opportunities for surveillance of the occurrence of brain tumors. Glioma constitutes the most likely tumor type to show a risk increase associated with mobile phone use, according to some reviews5–7 and the monograph meeting of the International Agency for Research on Cancer.8
Up to 2003, time trends in the incidence of glioma among adults in the Nordic countries did not indicate a detectable increase paralleling the increasing prevalence of mobile phone use.9 Time trends in malignant brain tumors were similarly flat in the United States up to 200610 and 2007,11 and in the United Kingdom up to 2007.12 Before these surveillance studies had been reported, the association between mobile phone use and brain tumor was investigated in cohort13,14 and case-control studies.15–21 One Swedish case-control study reported markedly elevated risks of malignant brain tumors (odds ratios [ORs] = 2–4) associated with overall use of mobile phones,20 and increased risks were also found in the analyses22–24 pooling this dataset with a previous study of the same group; however, most other studies showed no consistent excess risks.16–18,25 The largest study was the international Interphone case-control study, in which the odds ratio for glioma was decreased for overall use but elevated for heavy users of mobile phones (the 10% of regular users who had talked on their phone,25 a lifetime total of ≥1640 hours).
Mathematically, the OR reported in case-control studies of the association between mobile phone use and cancer should correspond to the ratio of the incidence rate in the subgroups of the population using mobile phones to the incidence rate in the subgroups not using mobile phones, assuming a true effect and no change in other factors affecting the disease. Indeed, an elevated OR for a specific exposure group in a case-control study should correspond to an increase in incidence rates of similar magnitude in the population with similar exposure.
It has been argued that the proportion of the population using mobile phones in the Nordic countries in 2003 was still too small to detect a postulated effect of mobile phone use at population level,26 although a RR of about 2.0 after 10 years of use would have already increased the IR by about 20% between 1999 and 2003 among the 40–59-year-old men.27 It remains, however, relevant to evaluate how large risks and how short induction periods yield detectable increases in incidence rates; or conversely, if no increases in secular trends are detectable, what levels of risk and induction periods can be excluded. We conducted a simulation study to address this.
To update the previous publication9 with data from 2004 to 2008, we analyzed time trends in the incidence rates of glioma among adults aged 20 to 79 years of the Nordic countries from 1979 to 2008. In addition, assuming that the use of mobile phones caused glioma, we simulated a population similar to 40–59-year-old Nordic men, and computed the probability of detecting the postulated association. Our aim was to define which levels of risks and induction periods would be compatible with the observed time trends in this population, and which would not be compatible.
We obtained the numbers of primary gliomas (International Classification of Disease for Oncology version 3, topographical codes C71 and morphologic codes between 938 and 948, and similar codes in the International Classification of Disease version 7 for the early period) in patients aged 20 to 79 years at diagnosis during 1979–2008, from the national cancer registries of Denmark, Finland, Norway, and Sweden. Sizes of the populations at risk by 5-year age groups were acquired from the national population registers for each calendar year.
Statistical Analysis of Observed Incidence Rates
Annual age-standardized incidence rates of gliomas per 100,000 person-years were calculated separately for men and women, standardized to the European standard population. The combined age range (20–79 years) and the 20-year age groups 20–39, 40–59, and 60–79 years were considered; country-specific and combined data were analyzed.
A piecewise log-linear model called joinpoint analysis (Joinpoint Regression Program, version 3.4.3 – April 2010; Statistical Research and Applications Branch, National Cancer Institute) without constraints on the positions of the nodes or joinpoints was used to identify trend changes and to estimate annual percent change in incidence rates over the period 1979–2008. The model was specified to include a maximum of 3 joinpoints, which could occur in the middle of a year or between 2 consecutive years. The model constrained the joinpoints to be at least a year and a half apart, and at least 2 years from the start and end of the study period. Under the assumption of heteroscedasticity and uncorrelated errors, the best-fitting model was searched on 4499 randomly permuted datasets using the grid search method. Tests, at an overall two-sided significance level of 0.05, were not adjusted for autocorrelation.28
We generated datasets under the hypothesis that there is a risk related to mobile phone use, assuming various risk scenarios (eg, a 2-fold increased risk 10 years after first use of mobile phones). In these simulated datasets, the association between mobile phones and brain tumors shows up as increased number of cases with a random distribution around the expected value of excess cases. Next, we estimated the risk parameter and its 95% confidence interval (CI) from the dataset and the model, pretending the true value was unknown. The risk estimated for each dataset was distributed around the value used in the scenario, and its 95% CI indicated whether the increase was statistically significant. We considered the proportion of datasets with statistically significant increases as the measure of our ability to detect an increased risk caused by mobile phones in the observed incidence rates, for each simulated scenario.
Prevalence of Use of Mobile Phones Used in Simulation Study
Information on use of mobile phones was obtained from self-reports in a sample of the general population of the Nordic countries interviewed for the Interphone study (ie, controls).21 We abstracted the proportion of persons who used mobile phones regularly—defined as using a mobile phone on average at least once a week over a period of at least 6 months–and the proportion of heavy users, at least 1640 hours of lifetime cumulative call time; these prevalences were computed for each year from 1980 to 2002 by sex and by age group, with lags of 1, 5, 10, and 15 years accounting for aging during the lag period. We focused our simulation study on men aged 40 to 59 years who had the highest use of mobile phones between 1980 and 1993.27 Data collection for the Interphone study stopped in 2002–2003. For the period 2003–2008, we extrapolated use in 2002 as rising 3% annually, a slightly slower increase than the average observed over years 1990–2000. To account for possible selection bias (ie, lower participation in mobile phone studies among controls not using phones29), lower prevalences were used in a sensitivity analysis under conservative assumptions.
Generation of Datasets
We sampled the number of cases as a Poisson distribution in a population of the same size as the population of men aged 40 to 59 years of the Nordic countries (from 2,461,722 persons in 1979 to 3,439,372 persons in 2008) with the same prevalence of use, with baseline incidence rate for nonexposed subjects of 8.88 cases per 100,000 person-years at risk, and with the assumption that the use of mobile phones increased the risk of glioma after an apparent induction period during which no increase occurred. Each scenario combined a value of RR (0.8, 1.1, 1.2, 1.5, and 2.0) that multiplied the baseline incidence rate for all mobile phone users or for heavy users only, with an induction period (1, 5, 10, and 15 years).
The following equation describes our model:
where i denotes the calendar year (1979–2008); J, the presumed induction period (1, 5, 10, or 15 years); O, the observed number of cases; b, the baseline incidence rate; RR, the relative risk; E, the number of persons exposed (all users or heavy users); and N, the number of persons unexposed (the remainder of the population). We generated 10,000 datasets per scenario.
Estimation of Outcome
For each simulated dataset, we estimated the relative risk (RR) under the Poisson model (1), with known exposed and unexposed populations (E and N) and unknown RR, and baseline incidence rate (b), treated as a nuisance parameter. Because the likelihood of model (1) was not standard, we obtained the estimates of the parameters with a maximum likelihood procedure (ml function of Stata; StataCorp. 2007. Stata Statistical Software: release 10. College Station, TX: StataCorp LP). Of the 10,000 estimates, the proportion of lower bounds of the 95% CI above 1.0 was a bootstrap estimate of the probability of obtaining a significantly increased RR, conditional on the scenario. For comparison purposes, we fitted the same model (1) on the observed number of cases among men aged 40–59 years.
Glioma Incidence Trends
This study was based on 35,250 glioma cases diagnosed from 1979 to 2008; the annual average number of glioma was 1175 in a population of 17 million adults aged 20–79 years. Sweden accounted for approximately 40% of the population and of the cases; the remaining countries (Denmark, Finland, and Norway) each contributed roughly 20% of the population and cases. The age-standardized incidence rates were higher in men (8.6 per 100,000 person-years) than women (6.0 per 100,000 person-years), and increased with age (Fig.). All countries had similar rates (eTable 1, http://links.lww.com/EDE/A543).
Time trends did not strongly increase at any point during the period 1979 to 2008 in any country (data not shown), among men, women, or any age-group (Table 1, Figure). Incidence rates were generally stable over the whole period, and increased gradually among older persons. A slight decrease in incidence rates was observed after the late 1980s among the younger men overall (annual percent change = –0.7% [95% CI = –1.4% to 0.1%]) and in Denmark and Sweden, but not in Finland and Norway (data not shown).
Prevalence of Regular Use
The self-reported prevalence of regular mobile phone use among men 40–59 years of age ranged from 2% in 1980 (which was the earliest year on which data were collected in the Interphone questionnaire) to 79% in 2002; we extrapolated this to 98% in 2008. The prevalence of heavy users was 0.2% in 1988 (the first year above 0%) to 16% in 2002 and extrapolated to 33% in 2008.
These exposure prevalences, combined with the induction periods of 1, 5, 10, and 15 years, did not result in statistically significant increased risks when model (1) was fitted on the observed number of glioma cases. The RRs were below 1.09 for all induction periods, for both all users and heavy users, with the 95% CIs consistently including 1.0 (data not shown).
Scenarios with shorter induction periods combined with higher RRs resulted in higher probabilities of detecting significant increases in estimated RRs (Table 2). For example, for the scenario in which all users of mobile phones were assumed to have 50% increased RR of glioma (RR = 1.5), 15 years after using their mobile phone for the first time (Table 2, row 2, column 6), 84% of the simulated datasets demonstrated a statistically significantly increased estimated relative risk. The remaining 16% had either nonsignificant or no increased risk for this scenario. Thus, if there was a true underlying risk increase 15 years after first mobile phone use, it would most likely be seen in the incidence rates (84%). When there was a doubling in risk among heavy users only, all datasets (100%) demonstrated a statistically significant effect of mobile phones with induction period of up to 5 years; this proportion decreased to 69% if induction was 10 years, and to 7% if induction time was 15 years (Table 2, row 5).
In scenarios with exposure defined as regular mobile phone use, statistically significantly increased estimated relative risks were obtained in all datasets for a risk of 2.0 and an induction period of maximum 15 years, for a risk of 1.5 and an induction period of maximum 10 years, and for a risk of 1.2 and an induction period of maximum 5 years. In scenarios where only heavy users were assumed to be at risk, statistically significantly increased risks were obtained in all or almost all the datasets for a risk of 2.0 and an induction period of maximum 5 years. All other combinations could be masked by the random fluctuations inherent in the occurrence of a rare disease in the population of middle-aged Nordic men. In particular, for a postulated risk of 1.5 among heavy users and an induction period of 1 year, statistically significantly increased risks were obtained in 98% of the datasets; thus, the probability of a true causal relative risk of 1.5 for the heaviest users of mobile phones, given the observed absence of an increase in incidence time trends, was estimated to be approximately 2%. When a true relative risk of 0.8 was modeled on all users, all datasets demonstrated significantly decreased risks for induction periods of up to 5 years. Essentially similar results were obtained in the sensitivity analyses under conservative assumptions on levels of use (eTable 2, http://links.lww.com/EDE/A544).
We detected no upward turn in the time trends of glioma incidence rates in the Nordic countries during 1979–2008, overall or in any subgroup by country, age, or sex among adults. Our results extend those of a previous study,9 of time trends in the Nordic countries up to 2003 by adding 5 more calendar years of follow-up. These analyses are based on the entire adult population of Denmark, Finland, Norway, and Sweden (17 million people) and are strengthened by the comprehensive high-quality cancer registration in these countries.30
If mobile phone use causes brain tumors, the change in prevalence of use from zero to nearly 100% over a 20-year period would eventually influence the incidence rates of these tumors. Conversely, a lack of change in the incidence time trends, at any point in time, would constitute evidence against this association. Given that the incidence time trends were stable up to 2008, several risk scenarios could be discarded with confidence, based on our simulations. The risk (should one exist) would have to be lower, or occur after a longer induction period, or act on a smaller population, than the scenarios that we discarded.
Several case-control and cohort studies have evaluated the association between mobile phone use and glioma risk in adults, covering cases ascertained up to 2004. The results of our simulation study are compatible with studies showing no increased risks of gliomas.16,17,31 In the Interphone study,25 the OR for glioma related to mobile phone use overall was statistically significantly decreased among regular users (OR = 0.8 [95% CI = 0.7 to 0.9]). This OR is also incompatible with the observed time trends and therefore, appears due to biases and errors.29,32–34 An elevated risk (1.4 [1.0 to 1.9]) among the 10% heaviest regular users was also reported, although bias was discussed as an alternative explanation. We evaluated the probability of a true OR of this magnitude (1.5) as approximately 2%, given the observed incidence trend. Therefore either the observed time trends reflect the lack of a moderate effect on the occurrence of glioma among heavy users, or these results are within the 2% margin of error that was the estimated probability for failing to detect such an effect in the current setting.
The association between mobile phone use and malignant brain tumors in adults has also been evaluated in 3 Swedish case-control studies.15,19,20 The results of the first 2 studies were compatible with our simulations, as no statistically significant increased risks were reported for overall use of mobile phones15,19 (OR of malignant tumors for use of analog phone = 1.1 [95% CI = 0.8 to 1.6]; OR for use of digital phones = 1.1 [0.9 to 1.5]). The third study, with case ascertainment in 2000–2003, yielded markedly elevated risks: OR for all analog phone use = 2.6 (1.5 to 4.3) and OR for all digital phone use = 1.9 (1.3 to 2.7), with elevated risks for all latency periods, but with even higher OR for a latency of >10 years.20 We found that such highly increased risks are strongly inconsistent with observed time trends in risk in the Nordic countries up to 2008. This may indicate the presence of biases and errors in self-reported mobile phone use, resulting in unrealistically high ORs. In the Interphone study, analyses restricted to the subpopulation who used mobile phones regularly were performed as a way to correct for participation bias. In these, the risk of glioma was increased for both short- and long-term use (OR for use 2–4 years = 1.7 [1.2 to 2.4]; OR for use 5–9 years = 1.5 [1.1 to 2.2]; OR for use 10 or more years = 2.2 [1.4 to 3.3]).25 Our simulation study indicated that these results were also unrealistically high. Several meta-analyses have been published,35–37 but such pooling of evidence does not remove systematic biases. Incidence rates of brain tumors in Swedish children and adolescents (5–19 years old) have recently been published for the period 1990 to 2008.38 Children and adolescents adopted mobile phone use later than adults, and so follow-up may be too short to observe an effect specific to this age group.
Our simulation study is based on certain assumptions. The induction period relating mobile phone use and glioma risk, if such an association exists, is unknown, as is the magnitude of risk, and the real patterns may be more complex than the scenarios we simulated. In addition, there are several factors that we did not account for. The coverage of the Nordic cancer registries is not complete; 1.5% to 6% of the malignant tumors are missed among men aged 40 to 59 years,39–41 and although cancer registration is thought to have improved, we have no information on possible changes in completeness over time. We assumed that other risk factors of the disease, as well as its detection, had no major impact over the period 1979–2008 in this age group, which is consistent with the observed stable rates. Some increase was present in the older age group, as reported in other studies,10 which could reflect improved imaging techniques and increased diagnostic activity. Our main results are based on the group 40–59 years of age and therefore should not have been affected by this phenomenon. We used self-reported use of mobile phones to evaluate the prevalence of use and heavy use up to 2002, and extrapolated the prevalence for the period 2003–2008 based on documented overall trends. The extrapolation over time affected only the results for the induction period of 1 year. The use of hands-free devices was not accounted for, but this was not frequent in these populations (data not shown). In sensitivity analyses, we investigated conservative assumptions on prevalence of use, with similar results. Indeed, Interphone controls were likely to overestimate prevalence and amount of mobile phone use in the entire population due to selection and information bias.29,32–34
Because we focused our simulation study on men aged 40–59 years, we did not account for the other age and sex subgroups that, independently, also did not show any change in their incidence time trends.
Our data indicate that, so far, no risk associated with mobile phone use has manifested in adult glioma incidence trends, although the induction period, if any, is unknown. Case-control studies conducted in 2000–2003 were not able to evaluate risks for exposures longer than 10–12 years. Our simulations show that many increased or decreased risks reported in case-control studies are implausible, implying that biases and errors in the self-reported use of mobile phone have likely distorted the findings.
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We thank the Cancer Registries of Denmark, Finland, Norway, and Sweden for providing the incidence data, Monika Moissonnier (IARC) for her help in running the simulations, and John Daniel (IARC) for careful editing of the manuscript.