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doi: 10.1097/EDE.0b013e31819c7ba8
Cancer: Original Article

Risk of Pituitary Tumors in Cellular Phone Users: A Case-Control Study

Schoemaker, Minouk J.; Swerdlow, Anthony J.

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Author Information

From the Institute of Cancer Research, Sutton, United Kingdom.

Submitted 15 July 2008; accepted 2 September 2008; posted 9 March 2009.

Supported by the European Commission Fifth Framework Program “Quality of Life and Management of Living Resources,” the International Union against Cancer (UICC), and the UK Mobile Telecommunications and Health (MTHR) Programme. The UICC received funds for this study from the Mobile Manufacturers’ Forum and the GSM Association. Provision of funds to the Interphone study investigators via UICC was governed by agreements that guaranteed Interphone's complete scientific independence. The Institute of Cancer Research acknowledges NHS funding to the NIHR Biomedical Research Centre. The views expressed in this publication are those of the authors and not necessarily of the funders.

Correspondence: Minouk J. Schoemaker, Institute of Cancer Research, Section of Epidemiology, Sir Richard Doll Building, Sutton, SM2 5NG Surrey, United Kingdom. E-mail:

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Background: There is public concern and scientific interest regarding a potential effect of cellular phone use on the risk of developing intracranial tumors. Tumors of the pituitary gland have barely been investigated in this context, but are of interest because of their intracranial location.

Methods: We conducted a population-based case-control study between 2001 and 2005 of the risk of developing pituitary tumors in relation to cellular phone use in Southeast England, with 291 cases and 630 controls. Detailed information on cellular phone use was collected by personal interview.

Results: Tumor risk was not associated with cellular phone use overall (adjusted odds ratio = 0.9, 95% confidence interval = 0.7–1.3), and was not appreciably increased 10 or more years after first use (1.0; 0.5–1.9), or after 10 or more years of cumulative use (1.1; 0.5–2.4). Odds ratios were 1.2 (0.7–1.9) for users in the highest quartile of cumulative number of calls and 1.1 (0.7–1.7) in the highest quartile of hours of use. Separate analyses of analog and digital phone use showed no associations with tumor risk.

Conclusions: We found no evidence that the risk of developing pituitary tumors is associated with cellular phone use for the induction time periods and intensities of use observed.

The increasing worldwide use of cellular telephones has generated public concern about exposure to radiofrequency fields as a potential risk factor for cancer. The number of worldwide cellular phone subscriptions was predicted to reach 4 billion at the end of 2008, and is still rising.1 Over the last decade, numerous epidemiologic studies have been published on the association of cancer risk with cellular phone use, largely reporting on risks of brain tumors and other intracranial tumors. Most studies focused on glioma, meningioma, and acoustic neuroma; the balance of evidence from these studies does not suggest an increased risk for the durations and intensities of use investigated so far.2,3 Exposure from cellular phones is greatest to the surface of the head and diminishes with distance penetrated.4,5 However, as there is no established biologic mechanism by which radiofrequency fields affect cancer risk,3,6–10 it is unknown whether different tissues might differ in sensitivity for any putative effect.

Pituitary tumors arise from the pituitary gland and comprise 10–15% of all diagnosed intracranial tumors.11,12 Despite their predominantly benign nature, they are associated with substantial morbidity because they compress surrounding structures and may cause oversecretion of pituitary hormones. Risk factors are largely unknown and cellular phone use as a risk factor for developing these tumors has barely been investigated, despite their intracranial location. A previous study on pituitary tumor risk in relation to cellular phone use in Japan13 showed no evidence for association. This study was relatively small (101 cases) with few subjects reporting long-term phone use.

We have conducted a population-based case-control study of pituitary tumors, including 291 cases and 630 controls, in Southeast England. We report here on the risk of developing pituitary tumors in relation to cellular phone use.

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We conducted a population-based case-control study of risk factors for pituitary tumors in the Thames regions of Southeast England. The study was conducted in parallel with case-control studies of several other types of intracranial tumors, which have previously been reported in pooled analyses with other centers.14–16 For studies of glioma, meningioma, and acoustic neuroma, we contributed data to the Interphone study, a 13-country case-control study of brain tumors in relation to cellular phone use.17 Our local study was extended to include pituitary tumors, a wider age range, and additional risk factors.

To be eligible, patients had to be 18–59 years of age and resident in the study region at the time of pituitary tumor diagnosis (defined according to the International Classification of Diseases revision 1018 as topography C75.1, D35.2, or D44.3) between 1 December 2000 and 28 February 2005. Recruitment started in January 2001 but a 1-month retrospective ascertainment was applied. The date of diagnosis was defined as the date of the first scan that revealed a space-occupying lesion of the pituitary gland or, if not available, the date of pathologic diagnosis. Cases were ascertained directly from neurosurgical centers and oncology units in the hospitals in the study region as well as from the Thames Cancer Registry to enable population-based ascertainment. Ethical approval for the study was obtained from the Southeast Multicentre Research Ethics Committee.

Control subjects were selected from patient lists at general practitioner practices in the study region. To select a control, practices were instructed to find the first person with a surname starting with 2 randomly selected letters (provided by the research team) and to search through the lists until they found a person who was eligible according to the supplied age and sex criteria. Controls were ascertained throughout the study period and selected controls were subject to the same eligibility criteria of age and residence as the cases. Controls were excluded if they reported a past diagnosis of an intracranial tumor. General practitioner lists were considered a good source of population-based controls because approximately 98% of the UK population was registered with a general practitioner in 1992.19 A single control group was recruited for the several case-control studies of intracranial tumors and were frequency matched on the sex, age, and health-authority distribution of the total group of cases.

Ascertained cases and controls were invited by letter to participate in the study. The letter included a reply slip and a stamped envelope to return to the research team. If no reply was received after several weeks, the subject was contacted by telephone, or if that was not possible, was sent a second letter.

Subjects willing to take part were interviewed face-to-face by trained research nurses at the subject's home or another place convenient to the subject, with the exception of 2 controls who were interviewed over the telephone. Written informed consent was obtained from all participants. The interview was conducted using a structured computer-assisted questionnaire, with answers being entered directly into the questionnaire program on a laptop computer. Subjects were asked whether they had “ever” used a cellular phone prior to the interview, and subjects who reported phone use for an average of 1 or more calls a week for at least 6 months were asked for full details of their phone use. Information on models and types of phones was obtained with the aid of photographs of marketed phones. For each phone model, we collected information on the start and end date of use, the average amount of time of use, and the average number of calls per day, week, or month. Duration of use could also be answered as time per call. Most questions allowed for the answer to be given as a range of values, if an exact answer could not be given. If there were substantial changes in use that lasted for more than 6 months, usage information was collected separately for these periods. Data were also collected on the cellular phone network provider and the extent of handsfree use. Information on the patient's age at diagnosis of a pituitary tumor was confirmed at interview. If it was discovered at interview that a tumor had been diagnosed prior to the start of the study (1 December 2000), earlier than was initially apparent from the case notes, the case was excluded from the analyses.

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Statistical Analysis

Any cellular phone use less than 1 year prior to diagnosis was ignored for the analysis because it was unlikely to be etiologically relevant, except that time since first phone use was evaluated up to the diagnosis date. All interviewed controls recruited for the entire set of intracranial tumor cases were used for this analysis to increase statistical power. As controls were not individually matched to cases, we calculated control exposure by deriving a reference date similar to that of the cases’ diagnosis date, based on the year of interview and how far back the subject was asked to recall exposures (described previously14).

We defined regular cellular phone use as having used a cellular phone for at least 6 months during the period more than 1 year prior to diagnosis (or equivalent reference date for controls). We derived lifetime cumulative number of hours of phone use and numbers of calls by summing the calculated number of hours of use and numbers of calls for each usage pattern of each phone model. Where number of calls or time of use was expressed as a range, the midrange value was used in the calculations shown in the Results, but analyses were repeated after using the lower and upper value of the range as a sensitivity check. When, for individual usage periods, subjects expressed time of use as duration per call, this duration was multiplied by the reported number of calls to obtain an estimate of total duration of use. Odds ratios of risk in relation to these cumulative variables were derived with and without adjustment for use of handsfree devices, based on the proportion of time the subject reported such use, as described previously.20

An unconditional logistic regression model was used to obtain odds ratios for risk of developing a pituitary tumor in relation to indices of cellular phone use. We adjusted all odds ratios for sex, 5-year age category at the reference date, reference date, Townsend deprivation score (a socioeconomic index based on the subject's residential postcode21), and geographic area within the study region (5 areas). Reference date was adjusted for in 6-month periods to control for rapid increases in cellular phone use over time and for the fact that cases were, on average, interviewed later than controls. Analyses were conducted for cellular phone use overall and separately by signal type (analog or digital), because analog phones have a higher average power output than digital phones.6 Tests for trend were done on the continuous values rather than on categorized variables. The statistical package Stata was used in all analyses.22

Because exposure to ionizing radiation is a risk factor for pituitary tumors,23 we considered the effect on the results after excluding subjects who reported having had radiotherapy to the head more than 10 years prior to the reference date. We repeated the analyses with a 5-year instead of a 1-year lag time and for men and women separately. Finally, we compared the results with and without adjustment for Townsend score to give some assessment of the importance of residual confounding by socioeconomic status in the main results.

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We ascertained 506 eligible cases with pituitary tumors, of whom 189 did not take part in the study because of refusal (n = 84), no reply to our request (n = 75), death or illness (n = 13), consultant refusal (n = 9), or other reasons (n = 8). We interviewed 317 cases, corresponding to a participation rate of 63%. After interview, 18 cases had to be excluded because a pituitary tumor had been diagnosed before the start of the study, and 8 cases were excluded because there were no controls in the corresponding stratum of reference period. This left 291 cases in the analysis. We ascertained 1464 eligible controls, and interviewed 630 (43%). Among nonparticipants, 364 controls replied that they were not willing to take part and 470 did not reply to the invitation letter.

Participating cases were somewhat more likely to be female than nonparticipating cases (53% vs. 47%), and were slightly older (mean age: 45 vs. 43 years), and more affluent (mean Townsend score: 2.9 vs. 3.5). Among controls, participants were considerably more likely to be female (proportion female: 53% vs. 37%), and older (mean age at ascertainment: 46 vs. 43 years), and more affluent (mean Townsend score: 2.5 vs. 2.9).

Among participants, the same proportions of cases and controls were women (Table 1). Controls were more affluent than cases (mean Townsend score: 2.5 vs. 2.9), but were similar with regard to age (mean age at the reference date: 46 vs. 45 years) and marital status. All cases had pituitary adenoma.

Table 1
Table 1
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The proportion of participants reporting “ever” use of a cellular phone prior to the interview was high (cases, 92%; controls, 91%). Sixty percent of cases and 61% of controls were classified as regular phone users in the period at least 1 year prior to the reference date. Risk of a pituitary tumor was not increased in regular phone users overall (odds ratio [OR] = 0.9; 95% confidence interval [CI]: 0.7– 1.3) (Table 2), or in subjects who started using a phone 10 or more years prior to diagnosis (1.0; 0.5–1.9). Odds ratios were around unity for nearly all indices of phone use, with odds ratios marginally above 1.0 for subjects with cumulative use of more than 10 years, and those above the median of cumulative number of calls or hours of use. Odds ratios for these cumulative measures did not materially change after adjusting for handsfree use. Subjects who had more than the median hours of use (51 hours) 10 or more years prior to diagnosis had an odds ratio of 1.6 (0.8–3.6).

Table 2
Table 2
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Analyses by type of phone showed odds ratios in relation to regular use of 1.0 (0.6–1.6) for analog phones and 0.9 (0.7–1.3) for digital phones (Table 3). Odds ratios were above 1.0 for several categories of long-term use or cumulative use, but with wide confidence intervals because they were based on small numbers.

Table 3
Table 3
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Phone use, averaged out over the entire period of use, was not particularly heavy; controls made a median of 2.0 calls for 5.3 minutes per day. However, 30 subjects (13 cases, 17 controls) had a reported daily average use of 2 or more hours (range: 2–16.6 hours) and 23 subjects (6 cases, 17 controls) had a reported average of 20 or more calls per day (20–158 calls). High estimates of average daily duration of use were primarily in subjects reporting duration of use as a range of time per call; for example, if call duration was reported by a subject to be 1 to 30 minutes per call, and there were 20 calls per day, this would give an estimated 4.8 hours of daily use based on the midrange value (20 ×, 14.5 minutes = 290 minutes or 4.8 hours). Repeating the analyses using the lower value of reported ranges showed odds ratios of 0.8, 1.2, and 1.1 for less than 65, 65–362, and more than 362 hours of use, respectively. For cumulative number of calls, these odds ratio estimates were 0.8, 1.2, and 1.0 for fewer than 1996, 1996–7396, and more than 7396 calls, respectively. Using the upper value of ranges resulted in somewhat reduced odds ratios compared with those in Table 2.

Only 1 case and 1 control had undergone radiotherapy to the head more than 10 years prior to the reference date; this did not affect the results. Odds ratios were lower for women than for men; for example, in relation to regular use the ORs were 0.7 (0.5–1.2): and 1.1 (0.7–1.9) for men, but with no statistical evidence for effect modification by sex. Odds ratios were somewhat lower than those presented above when analyses were repeated without adjustment for Townsend score, but were similar after excluding from the analyses phone use during a period of 5 years, rather than 1 year, before diagnosis.

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In this population-based case–control study, risk of developing a pituitary tumor was not associated with regular phone use overall, nor with duration of use, time since first use, or cumulative number of calls or hours of use. Twenty-four cases and 48 controls started using a cellular phone between 10 and 17 years prior to the study; risks after longer induction times would need to be investigated in future studies.

Only 1 previous study, the Interphone study in Japan, has reported on pituitary tumor risk in relation to cellular phone use.13 Risk was not associated with regular use (OR = 0.9; 0.5–1.6), based on 101 cases and 161 controls, and was not associated with cumulative years of use or call time, or with type of phone used. The number of long-term users was smaller than in our study.

The pituitary gland is located within the sella turcica of the sphenoid bone at the base of the skull. Given the strong gradient through the head of energy deposition from radiofrequency fields from cellular phone use,4,5 the pituitary gland is thought to receive a level of exposure that is considerably lower than that in the tissues on the surface of the head in the immediate vicinity of the phone. If radiofrequency exposure affected cancer risks in all tissues equally, one would expect much less effect in pituitary tumor risk than in risk of other types of intracranial tumor. Given the lack of an established mechanism in which radiofrequency exposure affects cancer risk,3,6–10 it is unknown how sensitivity might vary between tissues. Previous studies of other types of intracranial tumor2,3 have not been restricted to highly exposed parts of the brain, however, because they included tumors on the opposite side of the head and tumors in other parts of the brain with low exposure levels (eg, midline tumors) and because of uncertainties about reported side of phone use. Recent efforts to take account of spatial relationships between tumor location and the distribution of radiofrequency exposure have improved on this.13

Exposure assessment in our study was based entirely on self-reported information and is likely to have been subject to considerable measurement error. Data on short-term recall of phone use were assessed in the Interphone study and showed large random and systematic errors, with underestimation of number of calls and overestimation of call time, the latter being greatest in heavy phone users.24 Such misclassification in exposure, if similar between cases and controls, could lead to odds ratios biased toward null in the presence of a true effect, and is an alternative explanation for the null findings in our study. Some evidence for differential misclassification was also provided by the Interphone study in that for cases (but not controls), the degree of overestimation of call time increased with increasing time before interview.25 Although this was based on few subjects with long-term data, such a bias (if true) could inflate estimates of risk in relation to phone use in more distant time periods.

Odds ratios in our study were derived using the combined group of never and nonregular phone users as baseline, which was necessary due to the small numbers of never users. Nonregular users included subjects with very low intensities of use and those who started use less than 18 months prior to the reference date, a period which is unlikely to be etiologically relevant. It is unlikely, therefore, that the composition of the baseline group has affected our results.

Some subjects in our study reported very high levels of cumulative hours of phone use or number of calls. High estimates of cumulative duration of use, in particular, appeared to be a result of subjects reporting a range of call durations. Using the lower or upper value of reported ranges instead of the midrange value affected estimates of cumulative use, and resulted in somewhat different odds ratios. These odds ratios were still not materially increased, and therefore do not affect inferences about the results.

Several other types of bias might have influenced our results. Participation bias is possible because only 43% of mailed controls took part in our study, and there was a disparity in participation rates between cases and controls (63% vs. 43%). The low control participation rate is due largely to unwillingness of people to take part in unpaid medical studies and high residential mobility in Southeast England. The true participation rate is likely to be appreciably higher, however, because some people will not have received the invitation letter at all (eg, because they moved). From additional data on probable addresses, we calculate that up to 63% of those contacted took part in the study, depending on the proportion of subjects who did not reply who had actually moved. In several Interphone studies there have been indications, from surveys of phone use among a proportion of nonparticipant controls26–28 that cellular phone users may have been overrepresented among participating controls. This might be the reason for the reduced risks in cellular phone users observed in several studies.2 We have no such survey data from our study, but we cannot exclude this possibility, even though the study documentation to controls did not mention cellular phones as 1 of the study hypotheses (unlike in some other Interphone centers17), which should have reduced the potential for such bias. Participants in our study were biased toward more affluent individuals, and among participants, controls were more affluent than cases. As cellular phone use, especially in the early years, was associated with greater affluence, we adjusted the analyses for Townsend deprivation score, leading to less strongly reduced odds ratios; hence, the possibility of residual confounding by socioeconomic status remains. Latent disease bias is also possible in that the presence of the tumor before diagnosis might have affected the patient's behavior and use of cellular phones, especially with slow-growing tumors such as those of the pituitary gland. Analyses of phone use in the period more than 5 years prior to diagnosis, however, showed similar results.

Tumors arising from the pituitary gland are usually benign. It is thought, based on autopsy and radiology studies, that small asymptomatic pituitary tumors are common and frequently remain undiagnosed.29 Our study is necessarily of cases with pituitary tumors that have come to clinical attention. We made much effort to ensure that only study participants with tumors first diagnosed during the study period were included in the analyses, but it is possible that such tumors had been present a long time before diagnosis. The long latency time of these tumors makes the evaluation of risks after longer-term phone use particularly critical.

Risk factors for pituitary adenomas are largely unknown. The syndromes multiple endocrine neoplasia (MEN1) and Carney complex (CNC) are associated with increased susceptibility but account for a small proportion of cases.30,31 Germline mutations in the aryl hydrocarbon receptor interacting protein (AIP) gene have recently been reported to account for a sizeable proportion of growth hormone–secreting pituitary adenomas.32 Exposure to ionizing radiation might also increase risk.23 Few studies have addressed other potential risk factors; some reported on oral contraceptive use and parity,33–37 with no conclusive results. It seems unlikely that any of these factors might have materially affected our results in relation to cellular phone use. In conclusion, we found no evidence that cellular phone use is associated with risk of pituitary tumors for the induction periods and intensities of use observed.

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We thank E. Cardis and colleagues (International Agency for Research on Cancer), H. Møller, B. Plewa and S. Richards (Thames Cancer Registry), and S. Hepworth (University of Leeds) for their contribution; the study team, apart from the authors named above, consisted of D. Hogben, A. Butlin, A. Hart, M. Pelerin, R. Knight, C. Parsley, J. Owens, K. Sampson, and M. Swanwick. Additionally, we thank the following individuals for help they provided: M. Cronin, T. Foster, S. Furey, F. Jones, N. Mendoza, F. Taylor (Charing Cross Hospital); P. Bullock, R. Selway, L. Smith, N. Thomas (King's College Hospital); J. Grieve, T. Pearce, M Powell (National Hospital for Neurology and Neurosurgery); J. Benjamin, J. Pollock (Oldchurch Hospital); J. Armstrong, G. Critchley, C. Hardwidge, J. Norris (Princess Royal Hospital); M. Allen, T. Dale, N. Dorward, D. Farraday-Browne, D. McLaughlin, R. Maurice-Williams, K. Pigott, B. Reynolds, C. Shah, C. Shieff (Royal Free Hospital); F. Afshar, H. Sabin (Royal London Hospital); M. Brada, D. Guerrero, D. Traish (Royal Marsden Hospital); S. Whitaker (Royal Surrey County Hospital); P. Plowman (St. Bartholomew's Hospital); C. Bramwell, A. Bell, F. Johnston, H. Marsh, A. Martin, A. Moore, S. Stapleton, (St. George's Hospital); and R. Beaney (St Thomas’ Hospital).

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