IJE Advance Access originally published online on August 2, 2008
International Journal of Epidemiology 2008 37(6):1304-1313; doi:10.1093/ije/dyn155
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Meningioma and mobile phone use—a collaborative case-control study in five North European countries
1 STUK – Radiation and Nuclear Safety Authority, Helsinki, Finland.
2 Tampere School of Public Health, University of Tampere, Tampere, Finland.
3 Section of Epidemiology, Institute of Cancer Research, Sutton, UK.
4 Institute of Cancer Epidemiology, Danish Cancer Society, Copenhagen, Denmark.
5 Institute of Environmental Medicine, Karolinska Institutet, Stockholm, Sweden.
6 Institute of Population-based Cancer Research, The Cancer Registry of Norway, Oslo, Norway.
7 Norwegian Radiation Protection Authority, Østerås, Norway.
* Corresponding author. STUK-Radiation and Nuclear Safety Authority, P.O. Box 14, FIN-00881 Helsinki, Finland. E-mail: anna.lahkola{at}stuk.fi
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Abstract |
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Background Use of mobile telephones has been suggested as a possible risk factor for intracranial tumours. To evaluate the effect of mobile phones on risk of meningioma, we carried out an international, collaborative case-control study of 1209 meningioma cases and 3299 population-based controls.
Methods Population-based cases were identified, mostly from hospitals, and controls from national population registers and general practitioners’ patient lists. Detailed history of mobile phone use was obtained by personal interview. Regular mobile phone use (at least once a week for at least 6 months), duration of use, cumulative number and hours of use, and several other indicators of mobile phone use were assessed in relation to meningioma risk using conditional logistic regression with strata defined by age, sex, country and region.
Results Risk of meningioma among regular users of mobile phones was apparently lower than among never or non-regular users (odds ratio, OR = 0.76, 95% confidence interval, CI 0.65, 0.89). The risk was not increased in relation to years since first use, lifetime years of use, cumulative hours of use or cumulative number of calls. The findings were similar regardless of telephone network type (analogue/digital), age or sex.
Conclusions Our results do not provide support for an association between mobile phone use and risk of meningioma.
Keywords cellular phones, brain neoplasms, case-control studies
Accepted 30 June 2008
Meningiomas are neoplasms originating from the meningeal tissue covering the brain and spinal cord. They are usually benign, with 1–3% exhibiting malignant growth.1 The incidence of meningiomas varies between populations, being higher among women than men.2 The aetiology of meningiomas has remained elusive, with some hereditary syndromes (mainly neurofibromatosis type 2 and tuberous sclerosis) and high doses of ionizing radiation among the few established risk factors.3 Radiofrequency electromagnetic fields emitted by mobile telephones has been suggested as a possible risk factor for meningiomas, mainly based on the analogy with ionizing radiation and the proximity of the meningeal tissue to the handset, i.e. the source of the radiofrequency field.
Previous studies of meningioma risk in relation to mobile phone use have included a relatively small number of study subjects with long-term exposure. Although some positive findings have been reported, so far the totality of epidemiological evidence does not demonstrate an increase in risk of meningiomas related to mobile phone use.4–9 We conducted a large international study to assess the possible association between use of mobile phones and risk of intracranial meningioma, using a shared protocol of the INTERPHONE study coordinated by the International Agency for Research on Cancer.10 The data for this study were collected from five Northern European countries, Denmark, Finland, Norway, Sweden and Southeast England, where mobile phone use has been common for at least a decade.11 Results from the Danish, Swedish and Norwegian studies have been published earlier.12–14 Collaborative analyses of acoustic neuroma15 and glioma16 risks based on these studies have also been reported previously.
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Materials and methods |
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Selection of study subjects
The study was carried out in Denmark (nationwide), Finland (98% of the national population, excluding Northern Lapland and Åland), Norway (the Southern and Middle parts, representing 90% of population), Sweden (Umeå, Stockholm, Gothenburg and Lund regions, 65% of population) and the United Kingdom (Thames region of Southeast England, 23% of population).
Eligibility criteria for cases included age 20–69 years in the Nordic countries and 18–59 years in Southeast England, residence in the study area and diagnosis of intracranial meningioma (International Classification of Diseases for Oncology, Third Edition, codes 9530–9539) between 2000 and 2004 (the exact study periods were slightly different between countries). Cases were identified through neurosurgery, oncology and neurology departments of several hospitals in the study areas. In addition, the national or regional cancer registries were used to evaluate and enhance completeness of coverage. The inclusion criteria were similar to earlier publications12–14 and the slightly smaller numbers of cases than in national reports are due to revised diagnosis, date of diagnosis or history of previous brain tumour. The diagnosis was microscopically confirmed in 93% of the cases (Table 1).
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Controls were selected through the national population registers in the Nordic countries. As there is no such register in the UK, the matched controls were randomly selected from general practitioners’ patient lists. In all countries, the controls were frequency-matched to the cases by sex, 5-year age group (at ascertainment date) and region of residence. Eligible cases were approached either by mail, or personally at the clinics, while the controls were first approached by mail. If the subjects contacted by mail did not respond, another letter was sent or the subject was approached by telephone. Before asking for participation, study subjects received both an invitation letter and written information about the study. Informed consent was obtained from all study participants. The ethical review of the study protocol was carried out by local committees in each country.
Data collection
Exposure assessment was based on personal interview that was typically performed at a hospital or at the subject's home and conducted by trained interviewers. Proxy interviews were used for 1.6% of cases and 0.1% of controls. Telephone interviews were common in Norway, where 48% of cases and 45% of controls were interviewed over the telephone, but infrequent in the other countries (0–4% among cases and 0–6% controls). The interview covered use of hand-held mobile phones, medical history, education and family history of brain tumours. Regular use of mobile phones was defined as making or receiving calls at least once a week for at least 6 months. For regular mobile phone users, a detailed history of use was obtained, including start and end dates as well as the frequency and laterality of use, type of phone, use of hands-free devices and other factors, such as type of telephone network. Show cards were used to facilitate recall of the phone models used. Information on the model of phones, calendar period of use, operator and network code of the phone number was used to classify phones as analogue or digital. In some of the countries, some information was also collected about Digital Enhanced Cordless Technology (DECT) and other cordless phones but this information was not used in the analyses, because the average power that they transmit is only 0.01 W versus 0.25/0.125 W with GSM 900/1800 phones and 1 W with NMT 900 phones.
Data handling and statistical analysis
Frequency-matching employed throughout the INTERPHONE study allowed us to utilize the entire control group recruited for all intracranial tumours (glioma, meningioma and acoustic neuroma) in the matched strata of the meningioma cases, to increase statistical power. The analysis covered several features of reported mobile phone use that are potentially relevant for exposure to radiofrequency electromagnetic fields. Cumulative hours of use was calculated from average number and duration of calls. The analyses were performed using both continuous and categorical exposure variables. In analyses of categorical exposure variables, the cut-points were chosen based on the distribution among controls. The reference category consisted of the never and non-regular users, with the other cut-points defined by the 50 and 75th percentiles of the exposure distribution among regular mobile phone users. In the analyses of cumulative number of calls and cumulative hours of mobile phone use, the exposure was adjusted for the reported use of hands-free devices. The exposure was reduced by 100% if the subject reported use of hands-free devices all the time, by 75% if most of the time, 50% if half of the time and 25% if sometimes but less than half of the time. An additional analysis of the subgroup with the highest cumulative number of calls and cumulative hours of use was performed with the cut-point defined as the value among the 10% of controls with the heaviest mobile phone use (among regular users).
For calculation of exposure indices, a reference date was determined for each subject. For cases, the reference date was the date of diagnosis. As the controls were interviewed on average later than the cases and as the prevalence of mobile phone use increased rapidly over time, the reference date for controls was corrected for this delay. Thus, the reference date for controls was defined as the interview date adjusted for the mean interval between the diagnostic and interview date of cases, and the difference between the mean interview date of cases and controls (in days) i.e. refdatecontrol = intdatecontrol – (mean intdatecases – mean diagdatecases) – (mean intdatecases – mean intdatecontrols). All mobile phone use within 1 year prior to the reference date was excluded from analysis, except when calculating the years since first use, which was evaluated up to the reference date.
The odds ratios (OR) for meningioma associated with mobile phone use were estimated with conditional logistic regression, with strata defined by sex, 5-year age group, region and country. Based on previous literature on aetiology of meningioma, highest educational level attained, family history of meningioma, radiotherapy to the head and neck region (at least 10 years before the reference date), and past diagnosis of neurofibromatosis or tuberous sclerosis of the subject were regarded as potential confounders. All the analyses were conducted both with and without considering the effects of the potential confounding factors. Adjustment for family history and socio-economic status in the analyses did not affect the results, nor did exclusion of subjects with neurofibromatosis, tuberous sclerosis or a history of radiotherapy to the head and neck region. Therefore, all the results reported are from analyses taking into account only the stratification variables (sex, 5-year age group, region and country).
Analyses were conducted separately by type of phone (analogue and digital). Furthermore, analyses were performed both based on the whole dataset and individually by country, and also by sex and by 10-year age group. Heterogeneity in the results between countries, 10-year age groups and sexes was assessed with the likelihood ratio test by comparing nested models, with one including both main effects and an interaction between the stratification factor and the exposure indicator, and the other including only the main effects.
The tumour location was assessed in relation to the reported predominant side of mobile phone use, using previously described methods.5,13 The Inskip method5 is based on case-only design, while the Lönn approach13 includes also the controls in the analysis. For cases, ipsilateral location was defined as tumour on the same side where the mobile phone was mainly held, whereas contralateral location was defined as tumour on the opposite side of where the phone was primarily held. For subjects reporting bilateral use (similar amount of use on both sides), both hemispheres were regarded as exposed. As controls have no tumour, an index laterality was randomly allocated to each of them in the analyses with the Lönn method. In this study, the method used by Lönn et al.13 was modified so that the reference group comprised only never and non-regular users but not those using the phone on the other side that were included in the reference group in the original analysis. All the analyses were conducted with the statistical software STATA (version 9).17
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Results |
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A total of 1629 eligible cases and 6581 controls were identified. Of the potential cases, 74% participated (1209 subjects, range 55–90% between countries, Table 1) and of the controls, 50% (3299 subjects, range 42–69%). The most frequent reasons for non-participation were refusal (9% of cases and 33% of controls), inability to contact the subject (8% of cases and 15% of controls) and illness or death (3% of cases and 0.5% of controls). Since there were strata with either no cases or no controls, several study subjects were excluded, leaving finally 1204 cases (of the 1209 who participated) and 2945 controls (of the 3299 who participated) in the analysis. The recall of mobile phone use in the interview was judged to be ‘good’ or ‘very good’ by the interviewers for 82% of the cases and 85% of the controls. The demographic characteristics of the study subjects are shown in Table 2.
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Regular mobile phone use was associated with an apparently reduced risk of meningioma, (OR 0.76, 95% confidence interval, CI: 0.65, 0.89), based on 48% regular users among cases and 58% among controls (Table 3). Years since first use or lifetime years of use were not associated with an increased risk of meningioma, as the OR for both variables was 0.99 (0.96, 1.01) per year. Cumulative number of calls was not associated with the risk of meningioma (OR = 1.00 per 10 000 calls, 95% CI: 0.96, 1.05, adjusted for hands-free devices). Those with the highest number of calls or hours of use (>21 753 calls and >1476 h, as among the highest 10% of controls) did not have an increased risk of meningioma (OR = 0.86, 95% CI: 0.60, 1.24 and 1.13, 95% CI 0.82–1.57, respectively). When only mobile phone use at least 10 years prior to the index date was considered, the results were not materially changed. The OR for cumulative number of calls 10 or more years ago was 1.07 (0.80, 1.41) per 10 000 calls, whereas for cumulative hours of use it was 1.02 (0.99, 1.05) per 100 h.
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Even though the distribution of cumulative call hours was skewed among both cases and controls, we also explored the linear relation of the call hours as a continuous variable to meningioma risk. It revealed an apparently positive association (OR 1.005 per 100 h, 95% CI: 1.001, 1.010), but this was driven by a small number of very high values which in turn reflected subjects with implausibly high reported mean daily hours of use. The result for cumulative hours of use was mostly based on subjects with more than 3 h of daily use (the 99th percentile for daily hours of use was 2.4 for controls and 3.5 for cases). When we excluded the subjects with more than 2 h of daily use (44 cases and 27 controls, corresponding to 1.7% of observations), no relation was detected. In the analysis of log-transformed number of call hours, no obvious association was observed (results not shown).
When analogue and digital networks were investigated, the results did not differ substantially from each other, or from the results based on all mobile phones. For analogue telephones, the OR for years since first use was 0.99 per year (0.96, 1.01), whereas for digital telephones, it was 0.97 per year (0.64, 1.00). When cumulative hours of use was analysed as a continuous variable, the OR for analogue phones was 1.003 (0.992, 1.014) and for digital phones 1.008 (1.002, 1.014) per 100 h. In the analyses of categorical variables, only minor differences between the two telephone types emerged, possibly due to relatively small numbers of cases within exposure categories (Table 4).
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The country-specific meningioma ORs for regular use were 0.87 (0.60, 1.27) for Denmark, 0.75 (0.56, 1.01) for Finland, 0.85 (0.57, 1.29) for Norway, 0.68 (0.49, 0.94) for Sweden and 0.72 (0.51, 1.01) for Southeast England. No indication of heterogeneity between countries was detected in results for regular use (P = 0.84) or any other indicator of mobile phone use (all P-values >0.2). Sex did not modify the relationship between mobile phone use and meningioma risk, as the OR for regular use compared with never or non-regular use was 0.79 (0.59, 1.06) for men and 0.75 (0.62, 0.89) for women. There was no heterogeneity by sex or age in the results for regular use (P = 0.74 and 0.70, respectively) or in any other indicator of mobile phone use (results not shown). The possible effect of the interview type (hospital/home/telephone) was also investigated and found out to have only marginal impact on the results (not shown). Additional analyses were also performed excluding cases without microscopically confirmed diagnosis (n = 80), but this did not affect the results (not shown). We also conducted analyses including only those subjects whose recall of mobile phone use was reported by the interviewers to be good or very good but the results were not altered substantially (not shown).
In the laterality analyses, both regular ipsilateral use (OR = 0.81, 95% CI: 0.66, 0.99), and regular contralateral use (OR = 0.67, 0.54, 0.83) were associated with apparently lower meningioma risk than never or non-regular mobile phone use. Years since first or lifetime years of ipsilateral use were not clearly related to meningioma, as the ORs were 1.05 (0.67, 1.65) and 0.99 (0.57, 1.73), respectively, for the subjects with the longest (>10 years) exposure history (Table 5). The laterality analyses were also conducted excluding subjects who had used a mobile phone on both sides of the head, but this had little effect on the results (not shown). When the laterality analyses were conducted using only cases (by the method similar to Inskip et al.5), an overall relative risk (RR) of 1.09 (Fisher's exact test, P = 0.10, two sided) was obtained for meningioma in relation to ipsilateral phone use, based on 212 exposed and 184 unexposed cases. For subjects with exposure duration >10 years (n = 30), the RR was 1.61 (P = 0.11), and for subjects whose first use was 10 or more years ago (n = 52), the RR was 1.26 (P = 0.17).
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Discussion |
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We did not find evidence of increased risk of meningioma in relation to mobile phone use, as regular use, years since first use, lifetime years of use or cumulative number of calls, were not associated with an increased risk. Our results are consistent with most previous studies and reviews on the issue.4–6,8,9,18,19
Cumulative number of hours (and to some extent also cumulative number of calls) and lifetime years of use represent cumulative exposure, ipsilateral use and possibly analogue phones represent higher intensity (magnetic field strength) and years since first use represent induction period. No increased risk was found in relation to duration of use, ipsilateral use or use of analogue phones. In the analysis of cumulative hours of use based on a categorical exposure indicator, there was no increased risk for meningioma, not even in the highest (10%) exposure category. When the continuous variable was used, some indication of an association was found, that was, however, based on small number of extreme and possibly erroneous values. The cumulative hours of use may contain uncertainties as an exposure indicator, as it is derived from the number and the duration of calls multiplied by the duration of mobile phone use. Additionally, both parameters were frequently reported also as ranges from which an average was used for estimating calling time. This may easily result in overestimation of exposure unless number and duration of calls are independent. If daily call-time of 30 min is composed of one 30-min call or thirty 1-min calls, it could be misinterpreted as on average 15 calls of 15-min duration, i.e. 225 min (nearly 4 h) constituting substantial overestimation.
It is also possible that cases over-report their mobile phone use (recall bias). Some evidence of this was observed, as in the distribution of cumulative hours of use the highest values among cases clearly exceeded those among controls and were generally based on implausibly high reported hours of mean daily use over long periods. When we performed an additional analysis of cumulative hours of use and excluded the subjects with more than 2 h of daily use corresponding to only 1.7% of observations, the OR was no longer increased.
The strengths of the study include a larger number of meningioma cases than in any previous report. This allowed us to obtain precise risk estimates and conduct detailed analyses of various aspects of mobile phone use. Furthermore, the number of subjects with long-term exposure is higher than in the earlier studies, due to the recent study period and study populations where large-scale mobile phone use was adopted early. This is essential, as the induction period may be long for a slowly developing benign tumour such as meningioma. A related aspect is larger cumulative exposure, providing a better opportunity to identify possible effects in a highly exposed group. On the other hand, the weaknesses of our study are the low participation rate among controls and the fact that mobile phone use and other exposures are based on self-reports. These limitations are shared by most other studies on the subject.
An apparently reduced risk was found related to regular use of mobile phones. A likely explanation for at least some of the risk reduction is selection bias. There is some empirical evidence for higher participation among mobile phone users than non-users for both cases and controls in the study.13,20 When mobile phone users are more willing to participate in epidemiological studies than non-users, exposure in the source population is overestimated, which may result in underestimation of the risk in relation to mobile phone use and may partly explain the decreased risk estimates detected in this study, if the selection was differential between cases and controls. Participation varying with level of use may distort the dose–response relation and lead to a J-shaped pattern, consistent with some of our results. The slightly lower proportion of controls with basic education only may be a related finding, as participation in research is commonly associated with a high educational level.21
Latent disease bias, where early symptoms of the disease may make cases less likely to become mobile phone users in the period preceding the diagnosis, may also explain the reduced risk estimate. However, the most common symptoms of meningioma are seizures and headache,22,23 which are unlikely to affect mobile phone use. Furthermore, we found little indication of such bias since the proportion of cases was similar to that of controls across the start of use categories, and also among those that started mobile phone use within a year before the reference date (20% vs 21%).
There was no evidence for confounding, as the crude results were very similar to those of the analyses accounting for highest educational level, family history of meningiomas, tuberous sclerosis, neurofibromatosis or previous radiotherapy to the head and neck region. Also, when calculating the exposure histories, the fact that controls were interviewed on average later than the cases was taken into account by adjustment, which is crucial when investigating an exposure such as mobile phone use that increases rapidly over time.
Considerable random error has been demonstrated in self-reported amount of mobile phone use.24–26 That is understandable, as mobile phone use is nowadays an unremarkable aspect of everyday life. As mobile phone use has increased during the years of use, the retrospectively reported amount of use may be influenced by not only past, but also current mobile phone use. In the INTERPHONE study, over-reporting of duration of calls by up to 50–100% was common.27 Non-differential misclassification of exposure is likely to bias the results towards the null, should there be an effect, particularly when a dichotomous or continuous measure of exposure is used.28 This is not equally obvious when categorical (multinomial) exposure indicators are used, as in our study. Information about regular vs non-regular use reported by the study subjects is likely to be more reproducible and crude measures might therefore be more robust than detailed, quantitative indices. Yet, the aforementioned selection bias may distort those analyses. Obviously, an objective source of exposure information, independent of the study subject, would provide highly valuable information. Yet, only one small cohort study has been able to use quantitative exposure information from operators, but it had a very short follow-up and inadequate number of events.29 Another cohort has also obtained information from subscriber lists, but only about ownership of a subscription, without any traffic data.8
In conclusion, our study does not provide evidence for an increased risk of meningioma in relation to mobile phone use. An apparent association was detected for cumulative hours of use as a continuous exposure indicator but this was based on a small number of extreme and possibly erroneous values. No such association was detected in the categorical analyses. Thus, the present findings do not suggest mobile phone use is associated with an increased risk of meningioma.
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All countries: the Quality of Life and Management of Living Resources program of European Union and the International Union against Cancer (UICC) (RCA/01/08). 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 Finnish study: Academy of Finland (grant no. 80921), Emil Aaltonen Foundation and Doctoral Programs for Public Health.
The UK study: the Mobile Telecommunications, Health and Research (MTHR) programme.
The Swedish Study: the Swedish Research Council.
The Nordic–UK collaboration thanks the interviewers for their contribution for data collection and Dr Elisabeth Cardis and the rest of the IARC team for their input to this study. We are also grateful to James Doughty and Jan Ivar Martinsen for programming work.
The Finnish centre thanks Dr J Jääskeläinen (Helsinki University Hospital), Dr S Valtonen (Turku University Hospital), Prof. J Koivukangas (Oulu University Hospital), Prof. M Vapalahti (Kuopio University Hospital), Dr T Kuurne and Dr H Haapasalo (Tampere University Hospital) and Prof. R Sankila (Finnish Cancer Registry) for their contributions to collection of the material.
The Norwegian centre thanks Jan Ivar Martinsen for data management and Karl G Blaasaas for his contribution with data collection, data management and analyses.
The Swedish centre thanks the Swedish Regional Cancer Registries and the hospital staff; especially the following key persons at the hospitals: Dr J Boethius, Dr O Flodmark, Prof. I Langmoen, Dr A Lilja, Dr T Mathiesen, Dr I Olsson Lindblom and Dr H Stibler (Karolinska University Hospital), Dr J Lycke, Dr A Michanek and Prof. L Pellettieri (Sahlgrenska University Hospital), Prof. T Möller and Prof. L Salford (Lund University Hospital), Dr T Bergenheim, Dr L Damber, Prof. R Henriksson and Dr B Malmer (Umeå University Hospital).
The Southeast England centre would like to thank all participants for their contribution to the study. They also thank Prof. H Møller, Mr B Plewa and Mr S Richards from the Thames Cancer Registry and the following neuropathologists, neurosurgeons, neuro-oncologists, clinical oncologists, neurologists, administrators and secretaries for the help they provided: Mr DG Hardy, Mr PJ Kilpatrick, Mr R Macfarlane (Addenbrooke's Hospital); Ms M Cronin, Ms T Foster, Ms S Furey, Dr M G Glaser, Ms F Jones, Mr ND Mendoza, Prof. ES Newlands, Mr KS O'Neill, Mr D Peterson, Ms F Taylor, Prof. J van Dellon (Charing Cross Hospital); Dr JJ Bending (Eastbourne District Hospital); Mr PR Bullock, Mr C Chandler, Mr B Chitnavis, Mr L Doey, Mr RW Gullan, Prof. CE Polkey, Mr R Selway, Mr MM Sharr, Ms L Smith, Prof. AJ Strong, Mr N Thomas (King's College Hospital); Dr GM Sadler (Maidstone Hospital); Dr S Short (Mount Vernon Hospital); Prof. S Brandner, Mr AD Cheesman, Miss JP Grieve, Mr WJ Harkness, Dr R Kapoor, Mr ND Kitchen, Mrs T Pearce, Mr MP Powell, Dr J Rees, Prof. F Scaravilli, Prof. DT Thomas, Mr LD Watkins (National Hospital for Neurology and Neurosurgery); Mr AR Aspoas, Mr S Bavetta, Mr J C Benjamin, Mr KM David, Mr JR Pollock, Dr E Sims (Oldchurch Hospital); Mrs J Armstrong, Mr J Akinwunmi, Mr G Critchley, Mr L Gunasekera, Mr C Hardwidge, Mr JS Norris, Dr PE Rose, Mr PH Walter, Mr PJ Ward, Dr M Wilkins (Princess Royal Hospital); Prof. TZ Aziz, Prof. D Kerr, Mr PJ Teddy (Radcliffe Infirmary); Ms M Allen, Ms T Dale, Mr R Bradford, Prof. AP Dhillon, Mr NL Dorward, Ms D Farraday-Browne, Dr DJ McLaughlin, Mr RS Maurice-Williams, Dr K Pigott, Ms B Reynolds, Ms C Shah, Mr C Shieff, Dr EM Wilson (Royal Free Hospital); Mr F Afshar, Mr HE Ellamushi, Prof. PM Richardson, Mr HI Sabin, Mr J Wadley (Royal London Hospital); Prof. M Brada, Mr D Guerrero, Dr FH Saran, Mrs D Traish (Royal Marsden Hospital); Dr S Whitaker (Royal Surrey County Hospital); Dr PN Plowman (St Bartholomew's Hospital); Mrs Carole Bramwell, Prof. A Bell, Mr F Johnston, Mr H Marsh, Mr A Martin, Mr PS Minhas, Miss A Moore, Mr S Stapleton, Dr S Wilson (St George's Hospital); Dr RP Beaney (St Thomas’ Hospital).
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