Mary L. McBride

A pooled analysis of magnetic fields and childhood leukaemia

Previous studies have suggested an association between exposure to 50–60 Hz magnetic fields (EMF) and childhood leukaemia. We conducted a pooled analysis based on individual records from nine studies, including the most recent ones. Studies with 24/48-hour magnetic field measurements or calculated magnetic fields were included. We specified which data analyses we planned to do and how to do them before we commenced the work. The use of individual records allowed us to use the same exposure definitions, and the large numbers of subjects enabled more precise estimation of risks at high exposure levels. For the 3203 children with leukaemia and 10 338 control children with estimated residential magnetic field exposures levels < 0.4 μT, we observed risk estimates near the no effect level, while for the 44 children with leukaemia and 62 control children with estimated residential magnetic field exposures ≥ 0.4 μT the estimated summary relative risk was 2.00 (1.27–3.13), P value = 0.002). Adjustment for potential confounding variables did not appreciably change the results. For North American subjects whose residences were in the highest wire code category, the estimated summary relative risk was 1.24 (0.82–1.87). Thus, we found no evidence in the combined data for the existence of the so-called wire-code paradox. In summary, the 99.2% of children residing in homes with exposure levels < 0.4 μT had estimates compatible with no increased risk, while the 0.8% of children with exposures ≥ 0.4 μT had a relative risk estimate of approximately 2, which is unlikely to be due to random variability. The explanation for the elevated risk is unknown, but selection bias may have accounted for some of the increase.

The INTERPHONE study: design, epidemiological methods, and description of the study population

The very rapid worldwide increase in mobile phone use in the last decade has generated considerable interest in the possible health effects of exposure to radio frequency (RF) fields. A multinational case–control study, INTERPHONE, was set-up to investigate whether mobile phone use increases the risk of cancer and, more specifically, whether the RF fields emitted by mobile phones are carcinogenic. The study focused on tumours arising in the tissues most exposed to RF fields from mobile phones: glioma, meningioma, acoustic neurinoma and parotid gland tumours. In addition to a detailed history of mobile phone use, information was collected on a number of known and potential risk factors for these tumours. The study was conducted in 13 countries. Australia, Canada, Denmark, Finland, France, Germany, Israel, Italy, Japan, New Zealand, Norway, Sweden, and the UK using a common core protocol. This paper describes the study design and methods and the main characteristics of the study population. INTERPHONE is the largest case–control study to date investigating risks related to mobile phone use and to other potential risk factors for the tumours of interest and includes 2,765 glioma, 2,425 meningioma, 1,121 acoustic neurinoma, 109 malignant parotid gland tumour cases and 7,658 controls. Particular attention was paid to estimating the amount and direction of potential recall and participation biases and their impact on the study results.

Risk of second cancer among women with breast cancer

A large number of women survive a diagnosis of breast cancer. Knowledge of their risk of developing a new primary cancer is important not only in relation to potential side effects of their cancer treatment, but also in relation to the possibility of shared etiology with other types of cancer. A cohort of 525,527 women with primary breast cancer was identified from 13 population‐based cancer registries in Europe, Canada, Australia and Singapore, and followed for second primary cancers within the period 1943–2000. We used cancer incidence rates of first primary cancer for the calculation of standardized incidence ratios (SIRs) of second primary cancer. Risk of second primary breast cancer after various types of nonbreast cancer was also computed. For all second cancer sites combined, except contralateral breast cancer, we found a SIR of 1.25 (95% CI = 1.24–1.26) on the basis of 31,399 observed cases after first primary breast cancer. The overall risk increased with increasing time since breast cancer diagnosis and decreased by increasing age at breast cancer diagnosis. There were significant excesses of many different cancer sites; among these the excess was larger than 150 cases for stomach (SIR = 1.35), colorectal (SIR = 1.22), lung (SIR = 1.24), soft tissue sarcoma (SIR = 2.25), melanoma (SIR = 1.29), non‐melanoma skin (SIR = 1.58), endometrium (SIR = 1.52), ovary (SIR = 1.48), kidney (SIR = 1.27), thyroid gland (SIR = 1.62) and leukaemia (SIR = 1.52). The excess of cancer after a breast cancer diagnosis is likely to be explained by treatment for breast cancer and by shared genetic or environmental risk factors, although the general excess of cancer suggests that there may be additional explanations such as increased surveillance and general cancer susceptibility.

Recall bias in the assessment of exposure to mobile phones

Most studies of mobile phone use are case–control studies that rely on participants’ reports of past phone use for their exposure assessment. Differential errors in recalled phone use are a major concern in such studies. INTERPHONE, a multinational case–control study of brain tumour risk and mobile phone use, included validation studies to quantify such errors and evaluate the potential for recall bias. Mobile phone records of 212 cases and 296 controls were collected from network operators in three INTERPHONE countries over an average of 2 years, and compared with mobile phone use reported at interview. The ratio of reported to recorded phone use was analysed as measure of agreement. Mean ratios were virtually the same for cases and controls: both underestimated number of calls by a factor of 0.81 and overestimated call duration by a factor of 1.4. For cases, but not controls, ratios increased with increasing time before the interview; however, these trends were based on few subjects with long-term data. Ratios increased by level of use. Random recall errors were large. In conclusion, there was little evidence for differential recall errors overall or in recent time periods. However, apparent overestimation by cases in more distant time periods could cause positive bias in estimates of disease risk associated with mobile phone use.

Risk of brain tumours in relation to estimated RF dose from mobile phones: Results from five interphone countries

Objectives The objective of this study was to examine the associations of brain tumours with radio frequency (RF) fields from mobile phones. Methods Patients with brain tumour from the Australian, Canadian, French, Israeli and New Zealand components of the Interphone Study, whose tumours were localised by neuroradiologists, were analysed. Controls were matched on age, sex and region and allocated the ‘tumour location’ of their matched case. Analyses included 553 glioma and 676 meningioma cases and 1762 and 1911 controls, respectively. RF dose was estimated as total cumulative specific energy (TCSE; J/kg) absorbed at the tumours estimated centre taking into account multiple RF exposure determinants. Results ORs with ever having been a regular mobile phone user were 0.93 (95% CI 0.73 to 1.18) for glioma and 0.80 (95% CI 0.66 to 0.96) for meningioma. ORs for glioma were below 1 in the first four quintiles of TCSE but above 1 in the highest quintile, 1.35 (95% CI 0.96 to 1.90). The OR increased with increasing TCSE 7+ years before diagnosis (p-trend 0.01; OR 1.91, 95% CI 1.05 to 3.47 in the highest quintile). A complementary analysis in which 44 glioma and 135 meningioma cases in the most exposed area of the brain were compared with gliomas and meningiomas located elsewhere in the brain showed increased ORs for tumours in the most exposed part of the brain in those with 10+ years of mobile phone use (OR 2.80, 95% CI 1.13 to 6.94 for glioma). Patterns for meningioma were similar, but ORs were lower, many below 1.0. Conclusions There were suggestions of an increased risk of glioma in long-term mobile phone users with high RF exposure and of similar, but apparently much smaller, increases in meningioma risk. The uncertainty of these results requires that they be replicated before a causal interpretation can be made.

Childhood leukemia and socioeconomic status in Canada.

Background: Leukemia is one of the most common potentially fatal illnesses in children, and its causes are not well understood. Although socioeconomic status (SES) has been related to leukemia in some studies, this apparent association may be attributable to ascertainment or participation bias. This study was undertaken to determine whether there is a difference in incidence of childhood leukemia for different levels of SES, as measured by neighborhood income, in an unselected population case group. Methods: All cases of childhood leukemia diagnosed in the years 1985–2001 were identified from population-based cancer registries in Canada. Postal codes for the place of residence at diagnosis were used to ascertain the census neighborhoods for cases. We constructed neighborhood-based income quintiles from census population data, and stratified the population at risk by sex and 5-year age groupings. Age-standardized incidence rates and 95% confidence intervals (CIs) were calculated. We used Poisson regression to compare incidence rate ratios (RRs) across income quintiles. Results: A slightly lower relative risk of childhood leukemia was observed in the poorest quintile compared with the richest (RR = 0.87; 95% CI = 0.80–0.95). The lower risk in the poorest quintile was restricted to acute lymphoid leukemia (0.86; 0.78–0.95) and was strengthened slightly by restriction to urban areas (0.83; 0.74–0.93). Conclusions: This analysis suggests that high SES is a true risk factor for childhood leukemia and that inconsistent results from other studies may be related to differences in case ascertainment or study participation.

A population-based study of childhood myelodysplastic syndrome in British Columbia, Canada

Myelodysplastic syndrome (MDS) is considered to be very rare in children. However, the only two published population‐based studies reported widely divergent incidence figures. To further explore the epidemiology of childhood MDS and to evaluate the accuracy of cancer registry and treatment trial data, we conducted a population‐based study of children aged 0–14 years in British Columbia (BC), Canada, between 1982 and 1996. MDS was diagnosed in 31 cases corresponding to an annual incidence of 3.2 per million children or 6% of all leukaemias, compared with an incidence of 6.0/million for acute myeloid leukaemia (AML), and of 0.5/million for chronic myeloid leukaemia. There was a non‐significant (P = 0.19) trend toward an increase in MDS incidence with time, the increase was partly explained by an increasing number of patients with Down syndrome. Associated abnormalities were found in 48% of the MDS cases with Down syndrome as the most common (seven cases). Only one third of the MDS cases were correctly registered in the Cancer Registry and less than half of the eligible MDS patients were enrolled on a cooperative group study. Data on MDS from treatment‐based studies and cancer registries were inaccurate and seemed to significantly underestimate the incidence of MDS in children.

Second primary cancers among 109 000 cases of non-Hodgkin's lymphoma

An analysis of other primary cancers in individuals with non-Hodgkins lymphoma (NHL) can help to elucidate this cancer aetiology. In all, 109 451 first primary NHL were included in a pooled analysis of 13 cancer registries. The observed numbers of second cancers were compared to the expected numbers derived from the age-, sex-, calendar period- and registry-specific incidence rates. We also calculated the standardised incidence ratios for NHL as a second primary after other cancers. There was a 47% (95% confidence interval 43–51%) overall increase in the risk of a primary cancer after NHL. A strongly significant (P<0.001) increase was observed for cancers of the lip, tongue, oropharynx*, stomach, small intestine, colon*, liver, nasal cavity*, lung, soft tissues*, skin melanoma*, nonmelanoma skin*, bladder*, kidney*, thyroid*, Hodgkins lymphoma*, lymphoid leukaemia* and myeloid leukaemia. Non-Hodgkins lymphoma as a second primary was increased after cancers marked with an asterisk. Patterns of risk indicate a treatment effect for lung, bladder, stomach, Hodgkins lymphoma and myeloid leukaemia. Common risk factors may be involved for cancers of the lung, bladder, nasal cavity and for soft tissues, such as pesticides. Bidirectional effects for several cancer sites of potential viral origin argue strongly for a role for immune suppression in NHL.

Second malignancies among survivors of germ-cell testicular cancer: A pooled analysis between 13 cancer registries

We investigated the risk of second malignancies among 29,511 survivors of germ‐cell testicular cancer recorded in 13 cancer registries. Standardized incidence ratios (SIRs) were estimated comparing the observed numbers of second malignancies with the expected numbers obtained from sex‐, age‐, period‐ and population‐specific incidence rates. Seminomas and nonseminomas, the 2 main histological groups of testicular cancer, were analyzed separately. During a median follow‐up period of 8.3 years (0–35 years), we observed 1,811 second tumors, with a corresponding SIR of 1.65 (95% confidence interval (CI): 1.57–1.73). Statistically significant increased risks were found for fifteen cancer types, including SIRs of 2.0 or higher for cancers of the stomach, gallbladder and bile ducts, pancreas, bladder, kidney, thyroid, and for soft‐tissue sarcoma, nonmelanoma skin cancer and myeloid leukemia. The SIR for myeloid leukemia was 2.39 (95% CI: 1.41–3.77) after seminomas, and 6.77 (95% CI: 4.14–10.5) after nonseminomas. It increased to 37.9 (95% CI: 18.9–67.8; based on 11 observed cases of leukemia) among nonseminoma patients diagnosed since 1990. SIRs for most solid cancers increased with follow‐up duration, whereas they did not change with year of testicular cancer diagnosis. Among subjects diagnosed before 1980, 20 year survivors of seminoma had a cumulative risk of solid cancer of 9.6% (95% CI: 8.7–10.5%) vs. 6.5% expected, whereas 20 years survivors of nonseminoma had a risk of 5.0% (95% CI: 4.2–6.0%) vs. 3.1% expected. In conclusion, survivors of testicular cancers have an increased risk of several second primaries, where the effect of the treatment seems to play a major role.

Quantifying the Impact of Selection Bias Caused by Nonparticipation in a Case–Control Study of Mobile Phone Use

PURPOSE To quantitatively assess the impact of selection bias caused by nonparticipation in a multinational case-control study of mobile phone use and brain tumor. METHODS Non-response questionnaires (NRQ) were completed by a sub-set of nonparticipants. Selection bias factors were calculated based on the prevalence of mobile phone use reported by nonparticipants with NRQ data, and on scenarios of hypothetical exposure prevalence for other nonparticipants. RESULTS Regular mobile phone use was reported less frequently by controls and cases who completed the NRQ (controls, 56%; cases, 50%) than by those who completed the full interview (controls, 69%; cases, 66%). This relationship was consistent across study centers, sex, and age groups. Lower education and more recent start of mobile phone use were associated with refusal to participate. Bias factors varied between 0.87 and 0.92 in the most plausible scenarios. CONCLUSIONS Refusal to participate in brain tumor case-control studies seems to be related to less prevalent use of mobile phones, and this could result in a downward bias of around 10% in odds ratios for regular mobile phone use. The use of simple selection bias estimation methods in case-control studies can give important insights into the extent of any bias, even when nonparticipant information is incomplete.