We therefore attempted to clarify associations between the risk of prostate cancer before the age of 70 years and several aspects of sexual activity, with a special focus on total ejaculations rather than sexual intercourse alone. We also attempted to investigate whether associations with sexual activity were age‐dependent by estimating ejaculatory frequencies in the patients’ third, fourth and fifth decades.

For the association with sexual activity the review noted possible hormonal influences but concluded that; ‘the mechanism through which frequency of sexual activity may be related to prostate cancer is unclear’[ 2 ]. In the studies reviewed, the measurement of sexual activity was almost always restricted to episodes of sexual intercourse. If the relevant ‘exposure’ from sexual activity is the amount of prostatic secretion, then this restriction may lead to substantial misclassification, especially in early adulthood and in later life, when sexual intercourse may be less frequent.

Although it is generally accepted that prostate cancer is a hormone‐dependent malignancy, its causes remain poorly understood [ 1 ]. The idea that prostate cancer risk might be related to men's hormonal milieu, and therefore possibly to differences in sexual activity, has been pursued in several epidemiological investigations; a recent meta‐analysis [ 2 ] of published studies reported a higher risk associated with a history of sexually transmitted infections, with an odds ratio (OR) of 1.4 (95% CI 1.2–1.7), with frequency of sexual activity (1.2, 1.1–1.3 for an increase of three times a week) and with increasing numbers of sexual partners (1.2, 1.1–1.3 for an increase of 20 partners). Although ever‐married men were suggested to be at a slightly increased risk (1.17, 0.98–1.40), the meta‐analysis did not support associations with multiple marriages, age at first intercourse or age at first marriage. Those authors concluded that there was support for an association with sexually transmitted infection and with sexual activity. However, an infection theory is inconsistent with other observations, e.g. the lack of correlation between the incidence of prostate cancer and cancer of the cervix [ 1 ], and the increased mortality from prostate cancer in presumably celibate Roman Catholic priests [ 3 ].

Lastly, to examine the effect of the number of ejaculations in the third decade, considering the number of ejaculations in the fifth (as these two were the least correlated decades) a variable was fitted with 16 categories, each representing a combination of the quartiles for the two decades.

Unconditional multiple logistic regression was used, adjusting for reference age, study centre, calendar year, family history and country of birth. Polytomous logistic regression was used to assess differences in ORs by tumour grade [ 5 ]. All continuous variables were divided into quantiles of the distribution in the controls, except for the number of male partners, which was dichotomized because of the few exposed subjects. For a direct comparison, the number of female partners was presented in the categories used by Rosenblatt et al. [ 6 ] (1, 2–4, 5–14, 15–29, > 29).

Direct interviews were conducted, usually at the man's home, at the end of which a sexual history questionnaire was given to each subject to complete by himself while the interviewer checked through the dietary questionnaires for completeness and obvious errors. The interviewer was available to respond only to the man's questions of definition (each interviewer was provided with plain‐language explanations of terms used in the questions). The questionnaire was developed using focus groups to ensure that the language was appropriate and that the questions were acceptable, comprehensible and likely to be answered. On this basis, we chose not to ask questions about sexually transmitted diseases, but included questions about the number of sexual partners before and after the age of 30 years (male and female partners were asked about this separately). The questionnaire focused on ejaculation irrespective of the context in which it occurred (intercourse with another, masturbation, nocturnal emissions, etc.). Men were asked their age at first ejaculation, the maximum number of ejaculations ever experienced in 24 h, and to estimate the average number of times that they had ejaculated per week in their most sexually active year in each of three decades of age (i.e. third, fourth and fifth). To preserve privacy, the subject sealed the questionnaire in an unmarked envelope before returning it to the interviewer.

We carried out a population‐based, case‐control study of prostate cancer in Australia, details of which were published previously [ 4 ]. Its principal purpose was to examine associations between lifestyle factors and the diagnosis of ‘clinically important’ prostate cancer. To this end we excluded tumours that were well‐differentiated (i.e. low‐grade or Gleason score < 5). We also focused on early‐onset disease, as we were interested in finding factors relevant to the prevention of prostate cancer that would cause premature mortality. Eligible cases comprised all male residents of Melbourne, Sydney and Perth diagnosed from 1994 to 1997 who were aged < 70 years at diagnosis and who were registered to vote on the State Electoral Rolls (adult registration to vote is compulsory in Australia). All cases diagnosed before the age of 60 years and random samples of half of cases diagnosed aged 60–64 and 25% of cases diagnosed aged 65–69 years were selected. Controls were randomly selected from the current State Electoral Rolls, and were frequency‐matched to the age distribution of the cases in a ratio of one control per case. The response rate was 65% in cases and 50% in controls [ 4 ]. In cases the response rates declined with age (40–49, 74%; 50–59, 68%; 60–69, 62%) and this was also true in controls (56%, 53% and 49%, respectively). Prior approval of the study protocol was obtained from all relevant Human Research Ethics Committees [ 4 ].

When the log OR was estimated as a linear function of the quartile of the number of ejaculations, each quartile increase changed the OR by − 15% for the third ( P < 0.001), − 12% for the fourth ( P = 0.007) and − 7% for the fifth decade ( P = 0.1). When fitted in the same model, the estimates for the second decade remained at − 15% ( P = 0.007), while the estimates for the other two became negligible at − 3% ( P = 0.7) and + 2% ( P = 0.8), respectively. This observation was confirmed by using combinations of the quartiles in the third and fifth decades in the logistic model ( Table 3 ). The decrease in risk with increasing ejaculatory frequency in the third remained irrespective of the ejaculatory frequency in the fifth decade.

Adjusting for ejaculatory frequency had little influence on any associations with the other variables reported in Table 2 . Furthermore, analyses restricted to the married/living as married cases, and analyses in which educational status or marital status was also controlled for, gave estimates that differed negligibly from those presented. OR estimates did not differ between moderate and high‐grade disease.

There was no association with either age at first ejaculation or with maximum ejaculatory frequency in 24 h ( P = 0.8 and 0.2, respectively). Greater ejaculatory frequency in the most sexually active year in each of the three decades was associated with a significantly lower risk, with men in the upper quartile of ejaculatory frequency having about two‐thirds the risk of those in the lower quartile for the third and fourth decade, and four‐fifths the risk for the fifth. Ejaculatory frequencies in each of the decades were correlated with one another ( r = 0.5–0.7). A similar finding related to the total number of ejaculations over the three decades. Those who reported an average of four to five or more ejaculations per week had two‐thirds the risk compared with those who, on average, ejaculated less than three times per week.

Table 2 shows that those ‘never married’ had almost half the risk of the ‘ever married’ ( P = 0.05), but there was no evidence that the total number of female sexual partners was associated with risk ( P = 0.8). Men with a history of male sex partners had about two‐thirds the risk ( P = 0.2) but this was based on few men (≈ 2.5% of controls). Nor was there any evidence that the number of children was associated with risk; the OR (95% CI) for one, two, three and more than three children compared with no children were 0.97 (0.73–1.29), 1.09 (0.81–1.46) and 0.95 (0.70–1.28), respectively ( P trend = 0.61).

After excluding men with missing data on the variables to be controlled for in the analysis, there remained 1079 cases and 1259 controls. Table 1 shows that cases were more likely than controls to be born in Australia, and to have at least one first‐degree relative affected with prostate cancer. There were no major differences in age, education, smoking habit or marital status distributions between cases and controls.

DISCUSSION

In this large case‐control study of aspects of male sexual life and prostate cancer risk, there was no association with the number of female sexual partners or children, but a positive association with being married. There was also no association with the maximum number of ejaculations ever made in 24 h but a negative association with men's frequency of ejaculations, especially in their most sexually active year during their 20s.

We considered the extent to which our findings might be caused by bias or confounding. Although the response in controls was low, their sociodemographic profile was similar to that of men in the National Health Survey in 1995 [8] and similar to those found in some other case‐control studies in recent years [9-11]. Because of the widespread PSA testing that occurred during recruitment [12], we compared moderate‐ with high‐grade tumours; the associations were at least as strong for high‐grade prostate cancer. Having controlled for the strongest established risk factors (age and family history) and, given the lack of other known risk factors, we consider that confounding is unlikely to have influenced our findings. We have no information on the sexual histories of the cases and controls not responding, or whether they differ between cases and controls, and therefore cannot exclude the possibility of response bias influencing our findings.

COMPARISONS WITH OTHER STUDIES The present positive association with marital status was reported by several studies [2] but its meaningfulness today, given the contemporary heterogeneity of marital status, is difficult to define. Lack of association with the number of female sexual partners is consistent with about half of the published case‐control studies covered in the meta‐analysis [2] but contrasts strongly with a recent case‐control study in Seattle [6] to which we matched our analysis. There are several reports of positive associations with venereal disease and with high‐risk behaviour, e.g. intercourse with prostitutes, having sex without condoms and having many sex partners [2]. However, an infection hypothesis is not the only possible explanation for these observations, as the activities they describe could be markers for having a strong sex drive that may hypothetically be associated with increased prostate cancer risk via hormone levels, but this hypothesis is not consistent with the present decreased risk associated with increased number of ejaculations. There might be three major reasons why our findings for sexual activity are opposite to those from some other studies; (i) the scope of sexual activity included; (ii) the temporal frame covered; and (iii) how the questionnaire was administered. Most other studies have measured sexual activity by reference solely to sexual intercourse [2]. Few have asked about all forms of ejaculation [13, 14]. As prostatic secretion is not limited solely to episodes of sexual intercourse, studies that focus only on sexual intercourse do not fully ascertain the exposure, i.e. prostatic secretion by ejaculation. However, should there be a causal sexually transmitted infectious agent, our exposure measure of ‘total ejaculations’ would dilute any such exposure by including ejaculations experienced without having sexual intercourse. If this is the case, had we been able to remove ‘ejaculations associated with sexual intercourse’, there should have been an even stronger protective effect of other ejaculations. The few studies that measured total ejaculations are more consistent in their findings. Banerjee [13] described a significantly lower frequency of ejaculations in cases than in controls during the sexually active parts of their lives. Although statistically insignificant, Hsieh et al.[14] also found suggestive evidence that increased ejaculatory frequency early in life might reduce the risk. Our finding is also consistent with that of Steele et al.[15], of an increased risk of prostate cancer after some period of reduced sexual activity. It is also consistent with the hypothesis of Isaac [16], that infrequent ejaculation could increase the risk of prostate cancer because of the possible stagnation of carcinogenic secretions in the prostatic acini. Apart from age at first marriage, few studies [13, 14] have measured sexual activity at different times of life. We found that ejaculatory frequencies in the third to fifth decades were highly correlated, but when they were modelled together the strongest effect was for the number of ejaculations in the second, while the effects of the number of ejaculations at other ages became inconsequential. This observation once more indicates earlier life experiences as predictors of much later outcomes. In attempting to compare studies from different societies, defined both geographically and historically, the probability must be considered that sexual mores and behaviour are socio‐culturally determined and have changed over time. In many communities the method of measuring sexual activity might influence not only the scope of questions that might be asked but also the response. Other studies have used both direct (face‐to‐face) [13, 17] and telephone interviews [6, 18], compared with the present, in which we used a self‐administered questionnaire [4]. It is difficult to believe that men's responses to questions of sexual activity, use of prostitutes, sexual orientation and sexually transmitted disease would not be influenced by the age and gender of the interviewer in a direct or telephone setting, as there would be a strong inclination to give more socially acceptable answers, particularly by participants in the older studies and by older respondents.